The Regulation of Occupations and Labour Market Outcomes In

The Regulation of Occupations and Labour Market Outcomes in Canada: Three Essays on the Relationship between Occupational Licensing, Earnings, and Internal Labour Mobility

Tingting Zhang

A thesis submitted in conformity with the requirements for the degree of Doctor of Philosophy Centre for Industrial Relations and Human Resources University of Toronto

The Regulation of Occupations and Labour Market Outcomes in Canada: Three Essays on the Relationship between Occupational Licensing, Earnings, and Internal Labour Mobility

Tingting Zhang

Doctor of Philosophy

Centre for Industrial Relations and Human Resources University of Toronto

2017 Abstract

The thesis begins with an introductory chapter that discusses the current state of occupational licensing research and motivates the analysis through the importance of occupational licensing, similar to that of unionization, to both labour market theory and public policy. The first paper in this thesis asks whether or not Canadian licensed workers earn higher wages than unlicensed workers. The paper also compares licensing pay premium to union wage premium. Based on longitudinal data from the Canadian Survey of Labour and Income Dynamics (SLID) from

1993 to 2011, a pay premium of approximately 12.0% is estimated for occupational licensing, slightly higher than the union wage premium of 9.0%. These results are based on a cross- section of respondents using Ordinary Least Square (OLS) estimates. Fixed-effect estimates from the longitudinal data (2.6% and 4.0% for licensing and unionization, respectively), however, are much lower than the OLS amounts, suggesting the importance of unobservable factors that are correlated with licensing and union status in determining the wage premium of workers.

The second paper further investigates whether or not wage premiums are uniform across the wage distribution. Using unconditional quantile regression methods, I investigated how occupational licensing and unionization impact the wage distribution of Canadian workers between 1998 and 2014. Unionization decreases wage inequality in upper-wage earners, but increases wage inequality in lower-wage earners. Occupational licensing, on the other hand, increases inequality across the entire wage distribution.

The third paper focuses on the impacts of occupational licensing and unionization on young workers’ interprovincial mobility decisions. Using Canadian longitudinal data from the

Canadian Survey of Labour and Income Dynamics (SLID) from 1993 to 2010, the results of both multilevel modelling analyses and linear probability models with clustered standard errors show that, unlike unionization which restricts labour mobility, individuals’ licensing status is not correlated with the likelihood of moving across provincial boundaries for young

Canadians aged 21–34.

The final component of the thesis is an Appendix that presents a newly constructed occupational licensing index, which is based on an authoritative Canadian jurisdictional review.

I define occupational licensing based on the exclusive-right-to-practice clause, and occupational certification based on the exclusive-right-to-title clause in the legislation, and construct licensing indicators to carry the empirical analyses of the effects of occupational licensing in Canada.

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Acknowledgments

This thesis represents much of what I have learned at the Centre for Industrial Relations and Human Resources, University of Toronto. I consider myself very fortunate to have had individuals at the Centre who helped me immensely and taught me so much. This thesis would never have taken shape without their input.

I am indebted first and foremost to my thesis committee, including Professors Michele Campolieti, Elizabeth Dhuey, and my co-supervisors, Professors Morley Gunderson and Rafael Gomez. I want to thank them for guiding me through the circuitous path to the completion of my thesis. Professor Campolieti, in particular, gave so much valuable feedback on my papers, in person or in emails, as a function of not only being on my committee, but also of being a co-instructor of the PhD seminar. I benefitted a great deal from his expertise in the field. Moreover, through the PhD seminar in particular, I have learned what it takes to be a serious researcher. Professor Dhuey provided me with the opportunity to collaborate on cutting-edge research and taught me, step by step, how to complete a complex research project. Professor Gunderson not only shows his supervisees how to work professionally in academia but also how to be a better person. His expertise, understanding, and patience added considerably to my doctoral experience. I will never forget the time that we spent in his office reviewing my work sentence by sentence. Professor Gomez, too, has always been inspirational and encouraging. He taught me how to grasp the bigger picture, stay focused, set priorities, and get things done. I am very thankful to him for encouraging me to pursue new research approaches and for providing the infrastructure and resources to accomplish my research work.

Also, thanks are due to Professors Anil Verma and Harry Krashinsky for teaching and guiding me through these years. Professor Verma helped me secure a great internship as part of the team that set Ontario’s minimum wage policy in 2013. Working as a research assistant with Professor Krashinsky gave me the opportunity to expand my econometric skills and explore various research projects.

I greatly appreciate the support that I received from the members of the Centre community. Vicki Skelton and Monica Hypher are more like family than staff and were always there when I needed help. I will really miss our chats during my study breaks. I would like to thank Caitlin MacLeod, who not only helped copy-edit my work, but also taught me how I can improve my writing. iv

Carol Canzano-Hamala has been amazing throughout this whole process, keeping me focused on the finishing line; she was always ready to help. Michelle Petersen-Lee, too, has always been there whenever I needed help. Deb Campbell was such a valuable part of the Centre (I hope she is enjoying her retirement!) and I am grateful for her support.

I thank my officemates at the Centre: Rachel Aleks, Muhammad Umar Boodoo, Bruce Curran, Alycia Damp, Jennifer Harmer, Lydia He, Crystal Huang, Guenther Lomas, Elham Marzi, Joanna Pitek, Amal Radie, Tina Saksida, Alana Steinhauer, Yao Yao, and Qian Zhang for the laughter, the tears, the complaints, and the intellectually stimulating discussions. I am forever grateful to have been part of such an amazing group. Each of us brought our unique character and knowledge to this communal space. We exchanged our understandings of the field, our goals for the future, and our successes and challenges on the road to fulfilling careers.

I am especially indebted to many mentors during my degree program period. Professor Pat Sniderman and Professor Phanikiran Radhakrishnan both helped me to develop my teaching skills. I would also like to give a big shout out to Rupa Banerjee, Danielle Lamb, Lorenzo Frangi, and Dionne Pohler for their friendship and invaluable feedback. I am looking forward to more exciting research collaboration down the road.

I gratefully acknowledge the support from the Statistics Canada Research Data Centre Network (CRDCN), which allowed me access to the data used in this thesis work. In particular, I owe an enormous debt to my friend Carmina Ng at the Toronto regional office, who has been so supportive over the past four years.

Finally, I would like to thank my family for their love, support, and encouragement: my parents, who raised me to be curious, responsible, and resilient, and who supported me in all my pursuits during the ups and downs; my in-laws, who helped me to look after my boys whenever needed; my Uncle Zhang and Aunt Chu for being role models and for having faith in me all along; and my boys, Alex and Brendon, for being understanding of the many family- fun-time-without-mama moments, and for hugging, kissing and looking up to me as their role model. And most of all, I would like to extend my deepest thanks to my loving, cheerful, and patient husband, Qian, whose support has meant so much. Thank you.

Table of Contents

Acknowledgments ...... iv

List of Tables ...... ix

List of Figures ...... xi

Chapter 1 Introduction and Overview of the Thesis ...... 1

1.1 Occupational Licensing in Canada ...... 1

1.2 Theories of Occupational Licensing ...... 3

1.3 Licensing versus Unionization ...... 4

1.4 Labour Market Consequences of Occupational Licensing: What We Know ...... 5

1.5 Focus of the Thesis ...... 7

1.6 References...... 10

Chapter 2 Effects of Occupational Licensing and Unions on Labour Market Earnings in Canada ...... 17

2.1 Introduction...... 17

2.2 Literature on the Impact of Occupational Licensing and Unions on Wages ...... 18

2.3 Occupational Licensing in Canada ...... 20

2.4 Conceptual Framework ...... 23

2.5 Data and Methodology ...... 26

2.5.1 Data ...... 26

2.5.2 Measures ...... 27

2.6 Empirical Approach ...... 29

2.7 Results ...... 31

2.7.1 The overall pay premium ...... 31

2.7.2 Gender differences in wage premiums ...... 33

2.8 Conclusions ...... 34

2.9 References...... 37

Chapter 3 Effects of Occupational Licensing and Unionization on the Distribution of Labour Market Income ...... 51

3.1 Introduction...... 51

3.2 Literature Review ...... 54

3.2.1 The impact of occupational licensing on the distribution of labour income...... 54

3.2.2 The impact of unionization on the distribution of labour income ...... 56

3.3 Theoretical Framework ...... 58

3.4 Data and Empirical Methods ...... 61

3.4.1 Empirical methods ...... 61

3.4.2 Data ...... 63

3.5 Results ...... 67

3.5.1 Differences by gender ...... 70

3.6 Conclusions ...... 72

3.7 References...... 74

Chapter 4 Does Occupational Licensing Restrict the Interprovincial Labour Mobility of Young Workers? ...... 101

4.1 Introduction...... 101

4.2 Literature Review ...... 103

4.3 Theoretical Framework ...... 109

4.4 Research Methods and Data ...... 110

4.4.1 Data ...... 110

4.4.2 Empirical methodology ...... 112

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4.5 Results ...... 114

4.6 Discussion and Conclusion ...... 116

4.7 References...... 121

Chapter 5 Appendix: Occupational Licensing in Canada: A Jurisdictional Review, 1990–2016 ...... 134

5.1 Introduction...... 134

5.2 Occupational Regulation in Canada ...... 135

5.3 Literature Review: Licensing Laws and Right-to-practice ...... 137

5.4 Research Method ...... 143

5.5 Results ...... 144

5.5.1 Regulated occupations are not all licensed ...... 144

5.5.2 Licensing legislation: Exclusive right-to-practice ...... 145

5.5.3 Exclusive right-to-title ...... 147

5.5.4 Compulsory skilled trades ...... 148

5.5.5 Other market closure clauses ...... 150

5.6 Discussion and Conclusion ...... 151

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List of Tables

Table 2.1 Interprovincial Differences in Occupational Licensing (both Licensed Professions and Compulsory Trades and Certification), 2015……………………………………………42

Table 2.2 Descriptive Statistics for Pooled Sample…………………………………………43

Table 2.3 Ordinary Least Squares (OLS) Estimates of Log of Pay Premium: Licensed Occupations………………………………………………………………………………….45

Table 2.4 Fixed Effects (FE) Estimates of Log of Pay Premium: Licensed Occupations………………………………………………………………………………….47

Table 2.5 Ordinary Least Squares (OLS) Estimates of Log of Wage Premium, by Gender: Licensed Occupations………………………………………………………………………49

Table 2.6 Fixed Effects (FE) Estimates of Log of Wage Premium, by Gender: Licensed Occupations…………………………………………………………………………….……50

Table 3.1 Share of Licensed Workers Covered by Collective Agreement/Not Covered by the Collective Agreement, by Year…………………………………………………………….. 82

Table 3.2 Descriptive Statistics (All Paid Workers)…………………………………………83

Table 3.3 Unconditional Quantile Regression Estimates of Log Wage Premiums for Licensed Occupational and Union Status, by Year……………………………………………………90

Table 3.4 Unconditional Quantile Regression Estimates of Log Wage Premiums for Licensing and Union Status; by Gender………………………………………………………………....95

Table 3A.1 Unconditional Quantile Regression Estimates of Log Wages Premiums for Licensed Occupational and Union Status, Interaction Term Included………………………98

Table 3A.2 Unconditional Log Wage Premiums for Licensing and Union Status; Interaction Term Included, by Gender, 1998 vs. 2014………………………………..…………………100

Table 4.1 Descriptive Statistics for Pooled Sample………………………………..………..125

Table 4.2 Estimated Marginal Effects of the Likelihood of Moving to Another Province in a Year: Licensed Occupations……………………………………………………………..…126

Table 4.3 Estimated Marginal Effects of the Likelihood of Moving to Another Province in a Year: Licensed Occupations, by Gender………………………………………………..….127

Table 4A.1 Estimated Marginal Effects of Logit Estimates of Probability of Moving to Another Province in a Year (1993-2010): Licensed Occupations……………………………………128

Table 4A.2 Estimated Marginal Effects of Random-effects Models of Probability of Moving to Another Province in a Year (1993-2010): Licensed Occupations……………………….129

Table 4A.3 Logit and Random-Effects (FE) Estimates of Probability of Moving to Another Province in a Year (1993–2010), by Gender………………………………………………..130

Table 4A.4 Estimated Effects of the Likelihood of Moving to Another Province in a Year (1993–2010): Licensed Occupations………………………………………………..………132

Table 4A.5 Estimated Effects of the Likelihood of Moving to Another Province in a Year (1993-2010): Licensed Occupations, by Gender……………………………………………133

Table 5.1 Interprovincial Differences in Regulated Occupations, 2015…………...... ……...158

Table 5.2 Examples of Interprovincial Differences in Right-to-Practice and Right-to-Title Clauses……………………………………………………………………………………...159

Table 5.3 Interprovincial Differences in Licensed and Certified Professions, as of 2016………………………………………………………………………………………...160

Table 5.4 Years of Individual Trades Designated as Exclusive Right-to-practice by Provincial Apprenticeship Acts…………………………………………………….…..………………161

Table 5A.1 Examples of the Legislative Language Used in the Occupational Regulation Statutes……………………………………………………………………………………...162

List of Figures

Figure 2.1 Comparisons in the Time-Trends of Two Labour Market Institutions: Licensed Occupations and Unionization in Canada, 1998-2014…………………………...……..…...41

Figure 3.1 The Share of Workers Covered by Unions, and Licensed and Certified Workers in Canada, 1998-2014…………………………………………………………..…...…………81

Figure 3.2 Hourly Wage Distribution (All Paid Workers), by Year……………………..……84

Figure 3.3 Mean Difference of Logarithm of Real Hourly Wage over Wage Distribution: Licensed vs. Unlicensed; Union vs. Nonunion, by Year………………………………..…….87

Figure 3.4 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, All Paid Workers, by Year………………..…..…92

Figure 3.5 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, Male Paid Workers, by Year ...……...………...... 96

Figure 3.6 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, Female Paid Workers, by Year...……………...... 97

Chapter 1 Introduction and Overview of the Thesis

When applying for most jobs today, a person needs to submit a resume listing relevant educational attainment, previous work experience, and certifications. The prospective candidate then needs to pass a personnel selection process that involves tests, interviews, or work sampling. All of this information is accepted by both employees and employers as an assurance that the worker is qualified for the job. More and more often, however, this is not sufficient for a person to work in certain occupations where workers need an occupational credential that indicates that they meet the standard to perform the job.

1.1 Occupational Licensing in Canada

Licensed and certified occupations are occupational categorizations used in countries such as the United States, Britain, and Germany. Occupational licensing is the exclusive right to practice; in other words, a worker cannot practise in the occupation unless a license is acquired. Certification is, in fact, a title protection; individuals working in a certified occupation have a title, recognized by the market that is reserved for this group of practitioners. Others can practise in the profession or trade, but they cannot use that title.

In Canada, the federal government calls those occupations that “require a license to practice”1 regulated occupations, which includes both licensed and certified occupations. Licensing (as in exclusive-right-to-practice) and certification (as in exclusive right-to-title), however, should be separated. Licensing and certification have distinct impacts on occupational labour markets, although both lead to market segregation. Some scholars argue that certification is economically equivalent, if not superior, to licensing (e.g., Brockman, 1998; Kleiner, 2013; Wolfson, Trebilcock, & Tuohy, 1980). Certification directly resolves the issue of asymmetric

1 Based on information from the Government of Canada’s “Working in Canada Tool.” Retrieved from: http://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#R 1

information in the services or products market; is relatively flexible, as it preserves the right to free entry into the occupational market; and could prevent market concentration. However, certification may not effectively control the quality of services and products and, more importantly, certification has no authority to penalize or discipline wrongful activities. 2 Licensing, by contrast, allows the professional regulatory bodies to control the entry to the occupation,3 affect both the supply and demand sides of the product or services, and hence, capture the economic rent. Scholars express concern that, with the rising share of workers in the licensed occupations, the costs of licensing outweigh the benefits.

The decentralized nature of occupational regulation has been drawing growing attention in the political debate, mostly in the United States, but also more recently in Canada (Sweetman, McDonald, & Hawthorne, 2015). Although the licensing power falls mostly in the hand of provincial governments,4 in Canada, the federal government has taken action to address the downside of the occupational regulation by limiting the inconsistency of licensing requirements between provinces, developing alternative recognition procedures to lower the entry barriers for foreign-trained licensed professionals, providing more government oversight on licensing bodies, and becoming more critical of attempts to grant licensing power to unlicensed occupations. However, most of these political implementations lacked sufficient

2 Kleiner (2013) discusses three benefits of certification over licensing: (1) it reduces market intervention by reducing restrictions on competition; (2) it provides clearer signals to address information asymmetry; and (3) it reduces the unnecessary and often excessive political activities of professional associations that wish to gain licensing status. In addition, Milner (1979) suggests that statutory-title protection is the most efficient legal mechanism, compared to copyright, registered trademarks, and corporation status. Brockman (1998) argues that certification could be potentially “functionally equivalent to licensure” (p. 593) if consumers, including governments and corporations, perceive that the task can only be completed by certified practitioners. 3 Regulatory bodies, professional associations, and trade unions are interrelated organizations, although each serves different functions in the occupational regulation system. The majority of regulatory licensing agencies are professional associations. However, not all professional associations are granted the right by the government to be self-regulating. Trade unions can discipline workers for workplace-related issues and hold responsibilities to provide training. The distinction between these various institutions is conceptual rather than practical, and any organization may play more than one role. For example, the teachers’ union in Ontario, the Ontario Teachers’ Federation, used to serve all three functions. In 1996, the provincial government established the College of Teachers, which separated the union role from the regulatory college role. 4 Three occupations are reported to be regulated at the federal level: airplane pilots, air traffic controllers, and immigration consultants. 2

empirical evaluation. A systematic examination of the economic impacts of occupational licensing is necessary to support more evidence-based decision making in the Canadian context.

1.2 Theories of Occupational Licensing

Two theories could potentially explain the emergence of occupational licensing. In the first theory, consumer demand for protection and the greater public good are the main arguments for regulating occupations. Consumers are not able to fully protect themselves in the professional-services market. This is because they cannot evaluate the quality of these services, due to the presence of asymmetric information (Cox & Foster, 1990; Shaked & Sutton, 1981); they do not know how to protect themselves from the low-quality services provided by incompetent and/or unethical professionals; and they are not able to seek remedies after suffering damage from unqualified professional services. However, public-interest-protection arguments cannot always justify the decision to limit the scope of practices to a certain group of professionals. The second perspective, rent capture theory, asserts that licensing can provide the legitimacy and enforceability needed to close the market to unqualified professionals. This leads to higher economic welfare for licensed workers,5 but distorts the market by restricting entry to specific occupations, resulting in sub-optional human capital allocation and causing more harm than benefit. Both the public interest theory and the rent capture theory have been addressed and discussed over the past half-century. Combining both theories, licensure is a strong labour market institution that enjoys both jurisdictional protections and monopolistic power (e.g., Begun, Crowe, & Feldman, 1981; Carpenter, 2012; Kleiner, 2000; Moore, 1961; Paul, 1984; Wheelan, 1998).

5 Freidson (1986) argues that professional associations are powerful entities that construct social closure by “carving out a labor-market shelter” in order to protect their members (p. 59). 3

1.3 Licensing versus Unionization

Occupational licensing is an established labour market institution that many researchers have argued should be considered equally as important as unionization (Friedman & Friedman, 1980; Gittleman & Kleiner, 2016; Kleiner 2006; Zhou, 1993). In particular, the limited entry of licensed occupations, protected by the restrictive scopes of practices, are equivalent to the market closure set by the craft union, through which only members of the union can be employed to do certain tasks. Milton Friedman stated that the American Medical Association is “perhaps the strongest trade union in the United States” (Friedman, 1962, p. 150). Both labour market institutions have strong control over the supply of labour, resulting in improved welfare conditions of the members.

In the discussion of occupational licensing in the United States, the growing share of licensed workers is coupled with the sharp decline of union density in the past three decades (e.g., Gittleman & Kleiner, 2016; Kleiner, 2006; Kleiner & Krueger, 2013). There are, however, a few differences between unionization and occupational licensing. First, the two institutions seek divergent venues to protect members’ economic welfare. Zhou (1993) argues that “unskilled workers protected their interests by forming labor unions to deal directly with capitalists, while professional practitioners pursued their interests through state legislation” (p. 537). Secondly, unions increase members’ wages, both directly through collective bargaining for higher wages and fringe benefits, and indirectly by restricting supply in the case of craft unions. Regulatory bodies tend to increase the pay in their profession or trade through occupational licensing. In addition to decreased supply caused by the licensing barriers, another explanation is that occupational licensing increases demand for their profession or trade: only the licensed practitioners have the expertise to indicate how much of their service should be needed. The customers who rely on experts to guide their purchasing decisions will then require more services from the licensed practitioners. Thirdly, in his seminal book on occupational licensing, Kleiner (2006) argues that occupational licensing is a “more secure job classification” (p. 12) than unionization because unions are easily decertified in the United States, while occupations are rarely deregulated once the licensure laws are put in place. Only eight cases in which the state legislators revoked the licensing privilege have been observed in

the United States between the 1970s and the early 2000s (Thornton & Timmons, 2015). Since 2011, there have been 18 proposals in nine states to delicense certain occupations, the majority of which failed to move forward (Thornton, Timmons, & DeAntonio, 2017). By contrast, according to the National Labor Relations Board, 172 union-decertification elections were held in the fiscal year of 2016, of which only 39% were won by the union (National Labor Relations Board, 2016).6

Barbash (1987) argues that there is a “law of equilibrium” in labour markets such that “[i]f a union is not around to assert an institutional interest in equity… the state, management or an informal work society will try to fill the vacuum created by the union’s absence” (p. 172). As a result, when union power in contract negotiations weakens, the status of licensing would be expected to rise to fill the void in the equity function. Licensing associations may even have been seen by workers as a more attainable (if not preferable) form of labour market protection than unionization, especially in the American private sector where unions represent fewer than one in 10 workers. In Canada, the percentage of unionized workers and the percentage of licensed workers have remained steady over the past three decades (approximately 30% versus 11%, respectively). Combining the trends in both the United States and Canada, I believe that these two institutions are closely related. Another reason to look into these two institutions together is that unionization has been studied intensively, while occupational licensing has drawn less research attention in both countries. It would be helpful to use unionization as a bench mark to examine the impact of occupational licensing.

1.4 Labour Market Consequences of Occupational Licensing: What We Know

Occupational licensing, a more restrictive type of occupational regulation than certification, has been viewed by economists as an institutional barrier to a more efficient labour market. Many American studies provide empirical evidence that occupational licensing allows

6 Retrieved from https://www.nlrb.gov/sites/default/files/attachments/basic-page/node- 4626/Total%20Elections%202016.pdf 5

professional occupations to exert monopolistic power by restricting entry, reducing competition, 7 restricting labour mobility, 8 and increasing the price of the services these professions provide9 without necessarily improving the quality of these services.10 Also, rich empirical evidence suggests that occupational licensing increases practitioners’ earnings,11 decreases the quantity of services demanded,12 and results in welfare loss.13

In Canada, there has been no comparable examination of the labour market consequences of occupational licensing simply because there has been no original dataset of occupations’ licensing status. A few studies look into the employment or wage impacts of regulated occupations (Banerjee & Pham, 2014; Chen & Fougère, 2011; Girard & Smith, 2013; Gomez, Gunderson, Huang, & Zhang, 2015). In existing comparative studies of occupational licensing, scholars in other countries have often treated regulated occupations and licensed occupations as equivalent terms (e.g., Forth, Bryson, Humphris, Koumenta, & Kleiner, 2011). Although the estimated impacts of regulated occupations on wages and employment in Canada are similar to the impacts of licensing in countries such as the United States and the United Kingdom, it is necessary to go beyond similarity and ensure true equivalency when comparing terms across jurisdictions.

7 See, for example, Law and Kim (2005), Kugler and Sauer (2005), and Maurizi (1974). 8 See, for example, Federman, Harrington, and Krynski (2006), Kleiner, Gay, and Greene (1982), Kugler and Sauer (2005), Pashigian (1979), Pashigian (1980), and Peterson, Pandya, and Leblang (2014). 9 See, for example, Adams, Jackson, and Ekelund Jr. (2002), Cox and Foster (1990), Haas-Wilson (1986), Kleiner and Kudrle (2000), Kleiner, Marier, Park, and Wing (2016), Rottenberg (1980), Shepard (1978), and Wing and Marier (2014). 10 See, for example, Angrist and Guryan (2004), Carpenter (2012), Carroll and Gaston (1983), Haas-Wilson (1986), Kleiner (2006), Kleiner(2014), Kleiner and Kudrle (2000), Kleiner et al. (2016), Kleiner and Todd (2007), Kugler and Sauer (2005), Law and Kim (2005), Paul (1984), and Shapiro (1986). 11 See, for example, Gittleman and Kleiner (2016), Kleiner (2000), Kleiner (2006), Kleiner and Krueger (2013), Kleiner and Park (2010), Kleiner and Todd (2007), Kugler and Sauer (2005), Moore, Pearce, and Wilson (1981), Muzondo and Pazderka (1980), Pfeffer (1974), Schaefer and Zimmer (1995), Schaefer and Zimmer (2011), Timmons and Thornton (2010), Wheelan (1998), and White (1975). 12 See, for example, Adams et al. (2002). 13 See, for example, Adams et al. (2002) and Feldman and Begun (1985). 6

1.5 Focus of the Thesis

In my dissertation, I examine the impacts of occupational licensing on workers’ earnings, the distribution of labour market income, and young workers’ labour mobility decisions in Canada. To do this, I constructed a dataset of occupational licensing statuses nation-wide based on the exclusive-right-to-practice clause passed in provincial statutes and regulations. In addition to generating a better understanding of the relationship between occupational licensing and labour market outcomes, I also compare the Canadian evidence to the results found using the U.S. data.

One of the central focuses of licensing research is whether or not licensing will increase the economic welfare of licensed professionals. In Chapter Two, I study the wage impact of occupational licensing in Canada. Kleiner and Krueger (2013) report that occupational licensing is associated with a 15% wage premium in the U.S., similar to the union wage premium. My Ordinary Least Square estimates suggest that the estimated licensing wage premium in Canada is 12.0%, slightly higher than the union wage premium, which I estimate as 9.0%. The longitudinal evidence indicates that individual unobserved characteristics have substantial influence on individual wage premiums; the fixed-effects estimates drop to 2.6% for occupational licensing and 4.0% for unionization. The Canadian evidence supports the rent capture theory that licensed workers will earn more than unlicensed workers, controlling for other individual and job characteristics.

In the third chapter, I investigate whether or not occupational licensing premiums are the same for both high- and low-wage earners in the Canadian labour market. Unconditional Quantile Regression estimates indicate that licensing pay premiums prevail throughout the wage distribution but they are smaller at the low end of the pay distribution and larger at the higher end of the pay distribution, compared to unlicensed workers. Therefore, unlike unionization, which lowers inequality by providing higher wage-premiums at the lower bottom of the wage distribution while having lower or negative wage-premiums for the top of the wage distribution, occupational licensing increases inequality through the occurrence of varying wage premiums across the wage distribution. The relationship between occupational licensing

and wage premiums over the pay distribution is a significant finding since the overall pattern of licensing wage premiums fosters wage inequality. Occupational licensing, as a labour market institution, could potentially contribute to the rising labour earning inequality in addition to other mechanisms discussed intensively in the scholarly works.14

Another potential effect of occupational licensing is to restrict labour mobility, not only by imposing barriers to practice a certain occupation but also by limiting workers’ ability to move across jurisdictions and, in particular, to migrate internally. This proposition is supported by earlier U.S.-based empirical analyses. Earlier Canadian studies also report that licensed workers are less mobile than unlicensed workers. In the fourth chapter, I find that individuals’ licensing status is not correlated with the likelihood of moving across provincial boundaries for young Canadians aged 21˗34 using longitudinal microdata. By contrast, unionization significantly decreases the likelihood of internal migration for young Canadian workers.

In addition to studying the labour market impacts of occupational licensing, another focus in my dissertation is the gender differences of these impacts. As the theory predicts, the impacts of unionization are gender-neutral for wage gains. Occupational licensing, however, seems to favour female workers. The licensing wage premiums are higher for female workers than male workers, not only on average but also over the wage distribution. In terms of mobility, occupational licensing does not differentiate workers by gender; licensing status is not associated with the move-stay decision. However, the estimated likelihood of staying in the same province is significantly higher for male union members than female union members. The influence of gender on the impacts of occupational licensing has yet to be thoroughly

14 Scholars have identified various institutional factors as the potential causes of growing earning inequality (e.g. Lemieux, 2008; Fortin, Green, Lemieux, Milligan, & Riddell,2012), including government interventions such as income redistribution policies and loosening market regulation; and changes in labour-market institutions, such as the decline in the real value of minimum wage (e.g. Autor, Manning, & Smith, 2010; Bárány, 2016; Fortin & Lemieux, 2015; Western & Rosenfeld, 2011) and declining unionization (e.g. DiNardo, Fortin, & Lemieux , 1996; Lee, 1999). Others attribute rising earning inequality to the skills-biased technology change (e.g. Autor, Katz, & Krueger, 1998; Autor, Katz, & Kearney, 2008; Brown, 1999; Card & DiNardo, 2002; Galor & Moav, 2000). More recently, studies show that organizational practices, such as the growing use of contingent and precarious workers, and of performance pay, are also seen as factors that contribute to rising earnings inequality in the labour market(e.g. Bidwell, Briscoe, Fernandez-Mateo, & Sterling, 2013; Cobb, 2016; Riaz, 2015). 8

examined in the licensing literature. However, more attention is due given the significant increase in female labour force participation and the fact that the share of female licensed workers has grown significantly higher relative to their male counterparts. Whether or not occupational licensing, like unionization, can narrow the labour market pay gap between male and female requires more investigation.

The main obstacle to Canadian occupational licensing studies is the data limitation, since there has been no population survey asking directly about respondents’ licensing status. I begin unfolding the impacts of occupational licensing using the indicator I constructed and match it with the population survey. To address this data limitation in the future, we need to start asking about individual licensing status in our population survey, much as we do with union status, or find more accurate measures to quantify licensing status directly, such as leveraging more administrative data. More program evaluations and occupation-specific research down the road will help us better understand the causal relationship between occupational licensing and workers’ labour market outcomes.

We are marching into a new era where precarious and on-demand work is becoming increasingly common. The employment relationship has shifted into formats that are less standard, more flexible, and more diverse, but still highly specific to individual occupations. These changes in the market and in society have challenged the traditional shape and boundaries of labor markets and profoundly affected how employees and employers perform, organize, and experience work. In the face of such changes, evidence-based decision making is critical. My studies focusing on the implications of licensing on earnings, distribution of labour market income, and internal labour mobility will improve our understanding of the issues and how they are shaping employment in Canada.

1.6 References Adams III, A. F., Jackson, J. D., & Ekelund Jr, R. B. (2002). Occupational licensing in a "competitive" labor market: the case of cosmetology. Journal of Labor Research, 23(2), 261.

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Autor, D., Katz, L. F., & Kearney, M. S. (2008). Trends in US wage inequality: Revising the revisionists. Review of Economics and Statistics, 90(2), 300–323.

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Chapter 2 Effects of Occupational Licensing and Unions on Labour Market Earnings in Canada

2.1 Introduction

Both unions and occupational licensing are key labour market institutions that affect substantial portions of the workforce. They can affect members’ pay, the likelihood of accessing benefits, and even workers’ overall well-being. In the United States, the proportion of the workforce that is unionized fell steadily from about 28% in the 1960s to about 12% by 2008. Over that same period, the proportion of the workforce in a licensed occupation rose from about 10% in the 1960s to about 29% in 2008 (Kleiner & Krueger, 2013). This decline in unionization of approximately the same magnitude as the increase in occupational licensing has led some to speculate that both unionization and licensing may be substitute forms of labour market regulation, with the rise in licensing substituting for the decline in unionization (Gittleman & Kleiner, 2016; Kleiner, 2006; Kleiner & Krueger, 2010). In Canada, on the other hand, both licensing and unionization coverage have remained relatively steady over the past two decades, which seems to support the potential substitution effect between occupational licensing and unionization.

Both licensing and unionization generate positive wage premiums for members. In the U.S. context, both unions and licensing tend to increase members’ pay by roughly the same amount—around 15%—although there is considerable variation around those estimates (Kleiner & Krueger, 2010, p. 676; 2013, p. 188). Despite efforts to link these two labour market institutions, there is little evidence to illustrate how these two mechanisms jointly impact workers’ compensation packages. The purpose of this paper is to partially fill that gap. In addition, I intend, in this paper, to investigate further whether or not licensing wage premiums differ between males and females.

Based on the Ordinary Least Squares (OLS) estimates of cross-sectional data similar to the estimates found for the United States, the estimated pay premiums for occupational licensing 17

and unionization are approximately 12.0% and 9.0%, respectively. However, the fixed-effects estimates from the longitudinal data are less than half of those amounts for both licensing and unionization. This suggests the importance of unobservable factors for both licensing and unionization. The occupational-licensing pay premium is much higher for females than for males in both the OLS and fixed-effects estimates. The union pay premium, however, is slightly higher for females than for males in the OLS models, but similar in the fixed-effects models.

The remainder of the paper is organized as follows. Section 2.2 briefly reviews the recent literature on the impact of occupational licensing and unionization on wages and fringe benefits. Sections 2.3 and 2.4 follow with a general description of the current developments in Canadian occupational licensing and a brief discussion on the conceptual framework. Section 2.5 describes the empirical methodology and data, while Section 2.6 presents the empirical results, and the last section concludes.

2.2 Literature on the Impact of Occupational Licensing and Unions on Wages

Earlier research on wage premiums generated through occupational licensing have mostly focused on specific occupations (e.g., barbering, accounting, engineering, medicine, etc.) in the United States and showed minimal increases in wages after licensing restrictions were imposed (Law & Kim, 2005; Maurizi, 1974; Pfeffer, 1974; Rottenberg, 1962). However, more recent studies found significant positive licensing wage premiums, ranging from 7% to 22% for some specific occupations: 7% for radiological technology (Timmons & Thornton, 2008); 11% to 22% for barbering (Timmons & Thornton, 2010); 16% for massage therapy (Thornton & Timmons, 2013); 12% for dentistry, which has more restrictive licensing (Kleiner & Kudrle, 2000); and 14% for nurse practitioners, with the relaxation of occupational restrictions in competing professions, such as restrictions against independently providing prescriptions (Kleiner, Marier, Park, & Wing, 2016), and the expansion in scope of practices for this field. Later U.S. studies estimated the conditional average occupational-licensing wage premiums across all licensed occupations at approximately 15%, controlling for other wage-determining

factors, similar to union wage premiums. Specifically, they are 15% in Kleiner and Krueger (2010) and 18% in Kleiner and Krueger (2013).15

The above studies are based on the United States, and there remains a research gap in Canada concerning the wage effects of occupational licensing. Muzondo and Pazderka (1980) did not attempt to estimate the wage effect of occupational licensing. Rather, they used 1971 census data for 20 professions across 10 Canadian provinces (200 observations) to estimate the effect of licensing restrictions on competition. They found that those professions that restrict advertising earn 32.8% more than those who work in professions that do not restrict advertising, and that practitioners of professions that restrict fee competition earn 10.4% more than those employed in professions that do not restrict such competition.

A number of recent Canadian studies focused on barriers that skilled immigrants face in having their foreign credentials recognized and how this has inhibited their assimilation into the labour market (e.g., Augustine, 2015; Girard & Bauder, 2007; Jantzen, 2015; Owusu & Sweetman, 2015; Sweetman, McDonald, & Hawthorne, 2015). Using regulated occupations and licensed occupations interchangeably, Girard and Smith (2013), Banerjee and Phan (2014), and Gomez, Gunderson, Huang, & Zhang (2015) found that recent immigrants to Canada are significantly less likely to work in a regulated occupation than are native-born Canadians. However, immigrants who work in regulated occupations benefit more in wage gains than do native-born Canadians working in the same occupations (Gomez et al., 2015).

Positive union wage premiums have been well documented in empirical research. The cross- sectional estimates range from 14% to 17% in the United States (e.g., Blackburn, 2008; Budd & Na, 2000; Gittleman & Pierce, 2007; Hirsch & Schumacher, 2004), while the range of longitudinal estimates is from 7% to 11% (e.g., Card, 1996; Gabriel & Schmitz, 2014). In Canada, the union wage impact is generally fairly similar to the 15% premium that prevails in

15 Gittleman and Kleiner (2016) reported smaller licensing wage premiums, ranging from 7.4% to 11.8%, using longitudinal data for the period 1979 to 2010. These premiums are much smaller than previous estimates. The differences can be explained by the fact that the earlier Kleiner and Krueger(2010) estimates were based on data collected for the period from 2006 to 2008. 19

the United States, although there is considerable variation around those amounts (Benjamin, Gunderson, Lemieux, & Riddell, 2017, p. 452), and some recent estimates suggest that the premium may have fallen closer to 8–10% (e.g., Kuhn & Sweetman, 1998; Fang & Verma, 2002; Gunderson, Hyatt, & Riddell, 2000; Renaud, 1998).16

More recent U.S. studies used regression discontinuity (RD) to estimate the causal impact of unionization on workers’ wages by comparing workers’ wages before and after union certification in closely contested elections. DiNardo and Lee (2004) find the wages did not significantly change for unions that barely won certification, compared to those that barely lost. Similar results are found in Frandsen (2010) for workers who remained in the firms after certification. The authors indicate, however, that this negligible effect, based on Regression Discontinuity (RD) methodology, could simply reflect the limited short-run bargaining power of those that had just won certification; and that the impact could be larger in the long run, and for those that won by a large margin. In fact, in a more recent study that is based on an event- study methodology, Lee and Mas (2012) calculated impacts that implied union wage effects of about 10%, assuming firms’ future earnings were fully transferred to workers in the form of wages. The authors explained that the negligible effects found in DiNardo and Lee (2004) and in Frandsen (2010) could simply be the results of the weak bargaining power of new unions.

2.3 Occupational Licensing in Canada

A licensed occupation is defined as an occupation that “requires a government-issued license to do [the] job” (Kleiner & Krueger, 2013, p. 181). Based on data from a private survey in 2008, 29% of American workers reported that their job required a professional or government- issued license (Gittleman & Kleiner, 2016). In Canada, the most comparable concept of occupational licensing would be the “regulated occupations,” which are defined as

16 Kuhn and Sweetman (1998) report 9-14% union wage premiums estimated from the displaced workers' wage loss. Cleveland, Gunderson, and Hyatt (2003) also find 15% union wage premiums for Canadian child care workers who are in low-wage service sector. 20

“occupations that set their own standards and that require workers to have a license to practice.”17

If we take a closer look at the legislative language, we can see that the Canadian definition of regulated occupations includes both licensing and certification. Hence, studying the regulated occupations as a whole will not reveal the true impacts of occupational licensing on wages and other labour market outcomes. I conducted a jurisdictional review, wherein I identified the licensed occupations in all 10 Canadian provinces by referencing the specific legislative language of exclusive-right-to-practice, and constructed the occupational licensing codes. Hereafter, occupational licensing will be used to refer to those occupations that require a license to practise.18

Occupational licensing is the most restrictive professional regulation, since it confers the exclusive right-to-practice, whereby only those with the license can practise in the profession (e.g., surgeons, lawyers). This is more stringent than occupational certification, which provides only the exclusive right-to-title, whereby only those with the certification can use the title, whereas others can practise the profession (e.g., “certified human resources professional”) (Kleiner, 2006, p. 18).

Panel A of Figure 2.1 shows the historical trend of the proportion of unionized workers and workers within a licensed occupation in the Canadian labour force between 1998 and 2014.19 By 2014, approximately 11% of paid workers in Canada, both salaried and self-employed, were identified as working in an occupation with a license requirement,20 and 30% of workers were reported to be covered by a collective agreement (i.e., hereafter referred to as unionized, even though a small number who are covered by a collective agreement may not be members

17 https://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#R 18 See more detailed discussion in the Appendix chapter. 19 The graph was derived by the author using the weighted population sample of the Canadian Labour Force Survey, 1998–2014. 20 Occupational licensing can be granted by municipal, state, and federal governments in the United States, and hence, the coverage of occupational licensing is much larger than in Canada (29% and 11%, respectively). 21

of a union). In Panel B, workers in the private sector mimic the U.S. pattern with a moderate decline in union density and a sizeable increase in the share of licensed workers. The proportional changes for both licensing and unionization are quite similar. ------Insert Figure 2.1 about here ------

Currently, 106 unique occupations are licensed in at least one Canadian province (see Table 2.1). There are interprovincial variations in the level of regulation: Quebec has the highest number of licensed occupations, 98, twice that of Newfoundland and Labrador, which has only 47. ------Insert Table 2.1 about here ------

There are two types of licensed occupations in Canada—licensed professions and compulsory skilled trades. Licensed professions (e.g., surgery, law) “require several years of university or college education, practical experience under the supervision of a licensed worker in the chosen profession, and the successful completion of a licensure examination.”21 Regulatory bodies operate at the provincial level in Canada, where professional associations, authorized by their provincial governments through legislative acts and regulations, exert control over their occupations by defining the scope of practices, establishing entry standards, and most important, defending members’ exclusive rights to practise (e.g., Balthazard, 2014). Licensed skilled trades are also called compulsory apprenticeable trades, which are skilled jobs that require those in the trade to have manual skills and special training and also to be under the supervision of a licensed journeyperson (e.g., electricians, mobile crane operators). The regulatory bodies for the skilled trades are provincial apprenticeship and industry training (AIT) authorities, which assess an individual’s credentials, training, and experience before granting a license. The exclusive-right-to-practice of compulsory skilled trades is established

21 Based on information from the Government of Canada’s “Working in Canada Tool.” Retrieved from: http://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#R 22

through the restricted employment requirement in the provincial apprentice training and development statues. How AIT approaches occupational closure varies significantly from province to province. Quebec has the highest number of compulsory trades, 29, followed by Alberta with 21 compulsory trades. British Columbia established licensed trades in 1997, but completely abolished the compulsory trade system in 2003.22 Both licensed professions and compulsory skilled trades exhibit high levels of market closure despite the different characteristics of licensing practices.

2.4 Conceptual Framework

Unionization and occupational licensing have long been seen as closely related labour market institutions. Zhou finds that “[u]nskilled workers protected their interests by forming labor unions to deal directly with capitalists, while professional practitioners pursued their interests through state legislation” (Zhou, 1993, p. 537). Rottenberg (1962) documents the political partnership between the trade unions and state barber associations in the earlier 1960s in the United States.

Unions increase members’ wages, both directly through collective bargaining for higher wages and fringe benefits, and indirectly by restricting supply, as in the case of craft unions. There is a potential “spillover” effect as workers are pushed into nonunionized sectors from reduced employment in the union sector, which in turn produces downward pressure on nonunion wages. Second, the rising wage gap leads to a change in demand when firms substitute nonunionized workers for unionized workers; the wage in the nonunionized sectors may increase up to the point where the wage relatives between unionized and nonunionized sectors are restored. Nonetheless, unionization yields positive wage premiums for unionized members.

Regulatory bodies tend to increase the pay in their profession or trade through occupational licensing, which restricts supply and increases demand for their profession or trade. Regulatory bodies restrict supply through various means: requiring long training and education periods;

22 For a more detailed discussion, see the Appendix chapter. 23

restricting slots in their training and education programs; regulating exam pass rates; not recognizing foreign credentials or even the credentials of other jurisdictions within the country; regulating advertising and fees; and requiring lengthy hands-on experience supervised by a licensed professional. In the case of apprenticeship trades, the ratio of apprentices to journeypersons who supervise the apprentices can be fixed in such a fashion as to make it difficult for apprentices to have access to sufficient journeypersons to obtain the required experience.

The licensed occupations can also artificially increase demand for their services. The power to regulate the profession is often given to the self-governing licensing body because it is difficult for consumers to judge the quality of the service and health as well as the safety issues that may be involved. In such circumstances, consumers may have to rely on the licensed service provider to indicate how much of the service is necessary, and they may demand more of the service because of its perceived higher quality (e.g., Kleiner, 2006; Kleiner & Krueger, 2013). In situations where the licensing process can both restrict supply and artificially increase demand, the potential for excessive pay is obvious.

Finally, professional associations usually provide ongoing training and skills—upgrading that could improve members’ occupation-specific human capital (e.g., Moore, Pearce, & Wilson, 1981), which, in turn, leads to increased wages. In sum, all theories of occupational licensing suggest that there will be a positive licensing wage premium.

From an economic standpoint, the direction of any substitution between unions and licensing could go either way, or it could work in both directions. As union density has declined in the United States, workers may have sought occupational licensing as a way to protect their wages. Conversely, as occupational licensing has become more prominent, individuals may have turned away from unions to protect their interests. Obviously, it is possible for both forces to be at work at the same time. In many cases, professional practitioners will leverage both

mechanisms to improve their own economic well-being.23

Industrial relations theory also suggests that the substitution effect may be prominent, such that “there is an equity function to be represented in the workplace and that unions, employers and governments all compete to represent this function” (Lipset & Meltz, 2004). The “law of equilibrium” theory can also be used to explain the concurrence of an increase in licensing coverage and a decrease in unionization in the U.S. collective bargaining has faced increasing challenges, and it is getting harder to negotiate higher wages for union members. Since unions have not been able to represent the equity function and protect employees’ wages, licensing associations, acting as government agencies, have stepped up to fill this vacuum. Licensing as a supplemental mechanism could help to maintain the balance in workplace representation. As a result, when union power in contract negotiations weakened, the status of licensing significantly grew.

In Canada, the picture is less clear, in part because of limited data on the extent of occupational licensing. As indicated in Figure 2.1, Panel A, both the proportion of the workforce that is unionized (about 30%) and the proportion in a licensed occupation (about 11%) have been fairly constant since the early 1990s.

In Canada, the fact that both union density and occupational licensing have been fairly constant over recent decades is also consistent with the substitution story—in this case, the absence of any substitution. Unions have not declined substantially, so workers may have had less need to look to occupational licensing to protect their interests. Conversely, occupational licensing has not increased substantially, so workers may have had little reason to move away from unions to protect their interests.

This paper will address the following research questions:

23 One could argue that occupational licensing and unionization complement each other, as craft unions tend to have members working in the strictly licensed occupations (e.g., teachers, and nurses). However, the substitution story is more reasonable when observing the evolution of these two institutions over time. 25

1. Do workers who are in licensed occupations receive a significant wage premium compared to workers in similar unlicensed occupations?

2. How do the licensing wage premiums, if they exist, compare to union wage premiums?

3. Are there any significant interaction effects between licensing and unionization?

Despite similar economic experiences over the past a few decades, labour market institutions are different in the United States than in Canada. The differences in union density are mainly attributed to the different labour-relations institutions in both countries. Canada has stronger pro-union labour relations provisions than the United States (e.g., Godard, 2008; Lipset, 1998; Slinn & Hurd, 2011; Taras, 1997; Taras & Ponak, 2001), which clearly leads to higher union coverage and higher union bargaining power in Canada. Therefore, union wage premiums in Canada are expected to be greater than the estimates shown in the studies on the U.S. situation.

Occupational licensing systems, in contrast, are quite similar in both countries, except that occupation licensing is under provincial jurisdiction in Canada,24 whereas in the United States, a particular occupation can be licensed at the municipal, state, or federal level. In Canada, unlike the United States where they can be introduced in various ways, licensing requirements can only be introduced through provincial legislation. Assuming that tighter control of the labour supply will lead to higher wages, estimates of licensed-occupation wage effects are expected to be similar in both countries.

2.5 Data and Methodology

2.5.1 Data

The present study uses six cycles of Statistics Canada’s Survey of Labour and Income Dynamics (SLID). Participants in the SLID were drawn from the same stratified sampling

24 Three occupations are self-reported as licensed at the federal level: airplane pilots, air traffic controllers, and immigration consultants. However, I was only able to find licensing language for pilots. 26

frame as for the monthly Labour Force Survey (LFS). The SLID provides fairly recent longitudinal data on individual labour market activities in Canada. It comprises a national sample of 15,000–17,000 households across 10 Canadian provinces. 25 Each cycle of longitudinal data contains six years of individual labour market activities and earnings, which describe the labour market changes the individuals experienced over time. The survey data also includes demographic information, such as the individual’s gender, age, education, and marital status.

The focus of this study is on the labour activities of the individual’s main job (i.e., the job at which he/she works the most hours).26 The sample is restricted to paid workers (both salaried and self-employed)27 between the ages of 21 and 60 at the first year of the survey who reported labour outcomes for their main job for at least two out of the six years, so as to estimate the changes over time. The final pooled sample contains 249,000 observations, which represent 64,000 individual respondents.28

2.5.2 Measures

The dependent variable is the hourly wage earned in the main job. The key independent variables are union and occupational-licensing status. Union status is a binary variable coded as one, if the individual reported to be a union member or covered by a collective agreement, and zero otherwise. Similar to other Canadian labour-related surveys, such as the Labour Force Survey (LFS), the Workplace and Employee Survey (WES), and the census, there is no

25 This survey followed the responding households for six consecutive years. There are three-year overlaps between two consecutive cycles, by survey design, and thus five cycles of the SLID are covered between 1993 and 2011. 26 Defined by the Canadian Labour Force Survey (LFS), Statistics Canada. Retrieved from: http://www23.statcan.gc.ca/imdb-bmdi/document/3701_D3_T9_V1-eng.pdf 27 Studies looking into the union wage effects generally restrict their sample to male full-time workers. However, in the occupational licensing literature, scholars tend to include the part-time and/or self-employed workers because many licensed workers fall into these categories. 28 Numbers are rounded to the nearest 100 in accordance with the requirements of Statistics Canada’s Research Data Centre. 27

information in the SLID as to whether or not a particular occupation requires a license. Since licensing is under provincial jurisdiction, it was necessary to determine which occupations are licensed in each province along with the year of proclamation. As such, I manually matched the four-digit National Occupational Classification (NOC) of SLID with information from the legislative record of occupational licensing status.29

Table 2.2 presents the descriptive statistics. In the pooled sample, 38.9% (SD=0.488) of the individuals were union members or were covered by collective-bargaining agreements in Canada,30 and 11.4% (SD=0.317) of the participants held occupations that were licensed, and 2.6% (SD=0.158) of workers were in certified professions (e.g., accountants, psychologists in some provinces). The mean hourly wage of paid workers is $CAD20.97 (SD=$CAD11.15). In addition, 9.3% (SD=0.290) reported holding multiple jobs, 3.1% (SD=0.173) identified themselves as self-employed workers, 11.5% (SD=0.319) described themselves as part-timers, 28.0% (SD=0.449) worked in the public sector, and 65.0% (SD=0.467) held a permanent position. ------Insert Table 2.2 about here ------

Although the share of workers in licensed occupations was similar for males and females overall, there was a higher percentage of male workers than female workers in licensed occupations (11.7% and 11.0%, respectively), while a higher percentage of females than males worked in the certified professions (3.0% and 2.1%, respectively). There were also significant differences in wages and access to fringe benefits between male and female members. On

29 The list of licensed occupations is available from the author on request. Similar linking of an individual’s occupation to external information on it being a licensed occupation when the microdata itself does not include such information is presented in Gittleman and Kleiner (2016), and is based on the National Longitudinal Survey of Youth in the United States. Girard and Smith (2013) do a similar linking in their analysis of whether immigrants disproportionately work in regulated occupations, which is based on Canadian census data. 30 The share of unionized workers in the sample is higher than the population average. One possible explanation is the limit of the longitudinal data structure. The slippage rate, which measures the difference between the survey estimates and the census project, ranges from 7% to 13% across six cycles of the SLID. More detailed discussion can be found at http://www.statcan.gc.ca/pub/75f0002m/2008005/section4-eng.htm 28

average, male workers had higher hourly wages than did female workers. The proportion of female workers with college degrees and above is higher than that of their male counterparts (25.6% and 24.3%, respectively). Females are more likely to work as part-timers and in the public sector, and their full-time equivalence working experience is significantly less than that of males.

2.6 Empirical Approach

Fixed-effects estimates are used where identification is obtained from changes in the individual’s occupational-licensing status and union status. For comparison purposes, pooled OLS (ordinary least squares) models are also estimated. Any differences between the fixed- effects models and the OLS estimates are suggestive of the importance of the individual- specific unobservable factors that are controlled for in the fixed-effects regressions, but not in the OLS estimates. The fixed-effects model is:

푙표푔(푤푎푔푒푠) 푖푡 = 훼 + 훾 퐿푖푐푂푐푐푢 푖푡 + 훽 푈푛푖표푛 푖푡 + 휋 푈푛푖표푛 푖푡 ∗ 퐿푖푐푂푐푐푢 푖푡 + ′ Ω 퐶푒푟푡푖푓푖푒푑푃푟표푓푒푠푠푖표푛푠 푖푡+ 훿 푋 푖푡 + 푣 푖 + 휖 푖푡 (1), where v 푖 is an individual-specific residual that is constant for a particular individual and differs among people, and ϵ 푖푡 is the residual that is uncorrelated with X it. The dependent variable 푙표푔(푤푎푔푒푠) is the logarithm of real hourly wages; 퐿푖푐푂푐푐푢 represents whether the worker is employed in a licensed occupation that is granted exclusive right-to-practice by laws, including both licensed professions and compulsory trades; 푈푛푖표푛 indicates whether the worker is a union member or is covered by a collective agreement; 푈푛푖표푛 푖푡 ∗ 퐿푖푐푂푐푐푢 denotes the interaction term of occupational licensing and union status; 퐶푒푟푡푖푓푖푒푑푃푟표푓푒푠푠푖표푛푠 indicates whether the worker is working in a certified profession, which is granted exclusive right-to-title; and X represents a set of conventional personal, human capital and job characteristics that can affect earnings. The variables in X include age,

gender,31 education level, marital status, province of residence, years of work experience, dummy variables for the two-digit North American Industry Classification System (NAICS) industries (manufacturing is the excluded reference), dummy variables for the two-digit NOC occupations (management is the excluded reference benchmark group), dummy variables for multiple-job holders, the permanent-job-holder indicator, the part-time worker indicator, self- employed workers, working in the public sector, and a categorical variable for the number of workers in the workplace.

In the fixed-effects model, the estimated coefficient 훽̂ denotes the average within-group mean estimate of union membership on pay, and the estimated coefficient 훾̂ denotes the within- group estimate of occupational licensing on pay. In other words, the estimated coefficients indicate the average impact of changes in union status and occupational licensing status on wages. The estimated coefficient 휋̂ denotes the within-group mean estimate of the interaction term between occupational licensing and union membership on pay.

In Table 2.3, for the log of real hourly pay, the dependent variable of interest, Model [1] provides estimates of the gross relationship between pay and the two main predictors (licensed occupation status and union status) with only year controls.32 Model [2] adds controls for individual characteristics (i.e., gender, education level, marital status, province of residence, and years of work experience); Model [3] adds controls for job characteristics (multiple-job holding, part-time vs. full-time jobs, self-employed workers, permanent-job holding, public sector job, and categories of industries and plant size); Model [4] adds further categories for occupations. There is an issue as to whether it is appropriate to control for occupations to the extent that they may be endogenous mechanisms through which licensing affects earnings. However, I have only controlled for the aggregate occupational categories when identifying

31 Gender is not included in the fixed effects model because gender is constant over time. A fixed effects model will not incorporate such variables. To counter this issue, I ran the fixed effects models on subsamples by gender. 32 The model also included the interaction between licensing and unionization and certified profession status. 30

the effect of occupational licensing and, hence, the potential selection bias is minimized.33 For this reason, the preferred specification is Model [4], which includes controls for both personal, human-capital and job characteristics, including occupations. In Table 2.4, fixed-effects specifications are the same as in Table 2.3; these help to illustrate the magnitude of the impact of individual unobserved heterogeneity on the estimated wage premiums.34

2.7 Results

2.7.1 The overall pay premium35

As indicated in Table 2.3, the cross-sectional estimates (Model OLS[1]) indicate that the gross pay premium for workers in a licensed occupation, compared to those in an unlicensed occupation, is 30.4%. This decreases to 17.3% after controlling for individual and human capital characteristics (Model OLS[2]), highlighting that licensed workers have greater endowments of the characteristics that positively affect pay. The premium further decreases to 12.4% after further controlling for job characteristics (Model OLS[3]) and decreases to 12.0%, after controlling for additional occupation categories (Model OLS[4]). The preferred estimates are those of Model [4], which controls for personal, human capital and job characteristics. This estimate of 12.0% is very close to the common estimate of around 15% found in most U.S. studies that are based on OLS estimates.

33 See the discussion in Gittleman & Kleiner (2016, p. 15). The estimated coefficients of models [3] and [4] are very similar in the current study. I choose model [4] as the preferred specification, based on consistency between licensing and unionization. 34 The conventional mincer equations include the quadratic terms of age and experience to capture the curvy- linear nature of the impacts on wages. I also tested the models including the squared terms of age and experience. The estimates are significantly negative for the squared variables. However, adding the quadratic terms does not change the estimated results of occupational licensing and unionization. The linear model was selected here because the interpretation of estimated coefficients would be more straightforward than the nonlinear model. 35 For dummy independent variables (like licensed status and union status), the true proportional change in the dependent variable associated with being in the dummy variable category is exp(β) -1, where β is the estimated coefficient. For small values of β that close to zero, the approximation is very close, underestimating the true value by less than 0.02 for values of β < 0.20, which tends to be the case for our main variables of interest in our preferred specifications; hence, we report the β coefficients. 31

------Insert Table 2.3 about here ------

The fixed-effects estimates of the pay premium from licensing, shown in Table 2.4, are substantially lower than the OLS estimates. This suggests that licensed workers also have unobserved characteristics that are associated with higher pay. The preferred estimates that are based on controlling for personal, human capital, and job characteristics, including occupations (Model FE[4]), imply a licensing pay premium of 2.6%, which is less than one-quarter of the 12.0% found in the OLS estimates that did not control for such unobserved characteristics. The fact that licensed-occupation wage premiums significantly decrease after controlling for job characteristics, whether in the OLS or fixed-effects estimates, highlights that the positive wage premiums of occupational licensing substantially represent the premium of having a better job. ------Insert Table 2.4 about here ------

A fairly similar pattern prevails for the union pay premium. Based on the OLS estimates, the gross union-nonunion pay gap is 18.8% before controlling for other determinants of pay (Model OLS[1]). After controlling for the effect of personal and human capital characteristics (Model OLS[2]), the estimate decreases to 16.4%, and after further controlling for job characteristics, the union pay premium drops to 4.6%. The estimated union wage premiums go up to 9.0% after adding additional occupational controls. In the case of union pay premiums, the preferred estimates are those of Model OLS[4] that also controls for both job characteristics and occupations, since these are not likely to be mechanisms through which unions affect pay. The union pay premium of 9.0%, which is based on also controlling for the effect of aggregate occupations, is higher than the 4.6% obtained when such job characteristics are not controlled for, and both are lower than the 15% estimate commonly found in the U.S. literature, but are fairly close to the 8–10% found in the more recent Canadian evidence (Benjamin et al., 2017, p. 452).

The fixed-effects estimates of unionization in Table 2.4 are all smaller than the OLS estimates, although the differences between the OLS and fixed-effects estimates are not as great as the 32

occupational licensing estimates. This suggests that unionized workers also have unobserved characteristics that positively affect pay, such that when they are controlled for in the fixed- effects estimates, the pay premium is somewhat reduced. But the effect of the unobservables on the union pay premium is not as great as their effect on the occupational-licensing premium. The preferred fixed-effects estimates of 4.0% (Model FE[4]), based on also controlling for occupations, is very close to the fixed-effects estimates of 3.9% after controlling for personal, human capital and other job characteristics. Both are smaller than the approximately 9.0% in the preferred OLS estimates, and this is typical of the literature on union impact that is based on longitudinal data, where “[l]ongitudinal or panel data analyses in the United States…generally find that the estimated union-nonunion wage differential is about 10 percent, substantially smaller than cross-sectional estimates” (Benjamin et al., 2017, p. 448). In essence, the preferred OLS cross-section estimates of a union pay premium of 9.0% (Model OLS[4]) and fixed-effects estimates of 4.0% (Model FE[4]) are smaller than the U.S. OLS cross-section estimates of around 15% and fixed-effects estimates, based on longitudinal data, of 10%.

The cross-sectional estimates of the interaction term between unionization and occupational licensing are negative and generally statistically insignificant in both the OLS and fixed-effects estimates. The insignificance of the interaction term (both statistically and quantitatively) suggests that there is no additional effect from being both in a licensed occupation and covered by a collective agreement. As such, my discussion of the effects of licensing and unions focuses on their direct independent effects.

2.7.2 Gender differences in wage premiums

Table 2.5 provides the OLS and fixed-effects estimates of wage premiums by gender for both licensed occupation and union status. Based on Model [4] with all control variables, the OLS estimates of the effect of occupational licensing on pay are much larger for females (17.8%) than for males (9.6%). This is also the case for the fixed-effects estimates, although they are much smaller, at 5.0% for females and near zero for males. The smaller estimates for the fixed- effects models suggest that those in licensed occupations also have unobservable

characteristics that positively affect pay, so that when these are controlled for, the effect of licensing is somewhat reduced.

The union impacts on pay are comparable between male and female workers. Based on Model [4] with human capital and job characteristics controls, the union impact is 9.3% for females and 8.6% for males, when based on the OLS estimates; and 4.0% for females and 4.1% for males, when based on the fixed-effects estimates that better control for the effect of unobservables. Both confirm that the collective-bargaining process does not generally discriminate against female workers.

In addition, the estimated cross-sectional and fixed-effects coefficients of the interaction term are not statistically significant for either female or male workers after controls, which implies no support for the existence of substitution effects between occupational licensing and unionization. In sum, female workers benefit more than male workers from both licensed occupational status and union status. ------Insert Table 2.5 about here ------

2.8 Conclusions

The results for occupational licensing are based on the preferred specification of Model [4], which controls for both personal and human capital characteristics as well as job characteristics, including aggregate occupations. The following generalizations emerge:

(1) The pay premium for occupational licensing of 12.0%, based on pooled, cross-sectional OLS estimates, is reduced substantially to 2.6%, based on the fixed-effects longitudinal estimates that control for the effect of unobservable factors;

(2) The licensing premium is larger for females than for males in both the cross-section OLS estimates (17.8% for females and 9.6% for males) and the fixed-effects estimates (5.0% for females and zero for males);

(3) The union pay premium is 9.0%, based on the cross-sectional OLS estimates and 4.0% based on the longitudinal fixed-effects estimates, similar for females as for males in the fixed- effects models.

(4) The occupational licensing premiums are smaller than, but roughly in line with, the premiums in the United States, which tend to be around 15% for each, based on the cross- sectional data. The cross-sectional estimates of the union wage premium, using Canadian data, are also similar to the U.S. estimates. American studies of the union impact, based on the longitudinal fixed-effects estimates, also find similar impacts in the neighbourhood of 10%.

This study departs from and contributes to the previous literature in the following four ways. First, both occupational licensing and unionization have significant positive impacts on wages, and exert a similar influence on wage premiums. Second, the magnitudes of positive wage premiums of occupational licensing and unionization are smaller in Canada than in the United States (Kleiner & Krueger, 2010; 2013). Third, females gain higher wage premiums in both licensing and unionization compared to males. Last but not least, there is no empirical evidence that supports the notion that unionization and occupational licensing are intertwined and jointly impact workers’ wages.

An additional contribution of this study is the construction of Canadian occupational licensing data. Previous Canadian studies have used information from the Government of Canada’s website, WorkingCanada.org, which not only includes both licensed and certified occupations, but also assumes that occupations were licensed before the beginning of the data (Banerjee & Phan, 2014; Girard & Smith, 2013; Gomez et al., 2015). I constructed the new occupational licensing indicator based on a jurisdictional review of occupational statutes and acts, and recorded the year that occupation licensing laws were proclaimed for each particular occupation across all 10 Canadian provinces. Gittleman and Kleiner (2016) use the same approach; their cross-sectional estimate of occupational licensing is smaller than the estimate reported in this study (7.4% and 12.0% respectively), while their estimate of union wage premiums is higher in the United States than in Canada (18.3% and 9.0%, respectively). The magnitudes of fixed-effects estimates exhibit the same patterns, but the gaps are smaller (2.6%

in Canada and near zero in the United States for occupational licensing premiums, and 4.0% and 12.8% for union wage premiums, respectively).

The SLID data provide a rare opportunity to examine the impact of both occupational licensing and unions, and to do so separately for males and females. Our knowledge in this area, however, would be substantially enhanced by adding questions to the LFS on whether or not the individual is working in a licensed occupation (perhaps through the use of supplementary questionnaires if the response burden is too great to add to the regular survey). This is being done in the Current Population Survey (CPS) in the United States; and it would be useful to do the same in Canada.

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Figure 2.1 Comparisons in the Time-Trends of Two Labour Market Institutions: Licensed Occupations and Unionization: Canada 1998–2014.

Panel A: All Paid Workers

Panel B: Professional Workers in the Private Sector

Note: The figures were derived, by the author, using the weighted population sample of the Canadian Labour Force Survey, 1998–2014. Data is restricted to all paid workers in the survey data.

Table 2.1 Interprovincial Differences in Occupational Licensing (Both Licensed Professions and Compulsory Trades) and Certification, 2016

Province BC AB SK MB ON QC NB NS PEI NFL Panel A Number of licensed occupations (exclusive right-to-practice) 60 92 67 71 80 98 65 69 55 47 Number of licensed professions 54 64 53 56 63 61 54 53 44 40 Number of compulsory trades 0 21 6 9 9 29 6 9 6 0

Panel B Number of certified professions (right-to-title) 13 12 18 12 8 28 8 13 4 3

Note: The information was collected based on the author’s own jurisdictional reviews of legislative documents across 10 Canadian provinces. Some licensed occupations also have right-to-title clauses in their professional acts.36

36 Several occupations fell into neither licensed professions nor compulsory skilled trades such as real estate brokers, funeral directors, private investigators, and security guards. The main features of these occupations include: 1) do not have a self-regulatory association; 2) do not require apprentice training; 3) do not have specific educational requirements; but 4) members are expected to have exclusive-right-to-practice by holding a government-issued license. 42

Table 2.2 Descriptive Statistics for Pooled Sample

Pooled sample Males Females Variables Mean Std Dev Mean Std Dev Mean Std Dev Hourly wages ($) 20.97 [11.15] 23.28 [11.87] 18.53 [9.74] Log of hourly wages 2.89 [0.49] 3.00 [0.47] 2.77 [0.48] I (employment-based benefits) 0.762 [0.426] 0.805 [0.397] 0.718 [0.450] I (pension plan) 0.578 [0.494] 0.609 [0.488] 0.546 [0.498]

Licensed occupations (both 0.114 [0.317] 0.117 [0.321] 0.110 [0.313] licensed professions and compulsory trades) Certified professions 0.026 [0.158] 0.021 [0.144] 0.030 [0.172] Union members 0.389 [0.488] 0.392 [0.488] 0.386 [0.487]

Females 0.485 [0.500] Age (years) 41.7 [10.0] 41.7 [10.0] 41.8 [10.0] Married 0.716 [0.451] 0.735 [0.441] 0.695 [0.461] Education High school and less 0.369 [0.482] 0.385 [0.487] 0.351 [0.477] College 0.353 [0.478] 0.342 [0.474] 0.365 [0.482] University B.A. and above 0.250 [0.433] 0.243 [0.429] 0.256 [0.437]

Experience (years) 17.7 [9.7] 19.4 [10.0] 16.0 [9.1] Multiple-job holder 0.093 [0.290] 0.086 [0.280] 0.100 [0.301] Part-time status 0.115 [0.319] 0.036 [0.187] 0.198 [0.398] Self-employed 0.031 [0.173] 0.036 [0.187] 0.026 [0.158] Permanent job status 0.650 [0.477] 0.654 [0.476] 0.646 [0.478] Public sector job 0.280 [0.449] 0.229 [0.420] 0.334 [0.472]

Plant size Fewer than 20 employees 0.291 [0.454] 0.262 [0.440] 0.322 [0.467]

20–99 employees 0.309 [0.462] 0.308 [0.462] 0.310 [0.462] 100–499 employees 0.232 [0.422] 0.250 [0.433] 0.213 [0.409] More than 500 employees 0.066 [0.248] 0.071 [0.257] 0.061 [0.239] # observations37 249,000 123,000 126,100

37 Missing categories are created for binary and categorical variables (e.g., education, plant sizes, and job characteristics). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement. 44

Table 2.3 Ordinary Least Squares (OLS) Estimates of Log of Pay Premium (1993–2011): Licensed Occupations

Dv: log of hourly wage OLS[1] OLS[2] OLS[3] OLS[4] Licensed professions (licensed 0.304*** 0.173*** 0.124*** 0.120*** professions or compulsory trades) (0.063) (0.046) (0.044) (0.032) Union 0.188*** 0.164*** 0.046** 0.090*** (0.036) (0.027) (0.023) (0.014) Licensed profession*union -0.092* -0.044 0.010 -0.021 (0.055) (0.034) (0.033) (0.025) Certified professions 0.208*** 0.129*** 0.074*** 0.058** (0.050) (0.033) (0.024) (0.024) College degree 0.148*** 0.112*** 0.086*** (0.014) (0.010) (0.008) Bachelor’s degree + 0.422*** 0.362*** 0.277*** (0.025) (0.016) (0.008) Age (years) 0.007*** 0.000 0.000 (0.001) (0.000) (0.000) I(married) 0.095*** 0.078*** 0.069*** (0.010) (0.007) (0.005) I(self-employed) 0.089*** 0.076*** (0.010) (0.009) I(multiple-job holder) -0.083*** -0.074*** (0.009) (0.007) I(permanent job) 0.125*** 0.113*** (0.010) (0.008) I(part-timer) -0.096*** -0.068*** (0.022) (0.018) I(public sector) 0.104*** 0.091*** (0.014) (0.014) Experience (years) 0.009*** 0.008*** (full-time equivalence) (0.001) (0.000) Plant size (21–99) 0.020** 0.020** (0.010) (0.009) Plant size (100–499) 0.086*** 0.081*** (0.012) (0.012) Plant size (500+) 0.136*** 0.127*** (0.012) (0.011) Females -0.236*** -0.175*** -0.172*** (0.021) (0.010) (0.010) Constant Y Y Y Y Year controls Y Y Y Y Industry N N Y Y Occupation N N N Y Province N Y Y Y 45

# observations 249,000 249,000 249,000 249,000 2 R adjusted 0.081 0.288 0.409 0.458 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels.38 Occupational licensing is a step function that is equal to one if the individual works in a licensed occupation (licensed profession or compulsory trades), and zero otherwise.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2011).

38 I also clustered the stand errors at the individual level. The results are identical due to the large sample size. 46

Table 2.4 Fixed Effects (FE) Estimates of Log of Pay Premium (1993–2011): Licensed Occupations

Dv: log of hourly wage FE[1] FE[2] FE[3] FE[4] Licensed professions(licensed 0.048*** 0.045*** 0.030** 0.026** professions or compulsory trades) (0.014) (0.014) (0.012) (0.012) Union 0.062*** 0.061*** 0.039*** 0.040*** (0.007) (0.007) (0.006) (0.006) Licensed profession*union -0.001 0.000 0.001 -0.003 (0.012) (0.011) (0.011) (0.011) Certified professions 0.046*** 0.042*** 0.032** 0.022* (0.015) (0.014) (0.013) (0.013) College degree 0.036*** 0.028*** 0.026** (0.010) (0.010) (0.010) Bachelor’s degree + 0.120*** 0.095*** 0.088*** (0.016) (0.015) (0.014) Age -0.021* -0.071*** -0.034 (0.011) (0.025) (0.026) I(married) 0.032*** 0.027*** 0.027*** (0.005) (0.004) (0.004) I(self-employed) 0.014 0.013 (0.008) (0.008) I(multiple-job holder) -0.028*** -0.027*** (0.004) (0.004) I(permanent job) 0.025*** 0.025*** (0.005) (0.005) I(part-timer) -0.028*** -0.024*** (0.008) (0.007) I(public sector) 0.080*** 0.082*** (0.012) (0.012) Experience (years) 0.003*** 0.003*** (full-time equivalence) (0.001) (0.001) plant size (21-99) 0.004 0.005 (0.004) (0.004) plant size (100-499) 0.009*** 0.010*** (0.003) (0.003) plant size (500+) 0.016*** 0.016*** (0.004) (0.004) Constant Y Y Y Y Year controls Y Y Y Y Industry N N Y Y Occupation N N N Y Regions N Y Y Y # observations 249,000 249,000 249,000 249,000 # individuals 64,000 64,000 64,000 64,000 47

Rho 0.846 0.875 0.952 0.892 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function that is equal to one if the individual works in a licensed occupation, and zero otherwise. Number of observations was rounded to the nearest hundred, according to the Statistics Canada Research Data Centre requirement.] Source: Survey of Labour and Income Dynamics (SLID) (1993-2011).

Table 2.5 Ordinary Least Squares (OLS) Estimates of Log of Wage Premium, by Gender (1993–2011): Licensed Occupations

Pooled OLS Female Male DV: log of hourly wage OLS[1] OLS[3] OLS[4] OLS[1] OLS[3] OLS[4] Licensed (licensed 0.367*** 0.191*** 0.178*** 0.198*** 0.081** 0.096*** professions or (0.074) (0.058) (0.056) (0.072) (0.040) (0.023) compulsory trades) Union 0.243*** 0.058*** 0.093*** 0.119*** 0.037 0.086*** (0.037) (0.020) (0.014) (0.039) (0.029) (0.017) Licensed*union -0.051 -0.004 -0.039 -0.076 -0.011 -0.037 (0.066) (0.048) (0.036) (0.062) (0.039) (0.032) Certified professions 0.232*** 0.100*** 0.082** 0.221*** 0.042* 0.040** (0.059) (0.037) (0.037) (0.040) (0.024) (0.019) Personal and human capital controls N Y Y N Y Y Job characteristics N Y Y N Y Y Industry N Y Y N Y Y Occupation N N Y N N Y Province N Y Y N Y Y # observations 126,100 126,100 126,100 123,000 123,000 123,000 2 R adjusted 0.151 0.429 0.482 0.041 0.343 0.391

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function that is equal to one if the individual works in a licensed occupation (licensed professions or compulsory trades). Each regression includes a constant as well as controls for individual (education level, marital status, and experience) and job characteristics (self-employment status, multiple-job-holder status, permanent employment status, part time employment status, public sector indicator, plant size, and categories of industry and one-digit occupation). The control variables are included corresponding to the specification. Number of observations was rounded to the nearest hundred, according to the Statistics Canada Research Data Centre requirement.] Source: Survey of Labour and Income Dynamics (SLID) (1993-2011).

Table 2.6 Fixed Effects (FE) Estimates of Log of Wage Premium, by Gender (1993-2011): Licensed Occupations

Fixed effect DV: log of hourly Female Male wage FE[1] FE[3] FE[4] FE[1] FE[3] FE[4] Licensed (licensed 0.087*** 0.054** 0.050** 0.025* 0.016 0.012 professions or (0.025) (0.023) (0.023) (0.013) (0.013) (0.012) compulsory trades) Union 0.069*** 0.039*** 0.040*** 0.056*** 0.040*** 0.041*** (0.011) (0.008) (0.008) (0.007) (0.006) (0.006) Licensed*union -0.017 -0.013 -0.018 0.005 0.006 0.004 (0.019) (0.016) (0.015) (0.014) (0.014) (0.014) Certified professions 0.065*** 0.047** 0.041** 0.020 0.015 -0.001 (0.022) (0.020) (0.020) (0.020) (0.021) (0.019) Personal and human capital controls N Y Y N Y Y Job characteristics N Y Y N Y Y Industry N Y Y N Y Y Occupation N N Y N N Y Province N Y Y N Y Y # observations 126,100 126,100 126,100 123,000 123,000 123,000 # individuals 32,500 32,500 32,500 31,500 31,500 31,500 Rho 0.842 0.973 0.940 0.830 0.822 0.817 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function that is equal to one if the individual works in a licensed occupation (licensed professions or compulsory trades). Each regression includes a constant as well as controls for individual (education level, marital status, and experience) and job characteristics (self-employment status, multiple-job-holder status, permanent employment status, part time employment status, public sector indicator, plant size, and categories of industry and one-digit occupation). The control variables are included corresponding to the specification. Number of observations was rounded to the nearest hundred, according to the Statistics Canada Research Data Centre requirement.] Source: Survey of Labour and Income Dynamics (SLID) (1993-2011).

Chapter 3 Effects of Occupational Licensing and Unionization on the Distribution of Labour Market Income

3.1 Introduction

Rising earnings inequality has drawn significant attention in both public policy formation and academic research, especially since the financial crisis of 2008. Scholars have not only documented the rapidly increasing share of wealth in the top income percentile in developed countries,39 but they have also noted the erosion of the middle class40 over the past three decades in countries such as the United States and Canada (e.g., Atkinson & Piketty, 2010; Fortin, Green, Lemieux, Milligan, & Riddell, 2012; Foster & Wolfson, 2010; Jones, 2015; Piketty & Saez, 2014, 2017). Various mechanisms are considered to be the main factors that cause the growth in earnings inequality (Lemieux, 2008; Fortin et al., 2012). These include skill-biased technology change (e.g., Autor, Katz, & Krueger, 1998; Autor, Katz, & Kearney,

39 For example, Burkhauser, Feng, Jenkins, and Larrimore (2012) compare both a population survey and administrative data, and confirm that the share of total income going to the top income earners has increased sharply since the 1960s in the United States. Piketty and Saez (2014) chart long-term income inequality in Europe and the United States and show that the share of total income held by the top 10% of earners has increased in both regions since the 1970s. Their most recent updates (Piketty & Saez, 2017) suggest that the income share of the top decile grew from around 25% in the 1960s to around 35% between 2000 and 2011, but has varied significantly by year since the 2000s. Fortin, Green, Lemieux, Milligan, and Riddell (2012) document the sharp increase in income inequality in Canada. In particular, the share of total income earned by the top 1% closely mirrors the U.S. trend. Veall (2012) also reports the increase in the income share for the Canadian top income earners using longitudinal administrative data. Roine, Vlachos, and Waldenström (2009) analyze the share of income for the top income percentile as a proportion of the total income for 16 countries over the twentieth century, and find in recent decades income inequality has increased in all sixteen countries. Their findings suggest that economic development disproportionately favours top income earners (the 99th percentile), while government spending aids the bottom income earners (the 1st to 89th percentile), hurts the upper middle class (the 90th and 99th percentile), and does not impact the rich (the 99th percentile). 40 For example, Autor (2014) looks at the earning inequality among the bottom 99% and concludes that educational attainment drives the earning inequality. Autor, Katz, and Kearney (2008) report that skill-biased technology change negatively impacts occupations, particularly those clustered in the middle of the wage distribution; hence, the wage distribution becomes more “polarized” because of the disappearance of “middle class wage” jobs. 51

2008; Brown, 1999; Card & DiNardo, 2002; Galor & Moav, 2000); government interventions in cutting social welfare programs and loosening market regulation; and changes in labour market institutions, such as the decline in the real value of minimum wages (e.g., Autor, Manning, & Smith, 2010; Bárány, 2016; Fortin & Lemieux, 2015; Neumark, Schweitzer, & Wascher, 2004; Western & Rosenfeld, 2011) and declining unionization (e.g., DiNardo, Fortin, & Lemieux, 1996; Lee, 1999). Organizational practices, such as performance pay and the growing use of contingent and precarious workers, are also seen as factors that contribute to earnings inequality in the labour market (e.g., Bidwell, Briscoe, Fernandez-Mateo, & Sterling, 2013; Cobb, 2016; Riaz, 2015).

Only a few studies have examined the impact of occupational licensing on heterogeneity in employment earnings. They argue that social segmentation, of which occupational licensing is considered one form, is an important factor that contributes to earnings inequality (Kalleberg, Wallace, & Althauser, 1981; Weeden, 2002). Weeden (2002) identifies five mechanisms that lead to earnings inequality in the United States. Her findings show that licensing and educational credentialing are two means of achieving occupational closure. Consequently, licensing has “a particular strong effect on occupational rewards” because occupational licensing represents both “supply-side monopolization” and “jurisdictional protection” (Weeden, 2002, p. 69). Kalleberg et al. (1981) point out that the worker power that is derived from collective efforts, such as unionization and professional licensing, is one of the most important determinants of income variation.

Many scholars consider occupational licensing and unionization to be comparable labour market institutions. Friedman and Friedman (1980) refer to long-standing professional associations, such as the American Medical Association, as labour unions by nature. More recently, Kleiner (2000) discusses the similarities between professional associations and labour unions, both of which can close labour submarkets by restricting entry. Since the late 2000s, occupational licensing has drawn increasing research interest. Most previous empirical studies focus on wage gains as a consequence of implementing licensing restrictions in the United States; the mean estimate of the impact of licensing and wage premiums is about 15% (Kleiner & Krueger, 2010, 2013; Gittleman & Kleiner, 2016). In Canada, recent studies report

that wage premiums due to occupational licensing range between 7.7% and 13.2% (Gomez, Gunderson, Huang, & Zhang, 2015; Zhang, 2016). These studies support the existence of labour income differentials between licensed and unlicensed workers, which could be seen as contributing to earnings inequality.

Empirical research on the impact of occupational licensing on the wage distribution is rather limited. The few studies include Kleiner (2006) and Kleiner & Krueger (2013). To the best of my knowledge, no such study has been conducted in the Canadian context. I use Canadian data to analyse the impact of licensing on wage distribution, comparing it to the impact of unionization, on which there is more extensive literature.41 In line with the existing literature (e.g., Firpo, Fortin, & Lemieux, 2009), I find that unionization decreases wage inequality by incurring smaller—even negative—wage premiums for higher-wage earners, while raising wages for lower-wage earners. In contrast, occupational licensing slightly increases relative inequality by generating higher wage premiums for higher-paid workers. More importantly, occupational licensing causes the 90/10 ratio between wages at the top of the distribution (the 90th percentile) and the bottom of the distribution (the 10th percentile) to increase, while unionization causes the 90/10 ratio to decrease.

The wage-distribution effects also vary by gender. Unionization compresses the wage distribution for both male and female workers in Canada. The distributional effects of unionization for both male and female workers are similar to the effects on workers overall. The distributional effects of occupational licensing for male workers are also similar to the effects on all workers combined, showing relatively minor positive wage premiums of licensing along the majority of the wage distribution until a moderate gain for the top decile of male wage earners. By contrast, occupational licensing for female workers shows a bell- shaped pattern similar to the distributional effects of unionization, despite the peak wage premiums exhibited at the 70th percentile. For female workers across most of the wage distribution, the higher the earnings, and the greater the licensing wage premiums. This

41 See, for example, Benjamin, Gunderson, Lemieux, and Riddell (2017, p. 448) for a comprehensive summary of the union wage impacts for the United States and Canada. 53

suggests that licensing could be a possible mechanism to narrow the gender wage gap, even though it does not reduce earnings inequality.

3.2 Literature Review

3.2.1 The impact of occupational licensing on the distribution of labour income

There have been many theoretical discussions about how occupational licensing, as a labour market institution, could potentially impact the wage distribution (e.g., Bol & Weeden, 2014; McLaughlin & Stanley, 2016; Sørensen, 1996; Weeden & Grusky, 2014; Weeden, Kim, Di Carlo, & Grusky, 2007).42 However, there is limited empirical evidence on this topic, and the research mainly investigates the occupational labour market in the United States. (e.g., Kleiner, 2006; Kleiner & Krueger, 2013). Sørensen (1996) suggests that the rent generated by social closure, such as occupational licensing or unionization, limits entry and leads to inequality. Professional associations, as rent-generating organizations, will actively defend members’ interests by preventing unauthorized workers from becoming employed in areas that require skills that are restricted to that profession. Weeden and Grusky (2014) point out that such institutional barriers could potentially contribute to earnings inequality.43 They also argue that labour market institutions, such as unionization and licensing, squeeze out lower-wage earners and that over time more rent is created for the higher-wage earners. In particular, occupational licensing is “an important rent-generating institution” and, hence, “an important source of rising income inequality” (Weeden & Grusky, 2014, p. 482). Weeden et al. (2007) report that the increase in wage inequality between occupational groups in the United States is particularly evident for nonmanual workers between 1973 and 2005. The authors suggest that one potential

42 Many earlier theoretical and empirical studies confirm the positive wage premiums produced by occupational licensing for licensed workers, but they do not explicitly discuss whether or not the occupational licensing wage premiums are the same along the distribution of wages (see Kleiner, 2006, for a detailed discussion; also see the literature review in Chapter 1). 43 The authors note that the forms of labour market rent can include minimum wage, union wage premiums, interindustry wage differentials, and licensing and other occupational-closure wage premiums. 54

driving factor is occupational licensing, which enables skilled employees to control the supply of and demand for jobs in their licensed profession.44 Bol and Weeden (2014) find that wage inequality in Germany and the United Kingdom is positively associated with occupational closure among various labour market institutions, including educational credentialing, unionization, apprenticeship, and the occupational licensing, noting that this impact is stronger in the United Kingdom than in Germany. McLaughlin and Stanley (2016) report that business regulations that tighten entry are associated with the growing share of income going to the top income earners and with an increase in the Gini coefficient, a common measure of inequality.

Occupational licensing generally limits market entry by setting higher entrance standards, such as high minimum educational credentials. Some argue that wage differentials are the result of the higher educational attainment of licensed professionals. However, Eckstein and Nagypál (2004) find that skill-based educational attainment45 cannot explain the wage differentials that exist between licensed and unlicensed occupations, and that increases in entry barriers could be considered a potential determinant of growing earnings inequality.

The literature so far mainly focuses on the within-sector effects of occupational licensing. Using decomposition analysis and comparing several licensed occupations to unlicensed occupations with similar skill requirements and tasks on the job, Kleiner (2006) reports that licensing increases earnings along the distribution curve, assuming there are no spillover effects to nonlicensed occupations. 46 The recent empirical evidence suggests otherwise:

44 Mouw and Kalleberg (2010) report that the changes in wage inequality could be linked to wage differences between occupations, but they are not able to identify the mechanisms that contribute to the between- occupation differences. Bol and Weeden (2014) also find large between-occupation wage inequality in Germany and the United Kingdom. However, Kim and Sakamoto (2008) suggest that, during a similar period (1980s- 2000s), the main driver of wage inequality was the rising within-occupation wage difference. In addition, the authors find that unionization and a larger share of part-time workers are positively associated with wage inequality. 45 Skill-based educational attainment usually refers to levels or years of schooling, not the field of study. 46 Kleiner (2006) compares five licensed occupations—physicians, dentists, lawyers, teachers and cosmetologists—to comparable nonlicensed occupations, including biological and life scientists, economists, sociologists, public relations specialists, bartenders, waiters and waitresses, and maids and housemen. He also simulates the wage distribution of these licensed occupations if the occupations were not regulated. His visual presentation suggests that licensed wage distributions are more compressed than both nonlicensed comparison 55

Kleiner and Krueger (2013) find that licensing does not reduce the variation in wages, despite the positive impacts on wages of working in a licensed occupation. Gittleman and Kleiner (2016) report that occupational licensing is associated with greater wage dispersion, but that the difference is not statistically significant.47

Some studies go beyond the distributional impact of licensing in a particular occupation, or set of occupations, and comparisons between occupational incumbents and individuals who are either working in comparable occupations, practitioners working the same occupations in a close jurisdiction where a license is not required, or practitioners with estimated earnings if the occupation were not licensed. Instead, they compare the impact between sectors. Criscuolo and Garicano (2010) estimate that occupational licensing requirements will further increase inequality, even under globalization. They find that workers will receive higher wages because of licensing requirements when their jobs are not offshorable. They also find that workers not protected by licensing will incur lower wages, and that licensing will continue to broaden the wage gaps between workers in different occupations. In addition, they note that when jobs can be moved around globally, the employment gap will also widen between licensed and nonlicensed occupations. In summary, there is lack of systematic analyses of whether occupational licensing has between-sector effects and how between-sector and within-sector effects jointly impact the overall earnings distribution, which could further influence earnings inequality.

3.2.2 The impact of unionization on the distribution of labour income

Empirical evidence suggests that the earnings structure is flatter in unionized sectors and that low inequality is associated with higher union density (Chamberlain, 1994; Card, 1996, 2001; Card, Lemieux, & Riddell, 2004; DiNardo et al., 1996; DiNardo & Lemieux, 1997; Firpo et al., 2009; Gustafsson & Johansson, 1999). Using micro-level data, Freeman (1980) finds that

groups and the hypothetically “nonlicensed” licensed groups, and the tighter the occupational closure, the more concentrated the wage distributions are. 47 The authors also find union status is associated with lower wage dispersion. 56

unionization reduces the wage dispersion among male workers. Gosling and Machin (1995) document the rising earnings inequality in the United Kingdom between 1980 and 1990, and suggest that the earnings dispersion of workers in the unionized workplace is lower than those in the nonunionized workplace. Both Card (1996) and Freeman (1993) report that in the 1980s, de-unionization accounted for approximately 20% of the increase in wage variance in the United States. The results of DiNardo et al. (1996) are partially in line with Card (1996) and Freeman (1993). They find that de-unionization accounts for one-third of the increase in the 90/10 ratio between wages at the top of the distribution (the 90th percentile) and the bottom of the distribution (the 10th percentile) for male workers, between 1979 and 1988. At the same time, however, de-unionization reduced the 50/10 ratio between wages at the middle of the distribution (the 50th percentile) and the bottom of the distribution (the 10th percentile) for male workers. Dinardo and Lemieux (1997) compare the differences in wage inequality and union density between the United States and Canada, and find that unions account for a large portion of the widening wage inequality. Card et al. (2004) also show that a sharp increase in wage inequality coincides with a fast decrease in union density in the United States, the United Kingdom, and Canada. Lemieux (2008) provides a comprehensive review of explanations of the growth in wage inequality and suggests that more recent data should be used to strengthen the correlational relationship between a decrease in unionization and an increase in wage inequality. Using a newly developed unconditional quantile regression method, Firpo et al. (2009) also report that the union wage premium does not monotonically decrease from the lowest to the highest earnings distribution. They find that the wage-compression effects of unionization are more prominent at the high end of the earnings distribution, while unionization increases inequality at the lower end of the earnings distribution. Using a regression discontinuity that is designed to compare pre- and post-unionization results, Frandsen (2012) shows that the impact of unionization on average wages is insignificant, but that unionization significantly flattens the distribution of employee earnings.

In summary, previous analyses mainly focus on unionized vs. nonunionized male workers in the United States. The distributional effects of unionization on female workers have not yet been thoroughly studied. Card (2001) reports that a change in the unionization rate does not explain the rising inequality in the wages of female workers in the American private sector

from the early 1970s to the early 1990s. In the Canadian context, the distributional effects of unionization are primarily estimated for male workers (Card et al., 2004). To fill this research gap, there is a need to investigate the impacts on the wage distribution of female workers.

3.3 Theoretical Framework

In the early stages of the industrial revolution, unskilled workers sought welfare protection through unionization, while professional practitioners achieved occupational closure through legislative support, such as establishment of licensing law or granting self-regulatory authorities to professional associations (Zhou, 1993). Kleiner (2000) suggests that lower-wage workers benefit from unionization, while higher-wage workers benefit from occupational licensing. Unlike unionization, which decreases earnings inequality, occupational licensing increases inequality by “first keeping out persons from entering higher wage occupations, and then by raising wages for persons in these already high income occupations” (Kleiner, 2000, p. 196).

There is a significant correlation between the variation in wages across occupations and economic inequality (Le Grand & Tåhlin, 2013). Unionization compresses the wage structure by increasing wages at the lower end of the income spectrum, while decreasing wages at the higher end (Chamberlain, 1994; Card, 1996; Card et al., 2004; DiNardo et al., 1996; Firpo et al., 2009; Fortin et al., 2012). In a competitive-market model, unions impact wage inequality in two different venues. First, through collective bargaining mechanisms, unionization decreases within-sector inequality by increasing the pay of lower-wage workers by proportionately more than higher-wage workers. Second, unionization increases between- sector inequality by restricting the supply of unionized workers, thereby pushing excess labour into nonunionized sectors and lowering the wages in the nonunionized sectors, which results in positive wage premiums for the unionized workers. This leads to higher earnings inequality between unionized and nonunionized sectors. In more complicated models, however, some nonunionized sectors pay higher wages because of the threat effects posed by the unionized sector. That is, companies will pay higher wages and provide more benefits to their workers so as to keep them from unionizing, which could potentially lead to more equality between

some sectors. In general, the spillover effects of unionization still generate more inequality across the entire wage distribution because the between-sector effects dominate the within- sector effects (e.g., Fortin & Lemieux, 1997; Freeman, 1980).

Licensing associations are very similar to craft unions, which organize their members by trade (e.g., nursing, teaching, and skilled trades in the construction industry). Contrary to workers in industrial unions, who support wage-levelling policies because they lower the incentive for employers to replace them with cheaper workers, one of the main objectives of craft unions is to restrict membership and, thereby, restrict competition for labour. Studies on craft unions suggest that excluding semiskilled or unskilled workers from the union directly contributes to inequality (e.g., Buchmueller, DiNardo, & Valetta, 2004; Rothstein, 1992).

The ostensive reason for licensing an occupation is to protect the public interest. However, many argue that professional associations leverage their licensing power to capture economic rents for their members. Unlike unionization, which has the power to set pay criteria based on job descriptions and seniority at the plant/firm level through collective bargaining, licensing restricts the labour supply at the sector/occupation level, which might translate into higher wages because of the limited services that licensed practitioners provide.48

The impact of occupational licensing on the earnings distribution should also be discussed in two different venues. On one hand, the exclusive right-to-practice that is created by tightening the labour supply will decrease the within-sector wage difference. Occupational licensing excludes practitioners from practising lower-quality licensed work for lower prices (e.g., dental hygienists and barbers). The quality of services will be relatively more homogenous compared to nonlicensed occupations, and the wage distribution will be more compressed because practitioners will not be able to differentiate their services from competitors and charge higher service fees or demand higher wages. On the other hand, the variation of licensing requirements could be associated with within-sector wage inequality as well.

48 Some regulatory bodies even provide their members with a compensation guide in order to inform them of the wage standard within the respective industry and to provide wage protection. 59

According to rent capture theory, tighter occupational requirements will result in higher wage premiums; in geographic areas with higher licensing standards, practitioners will earn more. This could be seen as another source of within-sector earnings inequality. In sum, occupational licensing flattens the wage distribution within licensed occupations in the same jurisdiction, but may widen the wage dispersion across jurisdictions. The overall impacts of licensing on occupational specific wage distribution remain ambiguous.

Licensing increases the between-occupations wage gap in the form of positive wage premiums, similar to union wage premiums. Professional practitioners generally have high levels of human capital, which is rewarded in the labour market. Shapiro (1986) views occupational licensing as imposing minimum requirements on human capital investment. In most cases, licensing requires a certain level of education, a specific training program, passing a qualification exam, and years of experience in the licensed occupation. Therefore, it is reasonable to conclude that holding a license will imply a higher level of productivity that should be compensated. The higher the entry standard, the more human capital the members possess and the higher the returns. Hence, licensing could contribute to wage inequality between licensed and unlicensed workers. In other occupations, those in which practitioners do not fully comply with licensing law requirements or those in which members only gain the “exclusive right-to-title” through certification, rather than by crowding out the unlicensed workers who have the competency to perform the job, there will still be certain wage premiums because of the signaling effect of the higher value of human capital.49 However, Weeden and Grusky (2014) argue that earnings inequality cannot be sufficiently explained by the increasing demand for skills. Occupational incumbents need more powerful tools to control competition and restrict entry into their occupational market.

Licensing requirements also increase the between-occupations wage gap by restricting entry to the professions. Professional associations seek licensing power from the government and tend to restrict entry by imposing higher occupational standards (e.g., higher educational

49 Certification may also slightly reduce competition from those who are not certified, although this impact is likely to be small. 60

attainment, more mandatory skill sets, and tougher entrance examinations) so as to protect members’ higher economic welfare. For some occupations, such as physicians and lawyers, the tight licensing requirements weed out unlicensed practitioners, and so provide leverage for licensed workers to charge higher prices for services and gain higher wage premiums. Entry restrictions set by the licensing bodies may push workers who failed to acquire a license to seek employment in the nonlicensed sectors, increasing the supply of labour in nonlicensed occupations and dragging down the wages of unlicensed workers. This spillover effect is very similar to the spillover effects of unionization, but without the possibility of restoring wage gaps because licensed and unlicensed workers are not as substitutable as unionized and nonunionized workers. With the gap between licensed and unlicensed workers widened by the licensing requirements and specific skills needed to practise, employers cannot simply hire unlicensed workers to replace licensed workers. Because of the legitimization of the restricted supply of specific skills, occupational licensing, like craft unions, will result in a positive wage premium. Therefore, excluding unlicensed practitioners could lead to higher earnings inequality.

In both scenarios, occupational licensing will increase the between-sector wage inequality, in contrast to unionization. Kalleberg et al. (1981) point out that occupational licensing can enhance employees’ power, and support more economic segmentation in the labour market. In turn, it can lead to higher wage gains for licensed workers and higher income inequality between licensed and unlicensed workers. Hence, occupational closure explains the “persistently high and growing pay for skilled occupations” (Weeden & Grusky, 2014, p. 485). In this sense, the increase in occupational licensing requirements should be considered an important predictor for the rising earnings inequality of recent years.

3.4 Data and Empirical Methods

3.4.1 Empirical methods

Several quantile regression methods have been adopted to examine the differential relationship between changes in individual characteristics and changes in the wage distribution across quantiles (e.g., Borah & Basu, 2013; Budig & Hodges, 2014; Machado & Mata, 2005). Such 61

research methods are particularly meaningful when looking at the heterogeneous impacts of the predictors across different parts of the outcomes distribution. In this study, I use the recentered influence function (RIF) of the unconditional quantile regression to estimate the extent to which licensed occupations and unions have an impact across the wage distribution (Firpo et al., 2009, noted as FFL). As suggested by FFL, the RIF provides a clearer picture of how these two institutional mechanisms directly impact the log of the wage when the impacts are different at different points along the wage distribution. Conditional quantile regression estimates suggest that unions reduce within-group dispersion among union members, conditional on the members “who share of the same values of the covariates” (Firpo et al., 2009, p. 963) (e.g., education, skills, age, etc.). The RIF provides an enhanced understanding by incorporating the change in union density across workers’ wage distribution and hence incorporates the between-group effects. The unconditional effects are averages of the conditional effects (Frölich & Melly, 2013). Borah and Basu (2013) perform an empirical comparison between the conditional and unconditional quantile regressions, and conclude that the results of the conditional quantile regression may not be “generalizable and interpretable in a policy or a population context” (Borah & Basu, 2013, p. 1067). While the conditional and unconditional average treatment effects have similar meanings because of the linearity of the expectation operator, this is not the case for quantiles. In a simple example of gender wage gaps, the unconditional 0.9 quantile refers to the high-wage workers, whereas the conditional 0.9 quantile refers to the high-wage workers within each gender class, who may not necessarily be high earners, overall. Presuming that males earn higher wages than do females (the empirical evidence typically suggests gaps of 15 to 30%), it may well be that the 0.9 quantile among females is lower than the 0.7 quantile for all males. Therefore, the interpretation of the 0.9 quantile is different for the conditional and unconditional quantiles. Hence, the unconditional quantile regression is superior. In particular, when policy-makers need to provide a universal policy initiative across the whole distribution, their decision-making

process will benefit more from the unconditional quantile regression results because the results will not change due to changes in certain characteristics.50

The model I use in this research is as follows:

퐹푌(log_ 푤푎푔푒) = Pr(푙푖푐푒푛푠푒푑) ∗ 퐹푌|푙푖푐푒푛푠푖푛푔(log_ 푤푎푔푒 | 푙푖푐푒푛푠푒푑) + Pr(푢푛푙푖푐푒푛푠푒푑) ∗

퐹푌|푙푖푐푒푛푠푖푛푔(log_ 푤푎푔푒 | 푢푛푙푖푐푒푛푠푒푑) (1),

where the log_ 푤푎푔푒 is the dependent variable of the hourly wage in 2014 constant dollars.

퐹푌(log_ 푤푎푔푒) represents the unconditional population log wage distribution. The key independent variable 푙푖푐푒푛푠푖푛푔 is a dummy variable that is equal to one if the worker is licensed (either licensed professions or compulsory skilled trades), and zero otherwise. Union status is another key independent variable in this research, coded one if a member of a union or covered by a collective agreement, and zero otherwise. An interaction term between licensing and unionization is added to see whether or not there exist differential impacts of the joint influences of these two institutions. Standard control variables in the Mincer equation, such as individual characteristics (i.e., age, gender, individual educational attainment, marital status) and job characteristics (i.e., professional certificate status, self-employment status, part- time worker status, permanent job status, public sector indicator, aggregate industry, and occupation groups) are also included.

3.4.2 Data

The empirical analysis is based on the Canadian Labour Force Survey (LFS). The data describes participants, from all ten Canadian provinces, who were surveyed during the period 1998 to 2014. The LFS is a national sample of Canadians, aged 15 and older, who were not institutionalized. Year 1998 was chosen as the starting year because most of the licensure

50 Several recent researchers use RIF to investigate the heterogeneous impacts across the distribution of labour market measures (e.g., Aeberhardt, Givord, and Marbot (2012) on the minimum wage; Chi and Li (2008) on the gender wage gap; Gunderson and Krashinsky (2015) on returns to apprenticeship; Killewald and Bearak (2014), and Budig and Hodges (2014) on the motherhood penalty; and Lindqvist and Vestman (2011) on cognitive and noncognitive returns). 63

legislation had been enacted before the 1990s, and the question on union status was only included in the LFS after 1997. The Labour Force Survey is the main labour market source to generate monthly key indicators on the Canadian labour market, such as unemployment rate.51 The sample includes unique survey respondents and is population-weighted to be representative of the Canadian labour market.

I restrict the analysis to paid workers, who were between the ages of 21 and 65 and who received labour earnings in the survey year. I also limit the survey years to 1998, 2002, 2006, 2010, and 2014. This yields a working sample of 844,300 observations.52

------Insert Figure 3.1 about here ------

Figure 3.1 illustrates the historical trend of the share of unionized workers and licensed workers in the sample, between 1998 and 2014. In Canada, approximately 11% of paid workers were identified as working in an occupation with a license requirement; and more than 30% of workers reported being covered by a collective agreement (or were a union member). The share of unionized workers was relatively stable during the observation period, with a slight decline from approximately 33% in 1998 to 31% in 2014. The share of licensed workers is consistent between 1998 and 2006 and increased to about 12% in the more recent years covered in the data.

The logarithm of the hourly wage53 is used as the dependent variable, as in conventional human capital wage equations. Two key independent variables are included in the analysis: 1) whether the individual’s job was protected by a trade union (either as a union member or from

51 The LFS is structured to survey the same survey respondents monthly for six consecutive months. Hence, annual aggregations contain respondents that have been repeatedly surveyed. I reran the model using only the data for March and September to verify the impacts of repeated sampling. The results are identical. Hence, aggregate annual data (i.e., 12 months) are used in this study. 52 Results for other years are available upon request. Number of observations was rounded to the nearest hundred, according to the data-release requirement by the Statistics Canada Research Data Centre. 53 I focused on the real hourly wages, and so the reported hourly wages are converted to 2014 constant dollar. 64

being covered by a collective agreement, and 2) whether the individual worked in a licensed occupation. The independent variable─union membership─was drawn from the survey responses, while the licensing variable was constructed by linking the individuals’ respective occupations and provinces of residence, recorded in the survey, to the information I collected on the enacted legislations across the ten provinces, as discussed in the Appendix (i.e., Chapter 5 in the thesis). In the licensing literature, this linking procedure was used by Gittleman and Kleiner (2016). 54 The respondents’ occupations were identified based on the four-digit National Occupational Classification (NOC 2006). These two key independent variables are not mutually exclusive. Table 3.1 shows the share of licensed workers in both the unionized and nonunionized workplace, in the selected years. More licensed workers were employed in the unionized workplace than in the nonunionized workplace (about 15% and 7%, respectively). The shares of licensed workers also increased over the years under review in both types of workplace. ------Insert Table 3.1 about here ------

An array of variables is included as controls, including individual characteristics (gender, age, marital status, education level, years of work experience, and province of residence) and job characteristics (indicators of part-time status, multiple-job holders, self-employment status, and working in the public sector; categories of workplace size by number of employees; two- digit occupation categories; and two-digit industry dummies). Education level is coded as high school graduate or less; college diploma or equivalent; and university bachelor’s degree and above.

Table 3.2 shows the descriptive statistics for the full sample and split by gender. After converting to constant dollars, the mean hourly wage rate is $CAD25.54 (SD=$CAD13.01).

54 Girard and Smith (2013) use a similar approach to link regulated occupational status to the 2006 Canadian population census. They assume that in the year 2006 all regulated occupations are fully implemented. Thus, Gittleman and Kleiner’s (2016) approach is utilized to take into consideration of years that the corresponding licensing laws were proclaimed. 65

The mean age of workers in the private sector is 42.8 years (SD=12.6). Approximately 13% of workers are identified as working in a licensed occupation, and 32.1% of workers are unionized. The selected samples for both male and female workers are relatively balanced. The mean ages of the male and female workers in the sample are the same. The wage differential between male and female workers is high—the mean hourly wage rate for male workers is $CAD27.45 (SD=$CAD13.83), while the mean hourly wage rate for female workers is $CAD23.59 (SD=$CAD11.82). The share of workers employed in a unionized workplace or holding a licensed occupation is also different between males and females. The percentage of male licensed workers is larger than that for female workers, at 13.8% (SD=0.345) and 11.4% (SD=0.318), respectively, while a larger percentage of female workers (33.9%, SD=0.473) are working in a unionized workplace than male workers (30.4%, SD=0.460). ------Insert Table 3.2 about here ------

Figure 3.2 also illustrates the wage distributions of licensed and unlicensed workers in the selected years. The wage densities are quite different between licensed and unlicensed workers. The wage distribution of licensed workers is narrower and denser; while the wage distribution of unlicensed workers is more widely spread. Over time, the wage distribution of licensed workers shifts toward the right, which suggests that more licensed workers are earning higher wages. At the same time, the wage distribution of unlicensed workers remains stable in the top half, while at the bottom half, it is eroding. Both results suggest increasing inequality due to occupational licensing. ------Insert Figure 3.2 about here ------

Figure 3.3 illustrates the mean hourly wage differential between licensed and unlicensed, unionized and nonunionized workers over the wage distribution by year. The raw wage differentials between unionized and nonunionized workers increase for the bottom 20% of wage earners, peak at around the 20th percentile, then gradually decrease and drop to a negative number among the top 5% of wage earners. The raw wage differential between licensed and

unlicensed workers also exhibits a bell-shaped curve: increasing for the bottom 20% of the wage earners, remaining relatively steady or declining slightly, but overall staying positive. One particularly interesting observation is the pattern of wage differentials over time: the union wage differentials decrease at around 2000, but hold steady from 2010 onward over the whole wage distribution. The occupational licensing wage differentials are very stable year over year, and even increase slightly for the workers in the upper middle of the distribution, in particular for workers at the 55th–65th percentiles. In addition, occupational licensing wage differentials are higher than union wage differentials over almost the entire wage distribution, except for the bottom 15–20% of wage earners. The differences between two wage differentials have widened over the years along the wage distribution. ------Insert Figure 3.3 about here ------3.5 Results

Table 3.3 shows the RIF-OLS estimated coefficients of the log wages model for the 10th, 25th, 50th, 75th and 90th quantiles for selected years. The results are compared with the standard OLS estimates. The estimated coefficients of the impacts of both unionization and licensing status are also charted in Figure 3.4 in greater detail across the wage distribution.

The Standard OLS estimates show that, compared to nonunionized workers, unionized workers on average earned 10.8% (p<0.001)55 more in 1998, 10.1% (p<0.001) more in 2002, 9.3% (p<0.001) more in 2006, 8.3% (p<0.001) more in 2010 and 7.9% (p<0.001) more in 2014, respectively. In 1998, the quantile estimated wage premium of unionization increases from 13.0% (p<0.001) at the 10th percentile to 22.7% (p<0.001) at the 25th percentile and peaks at the 30th percentile (24.6%, p<0.001). The quantile estimated coefficients decrease between the 30th and 95th percentiles; the estimated RIF coefficient decreases to 17.4% (p<0.001) at the

55 In log earnings equations, the true proportional change is exp(β) − 1, where β is the estimated coefficient. For low values of β, the approximation is very close, underestimating the true value by less than 0.02 for values of β < 0.20, which is the case for our main variables of interest; hence, I report the β coefficients. 67

mean, continues to decline to 5.9% at the 75th percentile, and finally reaches negative numbers at the 90th percentile (-9.1%, p<0.001). The overall shapes of the estimated wage premiums over the entire distribution for each of other years are very similar to the shape of the estimated premiums for 1998. The highest wage premium of unionization lies in the middle of the bottom half of the distribution, at the 25th to 30th percentiles, and the estimated coefficients continue to decline between the 35th and 95th percentiles. For example, the estimated wage premium of unionization for 2014 first increases from 10.6% (p<0.001) at the 10th percentile to 17.8% (p<0.001) at the 25th percentile, peaks at the 30th percentile, decreases to 13.2% (p<0.001) at the mean, continuously declines to 3.5% at the 75th percentile, and eventually goes to a negative number at the 95th percentile (-8.1%, p<0.001). ------Insert Table 3.3 about here ------

Focusing on the top and bottom of the wage distribution, the unconditional quantile estimated coefficients suggest that there exists an equality-enhancing within-group wage compression effect of unionization. However, there is a strong inequality-enhancing between-group effect of unionization in the lower middle of the wage distribution. Therefore, in contrast to the conditional quantile estimates that uniformly support wage compression, the unconditional quantile estimates suggest that both the between-group and the within-group effects jointly result in different wage premiums at different points along the wage distribution. In addition, there is variation in union density across the wage distribution. The higher union density in the middle range of the wage distribution could potentially explain the increased between-group inequality suggested by the unconditional quantile estimates.

Based on the different wage premiums at the top and bottom of the wage distribution, I conclude that the wage-equalizing effects of unionization are not as strong as was previously estimated using conditional quantile regression techniques. To a certain degree, unionization still effectively reduces inequality between the lowest- and highest-wage earners. However, the middle-range-wage earners gain more from unionization than do workers at both ends of the wage distribution. I observe a decrease in union wage premiums over time for the middle range of the wage distribution, in which the peak wage premium is approximately 25% and 68

gradually declines to around 18%. The unconditional quantile wage premiums of unionization decrease minimally at the lower end while increasing slightly at the higher end of the wage distribution between 1998 and 2014, which implies that unionization cannot effectively explain the increasing earnings inequality in Canada over the past two decades.

The estimated wage premiums of holding a licensed occupation also vary across the wage distribution. The standard OLS estimates show that workers employed in a licensed occupation earned, on average, between 7.6% (p<0.001) and 13.9% (p<0.001) more than workers in a nonlicensed occupation in 1998 and 2014, respectively. The unconditional quantile estimates suggest that the impact of occupational licensing is also different at different points along the wage distribution. For the year 1998, the estimated wage effect increases from 1.2% (p<0.001) at the 10th percentile, gradually climbs to 3.7% (p<0.001) at the 25th percentile, continues to increase steadily between the 25th and 75th percentile, gradually declines between the 75th and the 95th percentiles, and eventually holds steady at approximately 5.9%. The overall shapes of the estimated wage premiums of occupational licensing over the entire distribution for each of other years are gradually shifting away from the shape of the estimated premiums in 1998, particularly in the upper half of the distribution: For example, the estimated wage premium for 2002 increases from 1.9% (p<0.001) at the 10th percentile, to approximately 3.3% at the 25th percentile, gradually increases to 17.9% at the 75th percentile, and then continually decreases to 12.4% at the 90th percentile. In more recent years, the shape of the licensing wage premiums also implies a minimal wage-enhancing effect at the bottom of the wage distribution, a moderate effect for the top of the wage distribution, and most significant, steadily increasing, wage gain in the middle. The estimates suggest that the between-group inequality-enhancing effects dominate across the majority of the wage distribution, and are particularly strong at the high end of the wage distribution. The higher-wage earners gain the most from occupational licensing—the higher the wage, the more licensing premiums those wage earners enjoy— while the workers at the bottom end of the wage distribution benefit only minimally from being employed in a licensed occupation. In sum, occupational licensing increases inequality.

In addition, the union wage premiums have eroded over the years, where licensing wage premiums seem to have increased over the past two decades. Between 1998 and 2002, there

was a sharp increase in licensing wage premiums for the upper half of wage earners, in particular between the 50th and 85th percentiles; the wage premiums almost doubled between 1998 and 2014. For example, the wage premium at the 75th percentile is 12.5% in 1998, and 26.4% in 2014. Reasons for this jump are not obvious.

The estimated effects, further taking the interaction term of licensing and union status into consideration, are reported in Table 1 of the Appendix 3A. The interaction effects of licensing and unionization show another interesting pattern. The estimated coefficients of the interaction terms are negative at the bottom half of the wage distribution, while they are positive at the top half. This suggests that the two labour market institutions, occupational licensing and unionization, have a rather complicated relationship. These two institutions act as a competing force for the lower income earners while they have an additive effect for the higher income earners.

After controls, the wage premiums are higher for unionization than for occupational licensing for the bottom half of the wage distribution, and lower for the upper half of the wage distribution. Thus, for the highest quartile of the wage distribution, unionization provides less wage protection than does occupational licensing, while unionization provides stronger wage protection than does occupational licensing for the lower half of wage earners. Combining the between- and within-effects, occupational licensing may lead to higher wage inequality, while the negative impact of unionization on wage inequality was not sufficiently evident to draw a conclusion that the declining union density is associated with increasing income inequality.

3.5.1 Differences by gender

I further split the sample by gender to investigate whether occupational licensing and unionization have different impacts on the wage distribution for male and female workers. Table 3.4 shows the unconditional quantile estimates of the log wages model for the 10th, 25th, 50th, 75th, and 90th quantiles, for paid male and female workers, for the years 1998 and 2014. The overall patterns of the estimated premiums for the male workers, for both unionization and occupational licensing across the wage distribution, are comparable to the full sample. For male workers, unionization yields strong positive premiums to wage earners in the lower mid-

range, compared to those at both ends of the wage distribution. At the same time, occupational licensing yields moderate positive premiums to the higher mid-range wage earners. In particular, the licensing wage premiums for male workers have been constantly increasing at the top quartile since 2002. Again, unionization provides much larger wage premiums than does occupational licensing for male lower-wage workers, compared to the base group. For example, the union premiums for male wage earners at the 25th percentile are 21.5% (p<0.001) and 16.0% (p<0.001) in 1998 and 2014, respectively; while the licensing premium is 2.3% (p<0.001) in 1998 and 6.2% (p<0.001) in 2014 for the same group of workers. The opposite is observed for male higher-wage earners: The union premiums at the 75th percentile are merely 1.0% (p<0.001) in 1998, and 2.5% (p<0.001) in 2014. However, the licensing wage premiums increase sharply from 7.5% (p<0.001) to 15.5% (p<0.001) in 1998 and 2014, respectively. The estimated coefficients of the impacts of both unionization and the licensed occupation for male workers are also presented in Figure 3.5, in greater detail across the wage distribution.

------Insert Table 3.4 about here ------

The overall pattern of the estimated union wage premiums for female paid workers is consistent with the pattern for all paid workers (see Figure 3.6). However, the equality- enhancing effects of unionization are only observed at the upper half of the wage distribution, while inequality increases at the lower half between unionized and nonunionized female workers. Surprisingly, occupational licensing increases inequality among female workers across the entire wage distribution. In particular, a sharp wage gain is observed between the 50th and 80th percentile. The wage premium of occupational licensing begins at 1.4% at the 10th percentile, continually increases to 14.1% at the mean, continues to increase to 24.2% at the 80th percentile, and then gradually declines to 18.7% at the 90th percentile. In 2014, the estimated wage premiums of licensing are higher for the higher-paid female workers than for higher-paid male workers: the wage premium is 1.2% (p<0.001) at the 10th percentile, increases to 6.8% (p<0.001) at the 25th percentile, and continues increasing to 22.0% (p<0.001) at the mean, before keeping rising to 43.3% (p<0.001) at the 75th percentile, and eventually decreasing to 10.2% (p<0.001) at the 90th percentile. These findings show that occupational

licensing yields higher premiums than unionization for female workers at the top half of the wage distribution, while unionization provides higher premiums for the rest of the female workers covered in the study. ------Insert Figure 3.5 & 3.6 about here ------3.6 Conclusions

The unconditional quantile (RIF) estimates suggest that unionization could contribute to an increase in inequality at the bottom half of the wage distribution and to a decrease in wage inequality at the top half. The estimates also suggest that occupational licensing could contribute to an increase in inequality across the overall wage distribution. Occupational licensing increases inequality through the occurrence of different wage premiums across the wage distribution. The lower-wage earners receive lower wage premiums, while the higher- wage earners received relatively higher wage premiums when working in a licensed occupation, compared to unlicensed workers.

The wage impacts of occupational licensing vary substantially by gender, although the wage impacts of unionization are more in line with the pooled sample. For male workers, the impacts of occupational licensing are similar to those for the full sample. However, for female workers, working in a licensed occupation contributes to a significant increase in wage inequality across the entire wage distribution, particularly between the 25th percentile and 80th percentile.

The empirical analysis indicates that both unions and occupational licensing raise the pay of their members – about 11% for unions and 9% for occupational licensing. Unions, however, tend to reduce earnings inequality while licensing tends to increase inequality. The union impact on pay tends to be similar for males and females, but occupations licensing tends to have a larger impact on the pay of females compared to males so it reduces the male-female pay gap.

The impacts of unionization on wage inequality have been investigated intensively in the literature. The impacts of occupational licensing on wage inequality, however, have yet to be 72

thoroughly examined. The reason that occupational licensing has been overlooked so far could be that, unlike trade unions, which directly negotiate wages for their members, and hence have a direct connection to earnings inequality, occupational licensing is not explicitly tied to members’ earnings. Another reason for the neglect of licensing may be the difficulty of designating licensed occupations, given differences across provinces as well as the fact that current population surveys tend not to designate licensed workers. As shown in this study, the theories of occupational licensing and the empirical evidence both support the rent-capture behaviours of occupational licensing. Future studies could look further into the issue because occupational licensing has, at least in the United States, become a more prominent labour market institution than unionization, and the share of licensed workers has sharply increased. In Canada, the economic impacts of occupational licensing have also drawn more attention from policy-makers, despite the fact that the share of licensed workers has remained stable over the past three decades. More empirical evidence would facilitate a better understanding of this issue in evidence-based policy-making.

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Figure 3.1 The Share of Workers Covered by Unions, and Licensed and Certified Workers in Canada, 1998–2014

Note: Licensed workers include both licensed professionals and workers in compulsory skilled trades. A certified worker denotes certified professions who have exclusive right-to-title protection by the provincial occupational legislation. Weighted sample of the cross-sectional Canadian Labour Force Survey (1998-2014). Data constructed based on the licensing and certification data that author collected and presented in Chapter 1. Detailed discussion about the construction of the licensing and certification indicators can be found in Chapter 5.

Table 3.1 Share of Licensed Workers Covered by Collective Agreement/Not Covered by the Collective Agreement, by Year

Year 1998 2002 2006 2010 2014 share of licensed workers in unionized workplace 0.145 0.140 0.148 0.165 0.173 Std dev [0.353] [0.347] [0.355] [0.371] [0.378] share of licensed workers in nonunionized workplace 0.068 0.067 0.070 0.086 0.085 Std dev [0.252] [0.251] [0.255] [0.281] [0.279]

Note: Weighted pooled sample of five years of the Canadian Labour Force Survey data (1998/2002/2006/2010/2014).

Table 3.2 Descriptive Statistics (All Paid Workers)

Pooled sample Male Female Variables Mean Std Dev Mean Std Dev Mean Std Dev Hourly wages ($) 25.54 [13.01] 27.45 [13.83] 23.59 [11.82] Log of hourly wages 3.123 [0.482] 3.196 [0.483] 3.048 [0.470] Licensed occupations (both licensed profession and compulsory skilled trades) 0.127 [0.333] 0.138 [0.345] 0.114 [0.318] Certified professions 0.034 [0.181] 0.020 [0.141] 0.049 [0.216] Union members 0.321 [0.467] 0.304 [0.460] 0.339 [0.473]

Female 0.500 [0.500] Age (years) 42.8 [12.6] 42.8 [12.6] 42.8 [12.6] Married 0.696 [0.460] 0.682 [0.466] 0.709 [0.454] Education High school or lower 0.292 [0.455] 0.314 [0.464] 0.270 [0.444] College 0.426 [0.494] 0.425 [0.494] 0.426 [0.494] University and above 0.282 [0.450] 0.260 [0.439] 0.304 [0.460]

Multiple-job holder 0.053 [0.223] 0.045 [0.207] 0.061 [0.239] Part-time status 0.151 [0.358] 0.082 [0.274] 0.226 [0.418] Temporary worker 0.114 [0.317] 0.108 [0.311] 0.119 [0.323] Public sector job 0.248 [0.432] 0.190 [0.391] 0.310 [0.463] Plant size Fewer than 20 employees 0.180 [0.384] 0.177 [0.382] 0.182 [0.386] 20–99 employees 0.161 [0.368] 0.176 [0.381] 0.147 [0.354] 100–499 employees 0.143 [0.351] 0.154 [0.361] 0.133 [0.339] More than 500 employees 0.515 [0.500] 0.493 [0.500] 0.538 [0.499] # observations 844,300 417,000 427,300 Note: Weighted pooled sample of five years of the Canadian Labour Force Survey data (1998/2002/2006/2010/2014). Descriptive statistics are rounded according to Statistics Canada protocol.

Figure 3.2 Hourly Wage Distribution (All Paid Workers), by Year

Figure 3.2 Hourly Wage Distribution (All Paid Workers), by Year (Cont’d)

Source: Labour Force Survey, 1998/2002/2006/2010/2014. Converted to 2014 constant dollars. Population weighted.

Figure 3.3 Mean Difference of Logarithm of Real Hourly Wage over Wage Distribution: Licensed vs. Unlicensed; Union vs. Nonunion, by Year

Figure 3.3 Mean Difference of Logarithm of Real Hourly Wage over Wage Distribution: Licensed vs. Unlicensed; Union vs. Nonunion, by Year (Cont’d)

Source: Labour Force Survey, 1998/2002/2006/2010/2014. Converted to 2014 constant dollars. Population weighted.

Table 3.3 Unconditional Quantile Regression Estimates of Log Wage Premiums for Licensed Occupational and Union Status, by Year

OLS Q10 Q25 Q50 Q75 Q90 1998 [N=528,900] Licensed 0.076*** 0.012*** 0.037*** 0.115*** 0.125*** 0.106*** occupation (0.003) (0.003) (0.003) (0.003) (0.004) (0.005) Union status 0.108*** 0.130*** 0.227*** 0.174*** 0.059*** -0.091*** (0.010) (0.002) (0.003) (0.003) (0.003) (0.003) R2 0.464 0.192 0.312 0.333 0.276 0.189

2002 [N=540,900] Licensed 0.096*** 0.019*** 0.033*** 0.137*** 0.179*** 0.124*** occupation (0.007) (0.003) (0.003) (0.003) (0.004) (0.006) Union status 0.101*** 0.147*** 0.205*** 0.164*** 0.048*** -0.086*** (0.010) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.476 0.199 0.303 0.342 0.295 0.216

2006 [N=654,900] Licensed occupation 0.111*** 0.017*** 0.044*** 0.152*** 0.201*** 0.139*** (0.007) (0.003) (0.003) (0.003) (0.004) (0.006) Union status 0.093*** 0.117*** 0.206*** 0.145*** 0.050*** -0.076*** (0.010) (0.002) (0.003) (0.003) (0.003) (0.003) R2 0.453 0.181 0.291 0.324 0.288 0.203

2010 [N=574,000] Licensed 0.124*** 0.010*** 0.048*** 0.179*** 0.239*** 0.130*** occupation (0.012) (0.003) (0.003) (0.003) (0.004) (0.005) Union status 0.083*** 0.112*** 0.173*** 0.136*** 0.033*** -0.073*** (0.005) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.450 0.178 0.284 0.328 0.288 0.190

2014 [N=572,700] Licensed 0.139*** 0.013*** 0.065*** 0.206*** 0.264*** 0.126*** occupation (0.009) (0.002) (0.003) (0.003) (0.005) (0.005) Union status 0.079*** 0.106*** 0.178*** 0.132*** 0.035*** -0.081*** (0.009) (0.002) (0.003) (0.003) (0.003) (0.003) R2 0.455 0.183 0.309 0.332 0.287 0.191

Individual characteristics Y Y Y Y Y Y Job characters Y Y Y Y Y Y

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation, zero otherwise. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and experience) and job characteristics (certification status, self-employment status, multiple-job-holder status, permanent employment status, part-time employment status, public sector dummy, plant size, and categories of industry and one-digit occupations). The number of observations was rounded to the nearest hundred, according to the data-release requirement by the Statistics Canada Research Data Centre.] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Figure 3.4 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, All Paid Workers, by Year

Figure 3.4 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, All Paid Workers, by Year (Cont’d)

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors. Occupational licensing is a step function equal to one if the individual works in a licensed occupation, and zero otherwise. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and work experience) and job characteristics (certification status, self-employment status, multiple- job-holder status, permanent-employment status, part-time-employment status, public sector dummy, plant size, and categories of industry and one-digit occupations).] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Table 3.4 Unconditional Quantile Regression Estimates of Log Wage Premiums for Licensing and Union Status, by Gender

Male Female Licensing Union status Licensing Union status Year=1998 Q10 0.005 0.175*** 0.014*** 0.103*** (0.006) (0.004) (0.003) (0.003) Q25 0.023*** 0.215*** 0.040*** 0.210*** (0.005) (0.004) (0.004) (0.004) Q50 0.073*** 0.162*** 0.141*** 0.201*** (0.004) (0.003) (0.005) (0.004) Q75 0.075*** 0.010*** 0.242*** 0.076*** (0.005) (0.003) (0.007) (0.004) Q90 0.043*** -0.135*** 0.195*** -0.030*** (0.006) (0.004) (0.010) (0.005) N 272,200 256,700

Year=2014 Q10 0.007 0.127*** 0.012*** 0.091*** (0.005) (0.005) (0.002) (0.003) Q25 0.062*** 0.160*** 0.068*** 0.187*** (0.004) (0.004) (0.004) (0.004) Q50 0.135*** 0.144*** 0.220*** 0.117*** (0.004) (0.004) (0.004) (0.004) Q75 0.155*** 0.025*** 0.433*** 0.010* (0.005) (0.004) (0.008) (0.005) Q90 0.090*** -0.089*** 0.227*** -0.050*** (0.006) (0.004) (0.008) (0.005) # observations 284,900 287,800 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors in parentheses. Occupational licensing is a step function equal one if the individual works in a licensed occupation, and zero otherwise. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and experience) and job characteristics (certification status, self- employment status, multiple-job-holder status, permanent-employment status, part-time-employment status, public dummy, plant size, and categories of industry and one-digit occupation). Number of observation was rounded to the nearest hundred according to the data-release requirement by Statistics Canada Research Data Centre.] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Figure 3.5 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, Male Paid Workers, by Year

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors. Occupational licensing is a step function equal one if the individual works in a licensed occupation, and zero otherwise. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and work experience) and job characteristics (certification status, self-employment status, multiple- job-holder status, permanent-employment status, part-time-employment status, public sector dummy, plant size, and categories of industry and one-digit occupation).] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Figure 3.6 Unconditional Quantile Regression Estimates of Log Wages Premiums for Occupational Licensing and Union Status, Female Paid Workers, by Year

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors. Occupational licensing is a step function equal one if the individual works in a licensed occupation, and zero else. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and work experience) and job characteristics (certification status, self-employment status, multiple-job- holder status, permanent-employment status, part-time-employment status, public sector dummy, plant size, and categories of industry and one-digit occupation).] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Appendix 3A

Table 3A.1 Unconditional Quantile Regression Estimates of Log Wages Premiums for Licensed Occupational and Union Status, Interaction Term Included

OLS Q10 Q25 Q50 Q75 Q90 1998 [N=528,900] Licensed 0.077*** 0.071*** 0.123*** 0.116*** 0.035*** 0.027*** occupation (0.006) (0.005) (0.005) (0.005) (0.006) (0.007) Licensed * union -0.005 -0.110*** -0.159*** -0.003 0.161*** 0.141*** (0.015) (0.005) (0.006) (0.003) (0.007) (0.010) Union status 0.111*** 0.147*** 0.250*** 0.177*** 0.040*** -0.107*** (0.013) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.464 0.193 0.313 0.333 0.277 0.189

2002 [N=540,900] Licensed 0.115*** 0.092*** 0.127*** 0.122*** 0.102*** 0.152*** occupation (0.007) (0.004) (0.005) (0.005) (0.006) (0.008) Licensed * union -0.037* -0.138*** -0.177*** 0.026*** 0.142*** -0.051*** (0.018) (0.005) (0.006) (0.006) (0.008) (0.011) Union status 0.106*** 0.166*** 0.231*** 0.162*** 0.030*** -0.081*** (0.011) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.477 0.200 0.305 0.342 0.296 0.216

2006 [N=654,900] Licensed occupation 0.123*** 0.077*** 0.135*** 0.138*** 0.112*** 0.160*** (0.004) (0.004) (0.005) (0.005) (0.006) (0.008) Licensed * union -0.026* -0.117*** -0.177*** 0.025*** 0.168*** -0.041*** (0.012) (0.005) (0.006) (0.006) (0.008) (0.011) Union status 0.099*** 0.134*** 0.233*** 0.144*** 0.028*** -0.072*** (0.010) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.454 0.182 0.293 0.324 0.289 0.203

2010 [N=574,000] Licensed 0.124*** 0.062*** 0.124*** 0.164*** 0.133*** 0.107*** occupation (0.014) (0.004) (0.004) (0.005) (0.006) (0.007) Licensed * union 0.001 -0.105*** -0.152*** 0.032*** 0.215*** 0.048*** (0.001) (0.004) (0.005) (0.006) (0.008) (0.009) Union status 0.083*** 0.130*** 0.199*** 0.129*** -0.004 -0.082*** (0.005) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.450 0.178 0.286 0.328 0.291 0.190

2014 [N=572,700] Licensed 0.131*** 0.057*** 0.134*** 0.163*** 0.144*** 0.133*** occupation (0.010) (0.004) (0.004) (0.005) (0.006) (0.007) Licensed * union 0.016** -0.091*** -0.143*** 0.088*** 0.248*** -0.012 (0.007) (0.004) (0.005) (0.006) (0.008) (0.009) Union status 0.076*** 0.121*** 0.204*** 0.117*** -0.009*** -0.080*** (0.008) (0.003) (0.003) (0.003) (0.003) (0.003) R2 0.450 0.183 0.310 0.333 0.290 0.193

Individual characteristics Y Y Y Y Y Y Job characteristics Y Y Y Y Y Y

Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation, and zero otherwise. Each quantile regression includes a constant as well as controls for individual (gender, education level, marital status, and experience) and job characteristics (certification status, self- employment status, multiple-job-holder status, permanent-employment status, part-time-employment status, public sector dummy, plant size, and categories of industry and one-digit occupations). The number of observations was rounded to the nearest hundred, according to the data-release requirement by the Statistics Canada Research Data Centre.] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

Table 3A.2 Unconditional Log Wage Premiums for Licensing and Union Status; Interaction Term Included, by Gender, 1998 vs. 2014

Male Female Licensing Union status Licensing Union status Year=1998 Q10 0.074*** 0.197*** 0.075*** 0.108*** (0.008) (0.005) (0.004) (0.003) Q25 0.079*** 0.234*** 0.209*** 0.223*** (0.007) (0.004) (0.007) (0.004) Q50 0.052*** 0.157*** 0.255*** 0.121*** (0.006) (0.004) (0.007) (0.004) Q75 0.015** -0.005 0.294*** -0.090*** (0.006) (0.003) (0.012) (0.004) Q90 0.020** -0.140*** 0.167*** -0.067*** (0.009) (0.004) (0.013) (0.004) N 272,200 256,700

Year=2014 Q10 0.038*** 0.142*** 0.075*** 0.108*** (0.007) (0.005) (0.004) (0.003) Q25 0.088*** 0.175*** 0.209*** 0.223*** (0.006) (0.004) (0.007) (0.004) Q50 0.103*** 0.131*** 0.255*** 0.121*** (0.006) (0.004) (0.007) (0.004) Q75 0.100*** -0.001 0.294*** -0.036*** (0.007) (0.004) (0.012) (0.005) Q90 0.096*** -0.088*** 0.167*** -0.067*** (0.008) (0.004) (0.013) (0.004) # observations 284,900 287,800 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Population weighted regression models. Clustered standard errors in parentheses. Occupational licensing is a step function equal one if the individual works in a licensed occupation, and zero otherwise. Each quantile regression includes interaction term of licensing and unionization, a constant as well as controls for individual (gender, education level, marital status, and experience) and job characteristics (certification status, self-employment status, multiple-job-holder status, permanent-employment status, part-time-employment status, public sector dummy, plant size, and categories of industry and one-digit occupation). Number of observation was rounded to the nearest hundred according to the data release requirement by Statistics Canada Research Data Centre.] Source: Labour Force Survey, 1998/2002/2006/2010/2014.

100

Chapter 4 Does Occupational Licensing Restrict the Interprovincial Labour Mobility of Young Workers?

4.1 Introduction

The economic impact of occupational regulations on workers’ labour market outcomes has drawn growing research interest in recent years. Despite the importance of mobility in reducing economic inefficiencies in the labour market, the relationship between labour mobility— particularly internal labour mobility—and occupational licensing has received little attention in the literature, compared to studies investigating the impacts of occupational licensing on other aspects of the economy, such as wage premiums, employment, or quality of services.56

Allowing human capital to move across geographic areas more freely can help resolve regional labour shortages in the receiving regions, and labour surpluses in the sending regions. In addition, relocating workers internally to address local labour market inefficiency is a quicker, more economically efficient, and more politically acceptable policy response than policies that bring in external human capital, such as the skilled-worker immigration program or the temporary foreign workers program.

Occupational licensing creates labour market segregation by limiting the entry into and practising of certain occupations; this leads to potential labour misallocation. Moreover, licensing requirements vary across jurisdictions, which may cause further mobility restrictions to licensed practitioners. Inconsistent licensing requirements across jurisdictions decrease the net gains of internal mobility for licensed workers by raising the cost of mobility—if workers need to be relicensed in the new jurisdiction—or lowering the potential wage premiums—if the licensed practitioners move to a jurisdiction with lower or no licensing requirements.

56 In his two books of compressive review of the antecedents and consequences of occupational licensing, Kleiner barely mentions labour mobility (Kleiner, 2006, 2013). 101

Policy-makers typically view reductions to interjurisdictional differences in licensing requirements and implementation of reciprocal recognition of licenses between jurisdictions as effective ways to limit the negative consequences of occupational licensing.

In its proposed budget for the 2016 fiscal year, the U.S. Department of Labor allocated an additional $15 million, over the previous year's budget, for grants “to identify, explore, and address areas where occupational licensing requirements create an unnecessary barrier to labor-market entry or labor mobility" (U.S. Department of Labor, 2015, p. 23).57 Despite policy-makers’ desire to understand the labour-mobility issue, the empirical evidence is both limited and inconclusive. Most studies find that inconsistent occupational licensing requirements across states reduce the mobility of licensed professionals (Arbury, Bonilla, Durfee, Johnson, & Lehninger, 2015; Holen, 1965; Johnson & Kleiner, 2015; Kleiner, 2015; Kleiner, Gay, & Greene, 1982; Mulholland & Young, 2016; Pashigian, 1979), while other studies report that no significant relationship can be drawn (DePasquale & Strange, 2016; Prantl & Spitz-Oener, 2009; Zapletal, 2014). In Canada, federal laws, such as the Agreement on Internal Trade (1995), were passed so as to reduce mobility barriers for workers in the occupations that are “certified or licensed by provincial or territorial authorities.”58 The limited Canadian evidence reports lower interprovincial mobility for workers in regulated professions than workers in unregulated occupations (Chen & Fougère, 2011; Grady & Macmillan, 2007; Muzondo & Pazderka, 1980).

This study aims to provide empirical evidence on whether or not occupational licensing affects the interprovincial mobility of young Canadian workers. To answer this question, multilevel modelling analysis with random effects is used. I analyze the longitudinal Canadian Survey of Labour and Income Dynamics data for the period 1993 to 2010, and focus on testing whether or not occupational licensing deters worker mobility, compared to analogous groups who are not licensed. The results are not in line with the theoretical prediction that licensing reduces

57 https://www.dol.gov/sites/default/files/documents/general/budget/FY2017BIB_0.pdf 58 http://www.ait-aci.ca/labour-mobility/ 102

interprovincial mobility after controlling for the effect of other factors affecting mobility. On the contrary, estimates suggest small, positive, but statistically insignificant relationships between licensing status and probability of moving to a different province. By contrast, unionization significantly limits workers’ likelihood to move across provincial boundaries. There are no gender differences in the impact of licensing requirements on mobility. The effect of union status, by contrast, is impacted by gender: while union status does negatively impact workers’ likelihood of internal migration significantly for both male and female workers, the magnitude of estimates are larger for male workers than their female counterparts. The empirical results here are different from those reported in earlier studies in the United States (e.g., Johnson & Kleiner, 2015; Kleiner et al., 1982; Pashigian, 1979; Pratt, 1980). One possible explanation is that the federal Internal Agreement of Trade effectively reverses the mobility restriction caused by various provincial licensing requirements.

The paper is organized as follows: in section 5.2 and 5.3, I discuss the theoretical framework and provide a review of the literature on the impact of occupational licensing on interjurisdictional labour mobility (both intercountry and interprovincial/interstate) in Canada and the United States. Subsequently, I discuss the quantitative research methods and the data used. In section 5.5, I report the results on the impact of occupational licensing on interprovincial mobility. Finally, I discuss the recent Canadian laws that aim to lower mobility barriers for workers in regulated occupations (mostly licensed) and the policy implications. I also propose future research.

4.2 Literature Review

The United States has the highest internal labour mobility rate among developed countries.59 However, U.S. internal mobility has significantly declined since 1990 (Kaplan & Schulhofer-

59 Internal mobility: The United States. World Bank Report. Retrieved from http://siteresources.worldbank.org/ECAEXT/Resources/258598.../7383639.../10_us.pdf. Also seen in “America settles down, the Economist (2012), retrieved from http://www.economist.com/blogs/freeexchange/2012/07/labour-mobility 103

Wohl, 2012). Internal labour mobility in Canada is much lower than in the United States (1% vs. 2.4% of the total population annually). 60 The rate of interprovincial migration has fluctuated over the years, but overall it has remained relatively steady, ranging from 1.2% to 1.5% between 1994 and 2005 (Chen & Fougère, 2011).

Neoclassical economic theories imply that people will move to places where they can gain higher utility, such as through the greater economic gains that accrue to those who work in licensed occupations. However, professional associations may negatively affect labour mobility by creating entry barriers to practising the licensed occupations. Maurizi (1974) collects the number of new applicants and the number of licenses granted for 36 occupations by state governments, and reports that, in the United States, the pass rate for occupational licensing decreases when the demand for the services of these occupations increases. Most of the empirical research focuses more on the intercountry mobility barriers imposed by different licensing requirements, particularly the impact of restricting entry on skilled immigrants (see Federman, Harrington, & Krynski, 2006; and Peterson, Pandya, & Leblang, 2014, for the United States;61 see Banerjee & Phan, 2014; Brown, 1975; Girard & Smith, 2013; Gomez, Gunderson, Huang, & Zhang, 2015; Hall & Sadouzai, 2010; Jantzen, 2015; Warman, Sweetman, & Goldmann, 2015; and Zietsma, 2010, for Canada).62

60 America settles down, the Economist (2012), retrieved from http://www.economist.com/blogs/freeexchange/2012/07/labour-mobility 61 Peterson et al. (2014) argue that professional associations’ rent-seeking rationale creates “a barrier to the cross-national mobility of human capital” (p. 45). The authors find that the imposition of more-demanding licensing requirements on foreign-trained physicians leads to few new immigrant doctors. Federman et al. (2006) find that licensing laws limit the number of new market entrants. They also find that tightening licensing regulations could encourage immigrant practitioners to move to less-regulated states to search for new market opportunities. 62 Statistics Canada reports that more immigrants studied in the fields of regulated occupations in 2006, compared to Canadians overall (41% vs. 39%); in contrast, immigrants who studied in the regulated fields outside of Canada were significantly less likely to work in the professions for which they trained, compared to Canadian-educated immigrants and Canadian-born workers (24% vs. 53% vs. 62%, respectively) (Zietsma, 2010). Both Girard and Smith (2013) and Gomez et al. (2015) look at the regulated occupations across the labour market, using nationally representative data. Girard and Smith (2013) use cross-sectional data to provide a snapshot of entry barriers that immigrants face in Canada. Gomez et al.’s (2015) fixed-effects approach sheds some light on the impact of occupational licensing on individuals over a six-year period. However, they were unable to identify the beginning of the trajectory; hence, it was difficult to conclude that immigrants face entry 104

Licensing restrictions also vary across jurisdictions within a country. Taylor (2006) studies the interstate mobility of speech-language pathologists and audiologists triggered by hurricanes Katrina and Rita, and reports that 56% of the states allowed registered practitioners from another state to practise on a temporary basis; even lower percentages of states allowed out- of-state practitioners to continue to practise when submitting an application for a license. These small differences across geographic areas could be another factor limiting the mobility of licensed workers. Ladinsky (1967) finds a few individual and job characteristics that are associated with professional workers’ mobility decisions. For example, in addition to state licensing law requirements, strong professional networks and business demand for reasonable costs of capital and connecting with clientele, are correlated with low mobility rates across local labour markets because practitioners need to establish these resources from scratch after relocation.

The empirical evidence on the impact of occupational licensing on interjurisdictional migration tends to support the conclusion that occupational licensing decreases internal labour mobility. Earlier U.S. studies focus on a few most restrictive occupations such as the legal profession,

barriers because of barriers associated with occupational regulation. Banerjee and Phan (2014) find that skilled immigrants who worked in regulated occupations prior to migration struggle, both initially and over the long term, from the sustained negative consequences of not gaining licensing recognition in Canada. Jantzen’s (2015) study uses the 2011 National Household Survey and Immigration Landing File and finds that the most recent cohort of immigrants who were working in regulated occupations before they migrated have a better chance of working in corresponding occupations while living in Canada because of their higher human capital and more transferable skills, looking at five occupations, including civil, mechanical, and electrical and electronics engineering, financial auditing and accountancy, and nursing. Although the author does not discuss this point, it seems there is a possible correlation between the occupational matching rate and the tightness of occupational licensing—registered nurses had the highest matching rate, followed by engineers, and accountants had the lowest matching rates. It is difficult to draw a conclusion based on five occupations across three cohorts of immigrants. In addition, at first glance, the conclusion seems to conflict with the theory and the existing empirical research. This observation suggests that further studies on this issue are needed. Using a sampling of regulated occupations (17 used in a national comparison with an additional 20 in Ontario), Augustine (2015) shows that almost 75% of foreign-educated professionals do not work in the fields in which they were trained, but “the number of internationally educated license holders grows at roughly twice the rate of growth for those internationally educated and employed in a regulated profession (22.7% vs. 11.5%)” (Augustine, 2015, p. S35). Hall and Sadouzai (2010) compare immigrants who previously worked in high-tech occupations and those who worked in regulated occupations, and conclude that transferable skills and relatively loose institutional requirements enable high-tech workers to integrate more easily into the Canadian labour market. 105

dentistry, and medicine, and confirm that occupational licensing negatively impacts interstate mobility (Holen, 1965; Johnson & Kleiner, 2015; Kleiner et al., 1982; Pashigian, 1979; Pratt 1980). Holen (1965) shows that licensing requirements limit interstate mobility for dentists and lawyers. Pashigian (1979) studies the legal profession, using 1970 U.S. population census data, and finds that occupational licensing significantly decreases interstate mobility, and that limiting interstate reciprocity further reduces mobility. Kleiner et al. (1982) compare the immigration and outmigration of physicians, and conclude that differences in residency requirements and requirements to retake parts of qualification exams are correlated with lower mobility across states. The increasing trend in occupational licensing since the 1980s has been congruent with the decreasing trend in interstate migration over the same period. Pratt (1980) finds that the number of states that license a profession is negatively correlated with the labour mobility of the licensed profession’s members. Johnson and Kleiner (2015), looking at the interstate migration rates of lawyers, barbers, teachers, nurses, and dentists within the United States over the past sixty years, also report this relationship between occupational licensing and interstate migration. They find that four of the five occupations studied have a lower than average propensity to move out of the state.

More recent studies expand the occupation coverage and use more rigorous empirical methodologies to establish a more causal interpretation of occupational licensing and interstate mobility. These results tend to be less conclusive. Arbury et al. (2015) found that both teachers and teaching assistants were less likely than the general population to relocate to a different state. In addition, they applied the data on the reciprocity and endorsement of teachers’ licensing among states since 2000, and found that policies that aimed to lift barriers to labour mobility had a statistically significant impact on teachers’ interstate mobility, although the magnitude is very small. DePasquale and Strange (2016) examine the impact of the Nurse Licensure Compact (NLC), which permits reciprocity among states for practising registered nurses, on the labour mobility of nurses. Their study finds no increase in labour mobility or labour supply after the implementation of the NLC. Focusing on cosmetology, Zapletal (2014) concludes that states with lower occupational-licensing requirements have more stable workforces, with lower entry and exit rates, which contradicts the theoretical and empirical expectation that prospective practitioners will prefer to move away from states with more

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demanding licensing requirements and into states with lower licensing standards.63 Outside the United States, Prantl and Spitz-Oener (2009) find that the higher educational requirements included in entry regulations in states in East Germany significantly reduce labour mobility and engagement in self-employment among employees working in licensed occupations as compared to those in West Germany.

In Canada, some studies address occupational licensing as one of the many factors potentially affecting labour mobility (Gomez & Gunderson, 2007; Gunderson, 1994). Grady and Macmillan (2007) report that Canada has high labour mobility because of the low level of occupational regulations in all provinces, and suggest three major barriers that prevent professionals from moving across provincial boarders: “residency requirements; certain practices regarding occupational licensing, certification and registration; and differences in how occupational qualifications are recognized” (p. 2).

Despite the continuing interest in looking at the impact of occupational licensing on the individual or aggregate labour mobility, there is limited empirical evidence relating to this phenomenon in Canada. Brown (1975) examines occupational-licensure controls for the medical profession and finds that the majority of Canadian-born physicians stay where they were born and trained. In addition, Brown suggests that the more economically attractive the provinces are, the more stringent their licensing requirements. This further distorts physician allocation—foreign-trained doctors tend to move to the Atlantic Provinces, where licensing requirements are relatively looser than in Ontario. Using the 1971 Canadian census, Muzondo and Pazderka (1980) studied the interprovincial and international constraints that occupational licensing imposes on professionals that could potentially limit their geographic mobility. Referring to professions, they defined as “restrictive” laws enacted to require out-of-province candidates to retake licensing examinations or to take additional training under supervision when they want to gain the same license that they had already obtained elsewhere (Muzondo

63 Inconsistent occupational licensing requirements also have a spill-over effect into the rest of the labour market. Mulholland and Young (2016) find that individuals with low education are more likely to move to states that require fewer occupations to be licensed. 107

& Pazderka, 1980). However, the authors did not find any significant impact on interjurisdictional mobility. Chen and Fougère (2011) simultaneously examine interprovincial and interindustry mobility, and report that regulated workers (both regulated professions and employees in apprentice trades) are less mobile with respect to industry than are nonregulated workers but not so across provinces.64 The authors find that the interprovincial mobility rates for workers in regulated professions, apprenticeable trades, and nonregulated occupations were similar throughout the 1990s and early 2000s. To the best of my knowledge, this is the only recent empirical analysis that systematically addresses the impact of occupational licensing on internal mobility. Although Chen and Fougère defined interprovincial mobility by comparing the provinces of residence between two consecutive years, their probit models did not fully utilize the longitudinal data structure. Also, the ambiguous ways in which the regulated occupations are defined in Canada make it difficult to utilize the research findings in studies of other countries, including the United States.65

Public policies have a strong influence on internal migration in Canada. Day and Winer (2012) provide a comprehensive review of the impact of public policy on interprovincial migration in Canada. Between 1990s and early 2000, studies using micro data report that public policies such as unemployment insurance benefits, property taxes, or income subsidies are significantly related to an individuals’ move-stay decision. Occupational licensing was not examined as a public policy that could induce internal mobility.

Other factors that contribute to interprovincial mobility have been examined in the Canadian context. Individual characteristics such age, gender, marital status, educational attainment, immigrant status, primary language, union status, and occupations are predictors of an individual’s mobility decision (e.g., Bernard, Finnie, & St-Jean, 2008; Chen & Fougère, 2011; Finnie, 2004; Ostrovsky, Hou, & Picot, 2011; Robinson & Tomes, 1982). In addition, studies

64 The authors also find that unionized workers are less likely to move either across industries or between provinces. 65 See detailed discussion in the Appendix chapter. 108

using aggregate migration rates suggest that local labour market performance, such as regional unemployment rates, significantly affects migration within Canada.66

4.3 Theoretical Framework

In theory, licensing restricts professionals’ labour mobility in three ways. First, when a previously licensed professional practitioner moves to a new jurisdiction, he or she must be relicensed. The relicensing process typically involves getting their educational credentials verified, taking additional courses or professional development workshops, retaking qualification exams, and/or working for a period of time under the supervision of a licensed worker in the host province. These processes increase the opportunity cost of internal migration, creating barriers to professionals. Also, licensed practitioners who barely passed their entry requirements may be reluctant to move to a jurisdiction with higher licensing standards for fear that they might not be able to become licensed there and would lose the chance to practise.

Second, additional cost may occur when a licensed professional moves to a different jurisdiction from where he or she was initially licensed. As discussed in Ladinsky (1967), licensed workers might need to rebuild their professional networks, and compete with the incumbent licensed workers for clientele after relocation. It would not be economically beneficial for licensed workers to relocate because migrating to a different province requires specific investment in the local market.

Third, assuming tightening licensing standards will lead to higher economic benefits for the incumbents, licensed workers would not want to move to a jurisdiction with lower licensing requirements because the rent captured by occupational-licensing status is lower. In sum, the internal geographical labour mobility decision is heavily influenced by the net gains of moving, in particular, the costs of getting relicensed in the new jurisdiction. If the professionals do not need to relicense across jurisdictions, a barrier to mobility is removed because the costs

66 See detailed discussion in Day and Winer (2012). 109

are lower. This paper aims to empirically examine whether or not licensed workers are less likely to move across provinces.

4.4 Research Methods and Data

4.4.1 Data

In this research, I use five waves of the Canadian Survey of Labour and Income Dynamics (SLID). The SLID is a nationally representative longitudinal dataset collected by Statistics Canada. It uses the same sampling framework as the Canadian Labour Force Survey, and samples households across all 10 Canadian provinces. The survey asks individuals in the study groups to respond to a constant set of questions each year for six consecutive years. These questions cover individual educational attainment, major life events, financial position, labour market outcomes, and mobility during the survey year. The first wave of data was collected between 1993 and 1998, and the fifth wave was collected between 2005 and 2010. There were three overlapping years between the two consecutive waves, and so the five waves of data cover 18 years in total. Since younger workers are more likely to move across geographical regions (e.g., Finnie, 2004), I restricted my sample to paid workers between ages 21 and 34 who reported both their province of residence and their occupation. The pooled sample contains 40,000 individual survey respondents and 128,800 observations, in total.67

To answer the question, I focused on interprovincial mobility and constructed a variable of individual interprovincial mobility. The longitudinal survey asked the respondents about the province of residence annually. By comparing the province in the focal year and the year after, I identified individuals who moved between two consecutive years. The dependent variable 퐼(푚표푣푒푑) is coded as one if the survey respondent moved the next year, and zero otherwise.

The independent variables are constructed using the data collected in the year before the year of possible relocation. The key independent variables are occupational licensing and union

67 Numbers are rounded to meet the disclosure requirement of the Statistics Canada Research Data Centre. 110

status. Union status is a dummy variable equal to one if the individual reported being a union member or working in a workplace that is covered by a collective agreement, and zero otherwise. Governments in Canada have neither designed nor conducted a population survey that asks questions on occupational licensing status. For this study, licensing status is constructed based on the exclusive right-to-practice clause enacted in provincial legislation. As discussed in the Appendix chapter, I constructed such data. Licensing status is a binary variable that is coded as one if the worker was in an occupation with the exclusive right-to- practice, and zero if they were not. I linked the licensing status to the SLID by coding the licensing status of an occupation in a specific province equal to one if there exists a right-to- practice licensing law, and zero otherwise.68

Table 4.1 shows the pooled descriptive statistics. Pooling together 18 years of data, the percentage of survey respondents who were licensed workers was 6.6% (SD=0.248), and 16.6% (SD=0.372) of the Canadians aged 21 to 34 surveyed were either union members or covered by a collective agreement. Those who reported working part-time amounted to 10.8% (SD=0.310); 30.4% (SD=0.460) reported holding down a permanent job; and 9.4% (SD=0.292) held multiple jobs. Approximately 2.4% (SD=0.152) of the survey respondents reported moving to a different province in the next year. The mean local labour market unemployment rate is 8.46% (SD=2.98). ------Insert Table 4.1 about here ------

There was not much of difference between the share of male workers and female workers who moved across provincial boundaries (2.5% (SD=0.155) and 2.3% (SD=0.149), respectively). Regardless of whether or not they moved, the percentage of workers who are licensed was higher for males than females (6.9% (SD=0.254) and 6.3% (SD=0.243), respectively), and there was the same percentage of male workers who reported to be covered by the collective

68 A certification dummy is constructed in the same fashion for occupations with title protection only. However, the certification variable is not the focus of the current study. 111

agreement as female workers (16.6% (SD=0.372) and 16.6% (SD=0.372), respectively). Female workers reported being more likely to hold a bachelor’s or higher degree, compared to male workers who were more likely to have high school or lower educational attainment. Female workers are more likely to work in jobs with lower quality in the year before moving to a different province; specifically, they are more likely to work part-time, and hold multiple jobs at the same time; while less likely to hold a permanent job. In addition, male workers had more full-year equivalent work experience than women (6.4 years (SD=3.8) and 5.5 years (SD=3.5), respectively).

4.4.2 Empirical methodology

This study first uses multilevel modelling (MLM) to investigate the impact of occupational licensing on individual labour mobility. As previously discussed, the SLID data provides repeated measures of responses. Multilevel modelling allows the covariates’ structure to be based on the distance (in year in the SLID), which makes the MLM preferable because it takes the over-dispersion of auto-correlated data into consideration. MLM is also more flexible because it allows the estimation of inter-individual differences in intra-individual change over time. MLM can include both random- and fixed-effects estimation: it allows random effects of the intercept, and fixed-effects estimators are derived to investigate the impact of changes in an individual’s status (either unionization or licensing) on the changes in individuals’ likelihood to move to a different province next year.69 Identification is obtained from changes in the individual’s occupational-licensing status and union status. For comparison purposes, logit specifications with random effects and pooled OLS models, and a linear probability model with robust standard errors clustered at the combination of provincial and occupational

69 Typically, fixed-effects models are used to examine the impact of changes in independent variables on changes in dependent variables. However, if the subjects examined change little, or not at all across time, fixed- effects may not work when the within-subject variability is not enough. The random-effects models are more likely to be biased compared to the fix-effects models because the fixed-effects models control for unobserved characteristics that are potentially correlated with the dependent variables. In this research, random-effects models are more appropriate, and to control for a certain degree of omitted variable biases, multilevel modelling is adopted. 112

levels are also estimated.70 The multilevel model is:

퐼(푚표푣푒푑 푡표 푎 푑푖푓푓푒푟푒푛푡 푝푟표푣푖푛푐푒) 푖푡 = 훼 + 훾 퐿푖푐푒푛푠푖푛푔 푖(푡−1) + 훽 푈푛푖표푛 푖(푡−1) + ′ 훿 푋 푖(푡−1) + 푣 푖 + 휖 푖푡 (1),

The dependent variable 퐼(푚표푣푒푑 푡표 푎 푑푖푓푓푒푟푒푛푡 푝푟표푣푖푛푐푒) is a dummy variable that captures interprovincial mobility. Independent variables are measures in the survey year before relocation. 퐿푖푐푒푛푠푖푛푔 indicates whether or not the respondent was a licensed worker (both licensed professionals and workers in compulsory trades); 푈푛푖표푛 indicates whether the worker was a union member or covered by a collective agreement; and X represents a set of conventional personal, human capital, and job characteristics that can affect interprovincial relocation. The variables in X include basic characteristics of respondents: age; gender; education level; marital status; province of residence; years of work experience; dummy variables for the relevant industries (excludes manufacturing as the reference category); dummy variables for the relevant occupations (excludes management as the reference group); three indicators each for multiple job holders, permanent job holders, and part-time workers; and a categorical variable for the number of workers in the workplace.

The estimated coefficient γ̂ denotes the within-group estimate of occupational licensing on interprovincial movement. β̂ is the estimated coefficient of the fixed-effects component in the multilevel model specifications; it represents the average within-group mean estimate of the impact of union membership on labour mobility. These estimated coefficients indicate the average impact of changes in occupational licensing status and union status on interprovincial mobility across provinces in Canada.

Table 4.2 presents the four model specifications. Model [1] provides estimates of the raw relationship between mobility and the main predictors (licensing and union status). In model [2], I added individual characteristics as controls, such as gender, education level, marital status, and region of residence. Model [3] further controls for job characteristics: holding

70 See estimated results in the Appendix 4A. 113

multiple jobs, part-time versus full-time employment, holding a permanent job, aggregate categories for industries, aggregate occupational classification, and categories of plant size. Model [4] includes a control for local economic condition. The provincial unemployment rates are used as a proxy for the local economic conditions. The preferred specification is model [4]. Linear probability models with clustered standard errors, random-effects Logit models, and Logit models are also presented to provide comparative results that may reflect the influence of errors that correlate with time.71

4.5 Results

Table 4.2 reports the marginal effects of multilevel modelling estimated coefficients. When the dependent variable is binary, the estimated coefficients of the models can be interpreted as the probability of an individual belonging to a particular group (either one or zero). The marginal effects, in this case, show that the change in the probability of moving (i.e., 퐼(푚표푣푒푑 푡표 푎 푑푖푓푓푒푟푒푛푡 푝푟표푣푖푛푐푒) = 1) when the binary independent variable changes from 0 to 1, holding all other variables at their means. The estimated coefficients of licensing status are positive but insignificant, which suggests that whether or not a worker is licensed as either a licensed professional or in a compulsory trade, is not a predictor of the decision to move cross provincial boundaries. The estimated coefficients of certified profession status are negative, but insignificant in all specifications. The marginal effects estimates of the coefficients of union status are significantly negative, showing that union members are less likely to move to a different province. Union members are 1.1% less likely to move across a provincial boundary, holding other variables constant.

------Insert Table 4.2 about here ------

71 I ran the probit and linear probability models as well. The estimated coefficients are similar to the logit results, although licensing status was negatively associated with internal mobility at a 5% significance level. The results can be provided upon request. 114

Despite the unexpected results in licensing and certification, the estimated coefficients of individual characteristics are in line with the existing literature, with factors including gender, age, and marital status significantly predicting an individual’s interprovincial mobility. Estimates of age, gender, and marital status are all negative and significant, suggesting that: female workers are less likely to relocate than male workers; the older the worker, the less likely they are to relocate; and married individuals are less likely than singles to migrate internally. However, the effect sizes of these variables are minimal. Educational attainment is associated with this decision as well. An individual with higher education, in particular with a bachelor’s degree or above, is 2% more likely to migrate internally than a worker with high school or lower education.

The job-related characteristics played a minor role in individual’s mobility decisions. Workers who held a permanent job are less likely to move, compared to their counterparts. Work experience is also negatively related to relocation. In addition, workers holding a job in a large establishment (i.e., more than 500 employees) are more likely to migrate across provinces. Again, the effect sizes are very small. Other job-related characteristics appear to have an insignificant relationship with a probability of relocating to a different province.

Consistent with the existing literature, local economic condition is significantly associated with the individual’s mobility decision. Workers living in areas occurring higher unemployment rates are more likely to move out of the province.

Table 4.3 reports the marginal-effects estimates of licensing status on interprovincial mobility by gender. The results suggest that occupational licensing has no relationship with an individuals’ decision whether or not to move. The results of unionization for both male and female workers are similar to the overall sample except that the magnitudes of estimated coefficients for male workers are much higher than for female workers. Male union employees are 1.4% less likely to move, compared to their nonunionized counterparts; women who are union members are only 0.9% less likely to relocate. These suggest that union status affects male workers’ decisions to relocate more than those of female workers.

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------Insert Table 4.3 about here ------

To summarize, licensed workers are no less likely to move to a different province than are unlicensed workers. In contrast, unionized workers are significantly less likely to relocate than nonunionized workers. The impacts are consistent for both male and female workers, but the magnitudes are different.

4.6 Discussion and Conclusion

One of the major concerns in theory is the inconsistent licensing standards across jurisdictions, which increase the costs of relocation and decrease the benefits of practising in licensed occupations. My results differ from most of the empirical studies on the issue of occupational licensing and labour mobility, which mostly focus on the United States. For Canadians aged 21 to 34, employees in a licensed occupation are not less likely to move across provinces than employees in a nonlicensed occupation. Individual characteristics, such as having a bachelor’s or higher degree, and being female, a younger age, or single, in general, increase the probability of an individual moving to a different province. In addition, job characteristics, such as being covered by a collective agreement, having a permanent job or having more work experience are all observed to be significantly negatively correlated with mobility measures. I take union status, permanent job status, and work experience to be proxies of job security and speculate that individuals who have a more secure job will be reluctant to move to a different province.

There are no gender differences between male and female workers regarding the impact of occupational licensing on interprovincial mobility. In general, both male and female workers who are union members or covered by a collective agreement or who hold a permanent position are less likely to move, but male workers are impacted more than female workers.

Empirical studies on the impact of occupational licensing on labour mobility are limited because of the difficulties in gathering the relevant data. Researchers face two major challenges. First, direct measures of occupational licensing have been lacking until recently.

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In the United States, questions about real individual licensing statuses have been added to population survey questionnaires since 2008 (Kleiner & Krueger, 2013). No such direct questions were included in any Canadian population survey. To add to the complexity of the situation, the Canadian definition of a regulated occupation is very broad and may not reflect licensed occupations. Although the data I collected using the provincial occupational licensing legislative provisions offers a more precise interpretation of the licensing phenomena of the past three decades, there are still issues with this approach. My list of licensed occupations may not capture other de facto licensed occupations. For example, dietitians are not licensed in every province, and dietitians in many provinces only have the exclusive right-to-title. However, the majority of dietitians are hired by hospitals, and employers like hospitals may be able to detect quality of services and hence require only the right-to-title.

Another issue is that legislative support can only go so far to legitimize professional organizations’ efforts to close the market. The power of market closure varies both across and within actual labour markets. One example is barbering. Although market control is achieved through employment in barber shops, unlicensed practitioners can provide services in the informal market without customers objecting to differences in the quality of service. Last but not least, there is a delay between the passing of legislation and the time it comes into effect. Although only a few occupations have obtained licensing status since the 1990s, and most of these occupations had already more or less closed the market before they received legislative support, there is typically a delay between the passing of a piece of legislation and the time an individual faces its impact. In a couple of extreme cases, there was an additional delay between the passage and proclamation of law. In sum, the licensing status used here is a proxy that comes with biases. However, I would argue that this approach is a reasonable starting point for looking into this question. Further investigation of the actual licensing status is surely needed.

The second challenge to labour mobility research is that the data requirements are very high. It is difficult to construct longitudinal data that tracks individuals over time, let alone to focus on a population that moves across provinces. My research indicates that less than 3% of Canadians move across provincial boundaries. Previous studies also report low interprovincial

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mobility (e.g., Chen & Fougère, 2011; Osberg, Gordon, & Lin, 1994).72 The limited mobility typically leads to very small sample size in sampled population survey.

An individual’s decision to move to a different province is very complex. The reduced forms of hypothesis-testing models cannot capture the full essence of this relationship. One of the limitations of the empirical approach is the endogeneity problem. There are potential confounding variables that challenge this research, such as provincial economic conditions. Licensing bodies tend to have higher entry standards in the provinces that are experiencing better economic performance. The estimates also confirm that individuals are more likely to move if the local unemployment is higher. The decision to work in a licensed occupation is not exogenous—individuals tend to choose to work in a licensed occupation because of the higher economic gains, and they move to provinces with more robust economies. The current research is an exploratory attempt to understand the relationship between occupational licensing status and individual interprovincial mobility; it does not claim causality.

In Canada, policy-makers are proactive in permitting citizens to move freely across provinces. Chapter 7 of the Agreement on Internal Trade (AIT) was enacted to counter the negative impact of occupational licensing on interprovincial labour mobility through cooperation between the federal government and all provinces and territories. In 2009, it was announced in the AIT that, “the revised Labour Mobility chapter of the AIT will enable any worker certified for an occupation by a regulatory authority of one Party to be recognized as qualified to practice that occupation by all Parties.”73 The Agreement particularly focuses on two crucial mechanisms of occupational licensing. First, the Agreement provides legislative guidance to professional associations for removing residency requirements in their licensing standards. Second, the government encourages regulatory bodies to establish mutual-recognition clauses to ease licensing barriers across provinces.

72 Chen and Fougère (2011) found that less than 1% of Canadians relocated to a different province between 1995 and 2005, and significant variations of provincial inflow and outflow migration are also observed for the period. Osberg et al. (1994) illustrate that the net flow of labour mobility across Canadian provinces for this period ranged from -1.8% to 0.36%. 73 http://www.ait-aci.ca/wp-content/pdfs/English/Progress/fmm_en.pdf 118

The Competition Bureau of Canada (2007) argued that several of the most regulated professions are considered to have very low productivity per hours worked and suggested relaxing occupational regulations. Their assessment of the impact of labour mobility is positive as mutual recognition agreements have been signed in most provinces, and geographical restrictions have been replaced by reciprocity clauses in provincial occupational regulation acts. In addition, many professional associations have established alternative assessment methods so as to decrease the complexity of the licensing process. Among the five professions the Competition Bureau investigated (i.e., accountants, lawyers, optometrists, pharmacists, and real estate agents), only pharmacy still adopts province-based professional evaluation processes and entry standards without implementing any reciprocity or endorsement mechanisms. The Bureau criticizes this variation among provinces because pharmacists perform essentially the same duties and responsibilities, regardless of their geographic location within Canada. 74 The Bureau recommends encouraging higher competition by loosening occupational regulations.

Unfortunately, the data utilized in the current research was not able to identify whether or not labour mobility changed after the implementation of the AIT. Lome (2011) compares the mobility and retention of registered nurses before and after the implementation of the AIT, and finds no significant difference. I found no additional negative consequences to licensed workers after having moved across provincial boundaries. However, I have not yet been able to answer whether these results are the outcomes of the AIT or whether there were no penalties imposed by variations in licensing requirements across provinces. Therefore, it will be crucial for future research to assess the impacts of the AIT and to determine whether or not the adverse impacts of occupational licensing could be eliminated.

My research provides empirical support for the impacts of labour market institutions, such as occupational licensing and unionization in Canada, on internal migration using micro data. I used a national longitudinal data to investigate whether or not licensing status and union

74 http://www.competitionbureau.gc.ca/eic/site/cb-bc.nsf/eng/03407.html 119

membership can affect workers’ decisions to move across provincial boundaries. Multilevel modelling estimates show that working in a licensed occupation does not restrict a worker to migrate internally in Canada; however, workers covered by a collective agreement are significantly less likely to relocate to a different province. The results also hold for both male and female workers. Taking the other significant predictors found in this study into consideration, I conclude that individuals’ move-stay decisions are very much a rational analysis by weighing the cost and benefits of moving—being older, married, having longer tenure, or currently holding a permanent job will increase the cost of internal migration, while higher human capital will raise the benefits of relocation.

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4.7 References Arbury, C., Bonilla, G., Durfee, T., Johnson, M., & Lehninger, R. (2015). The ABCs of regulation: The effects of occupational licensing and migration among teachers. Hubert H. Humphrey School of Public Affairs. Retrieved from the University of Minnesota Digital Conservancy, http://hdl.handle.net/11299/170743.

Augustine, J. (2015). Employment match rates in the regulated professions: Trends and policy implications. Canadian Public Policy, 41 (Supplement 1), S28–S47.

Banerjee, R., & Phan, M. (2014). Licensing requirements and occupational mobility among highly skilled new immigrants in Canada. Relations industrielles/ Industrial Relations, 69(2), 290–315.

Bernard, A., Finnie, R., & St-Jean, B. (2008). Interprovincial mobility and earnings. Perspectives on Labour and Income, 20(4), 27–37.

Brown, M. C. (1975). Some effects of physician licensing requirements on medical manpower flows in Canada. Relations industrielles/Industrial Relations, 30(3), 436–451.

Chen, X., & Fougère, M. (2011). Regulated occupations, immigration, and labour mobility in Canada, 1994–2005. Canadian Journal of Regional Science, 34(1), 33–43.

Competition Bureau of Canada. (2007). Self-regulated professions: Balancing competition and regulation. Gatineau, QC: Government of Canada.

Day, K. M., & Winer, S. (2012). Interregional migration and public policy in Canada: An empirical study (Vol. 223). McGill-Queen's Press-MQUP.

DePasquale, C., & Strange, K. (2016). Labor supply effects of occupational regulation: Evidence from the nurse licensure compact (No. w22344). National Bureau of Economic Research.

Federman, M. N., Harrington, D. E., & Krynski, K. J. (2006). Vietnamese manicurists: are immigrants displacing natives or finding new nails to polish?. Industrial & Labor Relations Review, 59(2), 302–318.

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Finnie, R. (2004). Who moves? A logit model analysis of inter-provincial migration in Canada. Applied Economics, 36(16), 1759–1779.

Girard, M., & Smith, M. (2013). Working in a regulated occupation in Canada: An immigrant– native born comparison. Journal of International Migration and Integration, 14(2), 219– 244.

Gomez, R., & Gunderson, M. (2007). Barriers to the inter-provincial mobility of labour. Working paper series. Ottawa: Industry Canada.

Gomez, R., Gunderson, M., Huang, X., & Zhang, T. (2015). Do immigrants gain or lose by occupational licensing? Canadian Public Policy, 41(Supplement 1), S80–S97.

Grady, P., & Macmillan, K. (2007). Interprovincial barriers to labour mobility in Canada: Policy, knowledge gaps and research issues. Ottawa: Industry Canada.

Gunderson, M. (1994). Barriers to interprovincial labour mobility. Interprovincial Trade Wars: Why the Blockade Must End, Vancouver: Fraser Institute, 131–154.

Hall, P. V., & Sadouzai, T. (2010). The value of “experience” and the labour market entry of new immigrants to Canada. Canadian Public Policy, 36(2), 181–198.

Holen, A. S. (1965). Effects of professional licensing arrangements on interstate labor mobility and resource allocation. Journal of Political Economy, 73(5), 492–498.

Johnson, J., & Kleiner, M. M. (2015). Does occupational licensing reduce interstate migration? Kalamazoo, MI: W. E. Upjohn Institute for Employment Research.

Kaplan, G., & Schulhofer-Wohl, S. (2012). Understanding the long-run decline in interstate migration (No. w18507). National Bureau of Economic Research.

Kleiner, M. M. (2006). Licensing occupations: Ensuring quality or restricting competition? Kalamazoo, MI: W. E. Upjohn Institute.

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Kleiner, M. M. (2013). Stages of occupational regulation: Analysis of case studies. Kalamazoo, MI: W. E. Upjohn Institute.

Kleiner, M. M. (2015). Border battles: The influence of occupational licensing on interstate migration. Employment Research Newsletter, 22(4), 2.

Kleiner, M. M., Gay, R. S., & Greene, K. (1982). Barriers to labor migration: The case of occupational licensing. Industrial Relations, 21(3), 383–391.

Kleiner, M. M., & Krueger, A. B. (2013). Analyzing the extent and influence of occupational licensing on the labor market. Journal of Labor Economics, 31(S1), S173–S202.

Ladinsky, J. (1967). Occupational determinants of geographic mobility among professional workers. American Sociological Review, 32(2), 253–264.

Lome, Z. O. (2011). Influence of education portability and geographical mobility on nursing retention in Canada. The Journal of Global Health Care Systems, 1(2), 1–15. ISSN 2159- 6743.

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Mulholland, S. E., & Young, A. T. (2016). Occupational licensing and interstate migration. Cato Journal, 36(1), 17–31.

Muzondo, T. R., & Pazderka, B. (1980). Occupational licensing and professional incomes in Canada. Canadian Journal of Economics, 13(4), 659–667.

Osberg, L., Gordon, D., & Lin, Z. (1994). Interregional migration and interindustry labour mobility in Canada: A simultaneous approach. Canadian Journal of Economics, 27(1), 58– 80.

Ostrovsky, Y., Hou, F., & Picot, G. (2011). Do immigrants respond to regional labor demand shocks? Growth and Change, 42(1), 23–47.

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Peterson, B. D., Pandya, S. S., & Leblang, D. (2014). Doctors with borders: Occupational licensing as an implicit barrier to high skill migration. Public Choice, 160(1-2), 45–63.

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Zapletal, M. (2014). Three Essays on Regulation and Entrepreneurship. Doctoral dissertation, University of Michigan, Ann Arbor, MI.

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Table 4.1 Descriptive Statistics for Pooled Sample

Pooled sample Male Female Variables Mean Std Dev Mean Std Dev Mean Std Dev I(moved a year after) 0.024 [0.152] 0.025 [0.155] 0.023 [0.149]

Licensed occupations 0.066 [0.248] 0.069 [0.254] 0.063 [0.243] Certified professions 0.013 [0.114] 0.010 [0.101] 0.016 [0.126] Union members 0.166 [0.372] 0.166 [0.372] 0.166 [0.372]

Female 0.517 [0.500] Age (years) 27.7 [4.0] 27.6 [4.0] 27.7 [4.0] Married 0.463 [0.499] 0.412 [0.492] 0.512 [0.500] Education High school or less 0.408 [0.491] 0.438 [0.496] 0.379 [0.485] College 0.295 [0.456] 0.279 [0.449] 0.310 [0.462] Bachelor’s degree and 0.154 [0.361] 0.131 [0.337] 0.176 [0.381] above

Experience (years) 5.9 [3.7] 6.4 [3.8] 5.5 [3.5] Multiple-job holder 0.094 [0.292] 0.086 [0.280] 0.102 [0.303] Part-time status 0.108 [0.310] 0.054 [0.225] 0.158 [0.365] Permanent job status 0.304 [0.460] 0.312 [0.463] 0.296 [0.457]

Provincial unemployment rate 8.458 [2.978] 8.452 [2.986] 8.464 [2.971] # observations 157,800 76,300 81,600 Note: The number of observations and descriptive statistics reflect the information collected between 1999 and 2010.

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Table 4.2 Estimated Marginal Effects of the Likelihood of Moving to Another Province in a Year (1993–2010): Licensed Occupations

Dv: I(moved a year after) MLM[1] MLM [2] MLM [3] MLM [4] Licensed occupations (both 0.107 0.067 0.034 0.038 licensed professions and compulsory trades) (0.092) (0.090) (0.104) (0.104) Union -0.526*** -0.414*** -0.456*** -0.472*** (0.070) (0.069) (0.078) (0.078) Certified professions -0.044 -0.122 -0.138 -0.129 (0.186) (0.181) (0.191) (0.191) College degree 0.081 0.073 0.071 (0.054) (0.056) (0.056) Bachelor’s degree + 0.939*** 0.835*** 0.845*** (0.059) (0.064) (0.064) Age (years) -0.088*** -0.074*** -0.074*** (0.006) (0.007) (0.007) I(married) -0.328*** -0.296*** -0.293*** (0.049) (0.051) (0.050) I(multiple-job holder) 0.011 -0.004 (0.073) (0.073) I(permanent job) -0.319** -0.309*** (0.079) (0.079) I(part-timer) -0.078 -0.079 (0.075) (0.074) Experience (years) -0.028*** -0.027*** (full-time equivalence) (0.009) (0.009) Plant size (21–99) 0.015 0.018 (0.061) (0.061) Plant size (100–499) 0.019 0.028 (0.078) (0.078) Plant size (500+) 0.351*** 0.358*** (0.121) (0.120) Females -0.117*** -0.092* -0.087* (0.045) (0.049) (0.049) 0.070*** Provincial unemployment rate (0.011) Industry N N Y Y Occupation N N Y Y # observations 157,800 157,800 157,800 157,800 # individuals 40,000 40,000 40,000 40,000 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Robust standard errors in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as regional dummies, categories of industry, and one-digit occupation. The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement. 126

Table 4.3 Estimated Marginal Effects of the Likelihood of Moving to Another Province in a Year (1993–2010): Licensed Occupations, by Gender

Male Female Dv: I(moved a year after) MLM[1] MLM[4] MLM[1] MLM[4] Licensed occupations (both 0.132 0.142 0.044 -0.064 licensed professions and (0.122) (0.137) (0.0140) (0.161) compulsory trades) Union -0.652*** -0.548*** -0.394*** -0.370*** (0.100) (0.108) (0.100) (0.113) Certified professions -0.229 -0.460 -0.067 0.055 (0.308) (0.315) (0.235) (0.242) College degree 0.088 0.061 (0.079) (0.080) Bachelor’s degree + 0.796*** 0.870*** (0.095) (0.087) Age -0.070*** -0.078*** (0.010) (0.010) I(married) -0.217*** -0.361*** (0.074) (0.069) I(multiple job holder) 0.039 -0.040 (0.107) (0.100) I(permanent job) -0.289*** -0.324*** (0.112) (0.111) I(part-timer) 0.052 -0.137 (0.128) (0.093) Experience (years) -0.034*** 0.018 (full-time equivalence) (0.012) (0.013) Provincial unemployment rate 0.063*** 0.078*** (0.016) (0.016)

Industry N Y N Y Occupations N Y N Y # observations 76,300 76,300 81,600 81,600 # individuals 19,400 19,400 20,600 20,600 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Robust standard errors in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as provincial dummies, categories of industry, and one-digit occupation. The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993– 2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement.

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Appendix 4A Table 4A.1 Estimated Marginal Effects of Logit Estimates of Probability of Moving to Another Province in a Year (1993-2010): Licensed Occupations Dv: I(moved a year after) Logit[1] Logit[2] Logit[3] Logit[4] Licensed occupations (both licensed 0.003* 0.001 0.001 0.001 professions and compulsory trades) (0.002) (0.001) (0.002) (0.002) Union -0.010*** -0.007*** -0.007*** -0.008*** (0.001) (0.001) (0.001) (0.001) Certified professions 0.001 -0.001 -0.001 -0.001 (0.004) (0.003) (0.003) (0.003) College degree 0.001 0.001* 0.001 (0.001) (0.001) (0.001) Bachelor’s degree + 0.019*** 0.016*** 0.016*** (0.001) (0.001) (0.001) Age (years) -0.001*** -0.001*** -0.001*** (0.000) (0.000) (0.000) I(married) -0.006*** -0.004*** -0.004*** (0.001) (0.001) (0.001) I(multiple-job holder) 0.001 -0.001 (0.001) (0.001) I(permanent job) -0.005*** -0.005*** (0.001) (0.001) I(part-timer) -0.002** -0.002** (0.001) (0.001) Experience (years) 0.001*** -0.001** (full-time equivalence) (0.000) (0.000) Plant size (21–99) 0.001 0.001 (0.001) (0.001) Plant size (100–499) 0.001 0.001 (0.001) (0.001) Plant size (500+) 0.007*** 0.007*** (0.002) (0.002) Females -0.002*** -0.002** -0.001** (0.001) (0.001) (0.001) Provincial unemployment rate 0.001*** (0.000) Industry N N Y Y Occupation N N Y Y # observations 157,800 157,800 157,800 157,800 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as provincial dummies, categories of industry, and one-digit occupation). The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada–Research Data Centre requirement.

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Table 4A.2 Estimated Marginal Effects of Random-effects Models of Probability of Moving to Another Province in a Year (1993-2010): Licensed Occupations

Dv: I(moved a year after) RE[1] RE[2] RE[3] RE[4]

Licensed occupations (both licensed 0.107 0.060 0.034 0.038 professions and compulsory trades) (0.092) (0.093) (0.104) (0.104) Union -0.526*** -0.398*** -0.456*** -0.472*** (0.186) (0.071) (0.078) (0.078) Certified professions -0.045 -0.123 -0.138 -0.129 (0.186) (0.186) (0.191) (0.191) College degree 0.040 0.073 0.071 (0.055) (0.056) (0.056) Bachelor’s degree + 0.888*** 0.836*** 0.846*** (0.060) (0.064) (0.064) Age (years) -0.085*** -0.074*** -0.074*** (0.006) (0.007) (0.007) I(married) -0.380*** -0.296*** -0.293*** (0.050) (0.051) (0.050) I(multiple-job holder) -0.011 -0.004 (0.073) (0.073) I(permanent job) -0.319*** -0.309*** (0.079) (0.079) I(part-timer) -0.078 -0.079 (0.075) (0.074) Experience (years) -0.028*** -0.027*** (full-time equivalence) (0.009) (0.009) Plant size (21-99) 0.015 0.018 (0.061) (0.061) Plant size (100-499) 0.019 0.028 (0.078) (0.078) Plant size (500+) 0.351*** 0.359*** (0.121) (0.121) Female -0.101** -0.092* -0.087* (0.046) (0.049) (0.049) 0.070*** Provincial unemployment rate (0.011) Industry N N Y Y Occupation N N Y Y # observations 157,800 157,800 157,800 157,800 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as provincial dummies, categories of industry, and one-digit occupation). The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement.

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Table 4A.3 Logit and Random-Effects (FE) Estimates of Probability of Moving to Another Province in a Year (1993–2010), by Gender

Random-effects Logit Male Female Male Female Dv: I(moved a year after) RE[1] RE[4] RE[1] RE[4] Logit[1] Logit[4] Logit[1] Logit[4] Licensed occupations 0.003 0.002 0.002 0.000 0.132 0.142 0.044 -0.064 (both licensed professions (0.003) (0.002) (0.003) (0.002) (0.123) (0.137) (0.140) (0.162) and compulsory trades) Union -0.012*** -0.009*** -0.008*** -0.006*** -0.652*** -0.548*** -0.393*** -0.370*** (0.001) (0.001) (0.001) (0.001) (0.100) (0.108) (0.100) (0.113) Certified professions -0.003 -0.005* -0.003 0.003 -0.229 -0.460 -0.066 0.055 (0.005) (0.003) (0.005) (0.004) (0.308) (0.315) (0.393) (0.242)

College degree 0.001 0.001 0.088 0.061 (0.001) (0.001) (0.079) (0.080) Bachelor’s degree + 0.015*** 0.017*** 0.796*** 0.871*** (0.002) (0.002) (0.095) (0.087) Age -0.001*** -0.001*** -0.070*** -0.078*** (0.000) (0.000) (0.010) (0.010) I(married) -0.003*** -0.005*** -0.217*** -0.361*** (0.001) (0.001) (0.074) (0.069) I(multiple-job holder) 0.000 -0.001 0.039 -0.040 (0.002) (0.001) (0.107) (0.100) I(permanent job) -0.005*** -0.005*** -0.289** -0.324*** (0.002) (0.001) (0.112) (0.111) I(part-timer) -0.001 -0.003** 0.052 -0.137 (0.002) (0.001) (0.128) (0.093) Experience (years) -0.001*** 0.000*** -0.034*** -0.018 (full-time equivalence) (0.000) (0.000) (0.012) (0.013)

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Provincial unemployment -0.001*** 0.001 0.063*** 0.078** rate (0.000) (0.000) (0.016) (0.016) Industry N Y N Y N Y N Y Occupations N Y N Y N Y N Y # observations 76,300 76,300 81,600 81,600 76,300 76,300 81,600 81,600 # individuals 19,400 19,400 20,600 20,600 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Weighted regression models. Robust standard errors in parentheses clustered at the occupation at provincial levels. Occupational licensing is a step function equal to zero if the individual works in a licensed occupation. Each regression includes a constant as well as province of residence, level of plant size, categories of industry, and one-digit occupation). The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement.

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Table 4A.4 Estimated Effects of the Likelihood of Moving to Another Province in a Year (1993–2010): Licensed Occupations

Dv: I(moved to a different provinces)t LPM[1] LPM [2] LPM [3] LPM [4] I(moved to a different provinces)t-1 0.126*** 0.118*** 0.118*** 0.118*** (0.007) (0.007) (0.007) (0.007) Licensed occupations(both licensed 0.000 -0.001 0.000 0.000 professions and compulsory trades) t-1 (0.002) (0.002) (0.002) (0.002) Union t-1 -0.007*** -0.005*** -0.005*** -0.005*** (0.001) (0.001) (0.001) (0.001) Certified professions t-1 -0.004 -0.006* -0.006 -0.006 (0.003) (0.003) (0.004) (0.004) College degree t-1 0.004*** 0.004*** 0.004*** (0.001) (0.001) (0.001) Bachelor’s degree + t-1 0.016*** 0.015*** 0.015*** (0.001) (0.002) (0.002) Age (years) t-1 -0.001*** -0.001*** -0.001*** (0.000) (0.000) (0.000) I(married) t-1 -0.003*** -0.002** -0.002* (0.001) (0.001) (0.001) I(multiple-job holder) t-1 -0.001 -0.001 (0.002) (0.002) I(permanent job) t-1 -0.009** -0.009** (0.002) (0.002) I(part-timer) t-1 -0.001 0.000 (0.001) (0.001) Experience t-1 (years) -0.000** -0.000*** (full-time equivalence) (0.000) (0.000) Plant size (21-99) t-1 0.000 0.000 (0.001) (0.001) Plant size (100-499) t-1 0.001 0.028 (0.001) (0.078) Plant size (500+) t-1 0.006*** 0.007*** (0.003) (0.003) Females -0.002*** -0.002* -0.002* (0.001) (0.001) (0.001) Provincial unemployment rate t-1 0.001** (0.000) Industry N N Y Y Occupation N N Y Y # observations 122,800 122,800 122,800 122,800 R2 0.018 0.027 0.029 0.029 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Robust standard errors clustered at the occupation and provinces level in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as provincial dummies, categories of industry, and one-digit occupation. The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993–2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement.

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Table 4A.5 Estimated Effects of the Likelihood of Moving to Another Province in a Year (1993-2010): Licensed Occupations, by Gender

Male Female Dv: I(moved to a different provinces)t LPM[1] LPM[4] LPM[1] LPM[4] I(moved to a different provinces)t-1 0.127*** 0.119*** 0.125*** 0.116*** (0.010) (0.010) (0.010) (0.010) Licensed occupations (both licensed -0.004* -0.006** 0.004 0.006*** professions and compulsory trades) t-1 (0.002) (0.002) (0.003) (0.003) Union t-1 -0.004*** -0.003 -0.009*** -0.007*** (0.001) (0.002) (0.002) (0.002) Certified professions t-1 -0.003 -0.003 -0.005 -0.008 (0.004) (0.005) (0.005) (0.005) College degree t-1 0.005*** 0.004** (0.001) (0.001) Bachelor’s degree + t-1 0.016*** 0.014*** (0.002) (0.003) Age t-1 -0.001*** -0.001*** (0.000) (0.000) I(married) t-1 -0.003*** -0.001 (0.001) (0.001) I(multiple-job holder) t-1 -0.001 -0.001 (0.002) (0.002) I(permanent job) t-1 -0.011*** -0.006** (0.003) (0.003) I(part-timer) t-1 -0.003* -0.004 (0.002) (0.003) Experience t-1 (years) -0.000*** 0.000 (full-time equivalence) (0.000) (0.000) Provincial unemployment rate t-1 0.001 0.001** (0.000) (0.001) Industry N Y N Y Occupations N Y N Y # observations 63,200 63,200 59,600 59,600 R2 0.018 0.029 0.018 0.031 Note: *** p < 0.001; ** p < 0.01; * p < 0.05. [Robust standard errors in parentheses. Occupational licensing is a step function equal to one if the individual works in a licensed occupation. Each regression includes a constant as well as provincial dummies, categories of industry, and one-digit occupation. The control variables are included corresponding to the specification.] Source: Survey of Labour and Income Dynamics (SLID) (1993– 2010). Number of observations was rounded to the nearest hundred according to the Statistics Canada Research Data Centre requirement.

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Chapter 5 Appendix: Occupational Licensing in Canada: A Jurisdictional Review,1990–2016

5.1 Introduction

In Canada, there is no official definition of occupational licensing. The closely comparable definition is regulated occupation, which the Government of Canada defines as “occupations that set their own standards and that require workers to have a license to practice.”75 At first glance, regulated occupations in Canada are equivalent to licensed occupations in the United States, which are identified as occupations with exclusive right-to-practice. However, some provincial government reports (e.g., Manitoba Law Reform Commission, 1994) have adopted the U.S. approach of separating licensing from certification, which is defined in the United States as exclusive right-to-title.76 Comparing the list of regulated occupations provided by the Canadian federal government, and the list of licensed and certified occupations in the Manitoba report shows that not all regulated occupations are licensed. The inconsistency between the federal and provincial definitions makes one question whether or not it is appropriate to compare the studies of economic impacts of occupation licensing in the United States and of occupational regulation in Canada. At the very least, researchers need to be careful when interpreting and comparing the results.

The regulated occupation has been the research focus in Canada for the past two decades.77 However, the definition has not yet been critically examined. All recent studies take the definition of regulated occupation as the de facto definition of occupational licensing.

75 https://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#Regulated_Occupations 76 Exclusive right-to-title is defined as the designation of a reserved or protected title to professionals practising in their relevant occupation. 77 See for example, Banerjee & Phan (2014); Chen & Fougère (2011); Girard & Smith, (2013); Gomez, Gunderson, Huang, & Zhang (2015); Jantzen (2015). 134

Although there has been no study looking into the licensing laws in Canada, a few studies tap into the legislative documents, and in particular, review the historical development of professional regulations by systematically categorizing the occupational statutes in a few Canadian provinces (Adams, 2009a, 2009b, 2010). Most of the reviews cover the period before the 1960s. No recent period has been examined.

In this jurisdictional review, I provide an updated and more comprehensive review of occupational regulations in Canada, focused on identifying the licensing and certification laws granted by the provincial governments. I also provide a holistic review of the development of occupational licensing legislation in the past three decades and discuss a few new trends. I constructed the licensing and certification indicator for the four-digit National Occupational Classification (NOC) by reviewing the occupational statutes and regulations across 10 Canadian provinces. I define occupational licensing based on the exclusive-right-to-practice clause, and occupational certification based on the exclusive-right-to-title clause in the legislation. I found that in the Canadian context, the regulatory acts that cover most licensed professions have clauses on both the right-to-practice and the right-to-title; the exception is for well-established professions, including medicine, law, and teaching. Consequently, the licensing indicator is constructed based solely on the right-to-practice. I restrict the certified profession indicator to be mutually exclusive from the licensed occupations, and include the professions with only the exclusive-right-to-title clause. This indicator is crucial to examining the economic impacts of occupational licensing in the Canadian context.

The paper is organized as follows: I begin with a brief review of occupational regulation in Canada. Second, I review the literature and discuss some conceptual understanding of the rationale for the government to establish licensing laws. Third, I briefly illustrate the content analysis method used in the jurisdictional review. Finally, I report the findings and discuss the implications of this review.

5.2 Occupational Regulation in Canada

In the Canadian context, licensing and certification are not separated. The federal definition of regulated occupations is “occupations that set their own standards and that require workers to 135

have a license to practice.”78 However, provincial governments decide which occupations are regulated in their own jurisdiction. Table 5.1 lists the number of regulated occupations by province, based on Employment and Social Development Canada’s regulated occupation data, which is published by the Canadian Job Bank. As of 2015, 106 occupations are regulated in at least one Canadian province. Quebec has the highest number of regulated occupations, 98, while Newfoundland and Labrador has 47, the lowest number among the provinces.

------Insert Table 5.1 about here ------

Regulated occupations in Canada differ from licensed or certified occupations in the United States in part because they are divided into two categories: regulated professions and compulsory trades. Regulated professions are occupations that “usually require several years of university or college education, practical experience under the supervision of a licensed worker in the chosen profession, and the successful completion of a licensure examination.”79 In Canada, professions such as medicine and surgery, law, dentistry, pharmacy, and veterinary medicine were established before 1900 through legislative recognition (Adams, 2009b, 2010). Following this trend, professions such as architecture, nursing, physiotherapy, and teaching, which require special skill sets to practise, obtained professional recognition in the 1930s by having their own occupational acts in various provinces. Over the years, many more professions have become regulated (e.g., engineering, occupational therapy, and psychology). As seen in Panel B of Table 5.1, numbers of regulated professions differ slightly across provinces, ranging from 64 in Alberta to 40 in Newfoundland and Labrador.

Compulsory trades are skilled trades that require a license to practise.80 Specifically, workers in compulsory trades must possess a qualification or apprenticeship certificate. Both compulsory and voluntary trades are regulated by the provincial trade boards, including

78 https://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#R 79 https://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#Regulated_Professions 80 https://www.jobbank.gc.ca/content_pieces-eng.do?cid=24#Apprenticeship 136

Alberta Apprenticeship and Industry Training, the Nova Scotia Apprenticeship Agency, and the Ontario College of Trades. Compulsory trades’ coverage varies significantly across provinces. According to the federally-regulated-occupation database, Quebec and Alberta each have more than 20 compulsory trades, while other provinces have significantly fewer (See Panel C, Table 5.1).81

5.3 Literature Review: Licensing Laws and Right-to- practice

Licensure law is the strongest mechanism with which to close occupational markets because licensure enjoys both jurisdictional protections and monopolistic power (e.g., Begun, Crowe, & Feldman, 1981; Carpenter, 2012; Kleiner, 2000; Moore, 1961; Paul, 1984). Exclusive right- to-practice, secured by passing licensure laws, is a crucial instrument that ensures professional associations will maintain control of occupational labour markets. Rubin (1980) identifies six critical features that appear in a statutory context, among which the prohibition of unauthorized individual-practice is the most powerful yet controversial one. 82 Increasing government occupational regulation has been observed both in the United States and in Canada in the past century. Kleiner (2006) reports “a consistent upward trend in the growth of the number of laws or administrative procedures that regulate occupations” (p. 105) and “much variation in the changes in the stringency of licensing requirements over time” (p. 105) for occupations such as dentists, teachers, cosmetologists, attorneys, and accountants in the United States. Adams (2009a) also reports that, in Canada, more self-regulating professions have become regulated over time and suggests that the licensing laws have been tightened since the 1940s and 1950s.

There has been a consensus among researchers looking at the American context that licensure

81 British Columbia established a compulsory trade system in 1997, and then abolished it completely in 2003. Common sense would suggest that no compulsory trades would exist in British Columbia after 2003; however the federal data suggest that six trades are compulsory in British Columbia, such as steamfitters, electricians, refrigeration, and air conditioning mechanics. More detailed discussion can be found in the discussion section, on page 19. 82 The other five features include (1) the creation of a regulatory body; (2) education and other entry standards; (3) grandfather clauses; (4) codes of conduct; and (5) disciplinary provisions (Rubin, 1980). 137

law grants privileges to specified professionals even though licensure laws are purportedly introduced to protect the public interest (e.g., Becker, 1986; Graddy, 1991; Kleiner, 2006; Maurizi, 1974; Rubin, 1980; Stigler, 1971). Despite the debate over whether or not licensing is appropriate for various occupations, almost all professional associations actively seek licensing status (e.g., Begun et al., 1981; Moore, 1961; Paul, 1984; Welsh, Kelner, Wellman, & Boon, 2004). As Welsh et al. (2004) argue, “for modern professions, regulation or state licensure is a primary way to achieve market closure” (p. 218). Title protection (i.e., certification) is considered “the first step in a push for full occupational licensing” (Beatty & Gunderson, 1979, p. 53).

Several professions have gained exclusive right-to-practice as early as the 19th century. Most of the American studies focus on these more established licensed occupations, such as medicine, dentistry, law, and accounting. 83 The occupations in the health care sector are examined more thoroughly because it is easy to defend the claim that health care professionals must have a high level of professional knowledge and that licensing is necessary to protect public safety.84 Like medicine and dentistry, law and accounting are among the professional occupations that achieved occupational-licensing status early on through state legislation. The rationale to license the legal profession is based on their proximity to governance and law enforcement, while the accounting profession was heavily licensed in the United States because of its significant impacts on business and the economy.85 These professions have received more scholarly attention than professions that achieved licensing status in later

83 Most of the American studies tend to focus on one occupation (e.g., medicine and surgery (Kleiner, 2000, 2013); dentistry (Boulier, 1980); nursing (in relation to registered nurses) (White, 1980); or optometry (Begun et al., 1981; Haas-Wilson, 1986)). 84 White (1980) documents the shift from certification to licensing for registered nurses across all states, which mainly occurred between the early 1950s and the late 1960s; the nurses’ association convinced lawmakers that certification could not prevent untrained individuals from performing nursing activities and that these activities should be regulated to protect patient safety. Boulier (1980) documents dentistry’s evolving licensing system. Most states licensed dentistry by 1900; the National Board of Dental Examiners, a professional association issuing qualification certificates to dental practitioners, was formed in 1928; and by 1976, its certificate was recognized by most states. Begun et al. (1981) describe a path model of optometry regulations in various states, and reported that urbanization and state economic conditions are positively associated with regulators being in favour of licensing. 85 See, for example, Kleiner (2006), Richardson (1987), and Young (1991). 138

cohorts because their licensure law provisions have gone through multiple iterations of amendments, and so contain fewer loopholes and are more restrictive than those concerning occupations licensed later.86

Licensed low-skilled occupations are mainly examined in the United States when multiple occupations are reviewed,87 except barbering, which has been one focus of licensing in the United States since the late 1980s (e.g., Adams, Jackson, & Ekelund, 2002; Carpenter, Knepper, Erickson, & Ross, 2015; Hall & Pokharel, 2016; Thornton & Weintraub, 1979). This occupation is the only licensed low-skilled occupation that reports the requirement of apprenticeship training among those low-skilled occupations that are studied most often in the United States.88

In the Canadian context, researchers have focused on the regulated occupations. Similar to the U.S. studies, most Canadian researchers examine a single occupation from the health care profession cluster, in one or several particular jurisdictions. 89 Since the occupations investigated all fall into the group of exclusive-right-to-practice, the difference in definitions of licensing and occupational regulation is muted.

86 Zacharias’s (2009) data search covers more than 500 law review articles that were written between 1978 and 1988 and which refer to the legal profession as a “self-regulated profession.” 87 Accountants still have limited access to exclusive right-to-practice, which applies to such activities as auditing and public accountancy, despite the professional associations’ efforts to gain licensing recognition. Byington, Sutton, and Munter (1990) discuss the accountancy profession’s need to convince governments to allow them to retain monopolistic power. Byington and Sutton (1991) point out that lawmakers’ decisions to license accountants primarily rely on the costs and benefits of government intervention. Public trust is one such benefit of licensing. That is, in order to gain the public trust, accounting associations and licensing boards have imposed higher professional standards so as to defend self-regulation. Gorman (2014) studies professional self- regulation, focusing on both law and accounting, and argues that both professions are under attack due to unsuccessful self-regulation. In particular, he points to external governmental pressure to limit self-regulatory capacity. 88 In the United States, apprenticeship training is governed by the Federal Department of Labor, and this department generally does not interact with state legislatures and licensing bodies except in the case of barbering. Thornton and Weintraub (1979) report that 46 states require that barbers complete apprenticeship programs, which ranged from six to 36 months, in 1976. 89 See, for example, Bourgeault (2000); Hadley (1995); Martin, Turcotte, Matte, & Shepard (2013); McKendry & Langford (2001); Schiller (2014); and Welsh et al. (2004). 139

Some studies focus on regulated occupations that have overlapping practice components (e.g., Adams (2004) and Kleiner & Park (2010) for dentists and dental hygienists; Baldwin, Hutchinson, & Rosenblatt (1992) for midwives and physicians; Begun & Lippincott (1987) for optometrists’ efforts to gain prescription rights; and Kleiner, Marier, Park, & Wing (2016) for physicians and nurses). Shaked and Sutton (1981) theorize the welfare outcomes of the entry of rival unlicensed professionals. The authors argue that policy-makers face only one question when dealing with occupational licensing, “whether the profession should, or should not, be allowed to retain monopolistic powers” (Shaked & Sutton, 1981, p. 217). However, when occupations exist that perform similar tasks, lawmakers are frequently put in the position of adjudicating the boundaries of practice between competing professions. Several studies document the regulatory disputes that occurred between regulated professions that perform similar activities.90

Several studies have considered occupational regulations across multiple professions, within or across Canadian jurisdictions that were in the early stages of developing occupational regulations (Adams, 2009a, 2009b, 2010; Brockman, 1998; Wolfson, Trebilcock, & Tuohy, 1980). Reviewing provincial acts and regulations dating from 1867 to 1967 to investigate professional regulations in five provinces, Adams (2010) defines professional privilege as both the establishment of a regulatory body and the authority to determine the training required for entry into practice. Her investigation covers the early development of professional regulations found in several legal provisions. In her earlier work, Adams (2009a) assumes the common definition of occupation regulation and does not distinguish between licensing and certification. More importantly, she places emphasis on the self-regulatory professions instead of the full spectrum of occupations, and consequently leaves out the compulsory trades.

A few studies have documented the inconsistent occupational licensing requirements enacted by legislators across jurisdictions (Carpenter et al., 2015; Moore, 1961; Nafziger & Hiscox, 1976; Zhou, 1993). Moore (1961) looks into the licensing law legislated by the state of Illinois

90 For example, dentistry and dental hygiene (Adams, 2004), medicine and midwifery (Baldwin et al. 1992), and medicine and nursing (Kleiner et al., 2016). 140

and the city of Chicago and suggests that state licensing laws are most likely the outcomes of professional lobbying while municipal licensure laws aim more at protecting the public interest. Carpenter et al. (2015) study the scope of practice among the low-level occupations in the United States and find that the requirements vary not only between occupations, but also within the same occupations in different jurisdictions.

Licensing laws also evolve over time. Kleiner (2006) describes the evolution of licensing legislation in Minnesota. There was a moderate increase prior to 1950, followed by rapid growth from 1968 to 1990; the growth in the number of newly licensed occupations was much lower between 1998 and 2004 compared to the previous periods, and accounted only for approximately 0.25% of the overall growth in state employment. Kleiner also reports the number of proposed modifications to the existing licensure laws in Minnesota and notes that fewer occupations were licensed around the year 2000 than in the 1980s and 1990s (Kleiner, 2006).

Lawmakers in different jurisdictions also learn from each other when licensing an occupation. Zhou (1993) studied the institutional diffusion of occupational licensing across American states between 1890 and 1950. Using institution theory, the author suggests that state governments adopted occupational-licensure laws not only because of demands coming from the professionals in those occupations or from the general public, but also because of the legitimacy granted by other legislators in other jurisdictions. As a result, “licensing is more likely to diffuse across states for particular occupations rather than within a state” (Zhou, 1993, p. 549). The patterns of diffusion, however, are inconsistent across occupations. To the best of my knowledge, Zhou (1993) is the only empirical work that focuses on how occupational legislation diffuses across jurisdictions. However, it is hard to replicate such tests in the Canadian context because of the small number of provinces, which means that there is insufficient variation to look into such an issue empirically. A qualitative study might shed some light on the subject.

In Canada, the government is heavily involved in occupational regulation, particularly in the health care sector, which has been rigorously seeking a very strictly defined scope of practice

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in licensure laws.91 This is because the government “has a greater role in health care, not only as the regulator of health professions but also as a third-party payer of medical services” (Bourgeault & Fynes, 1997, p. 1060). The tendency of the government to regulate lower-tier health care professions is rationalized as providing a more cost-effective way of providing health care services.92

In sum, the literature review suggests that exclusive-right-to-practice is valid when defining occupational licensing status. Scholars are interested in the legislative development of occupational licensing, but have yet provided a comprehensive review, particularly in Canada.

Research Questions

This research focuses on distinguishing occupational licensing from certification in Canada. There has been some confusion over which occupations in Canada are, in fact, regulated according to the law. For example, the common understanding of mandatory trades and voluntary trades differs from the list of trades that are legislated. At present, no explicit definition of occupational licensing exists, and previous research has not fully established the distinction.

In this study, I focus on the specific legislative language found in the statutes (and the regulations, in certain cases) that specify the exclusive right-to-title and right-to-practice for all occupations based on four-digit National Occupational Classification (NOC). I am interested in the following two questions:

1) What professional associations were granted a right-to-practice and right-to-title clause in the legislation that pertains to their specific licensure, and were these rights granted after

91 For example, as of 2016, among 40 licensing bodies recognized by the Ontario government, 26 colleges represent health care professionals. http://regulatorsforaccess.ca/resources/regontario.aspx 92 This was manifested in the process of licensing midwifery in Saskatchewan, despite the very small number of practitioners in the province. Mckendry and Langford (2001) suggest that the recognition of the midwifery profession is to provide consumers with alternative choices to medical dominance, while Bourgeault and Fynes (1997) argue that lawmakers want to have more economically efficient/cheaper alternatives. 142

the 1990s?

2) Are there any other restrictions that could lead to more monopolistic power?

5.4 Research Method

This investigation applies the directed content analysis method (Hsieh & Shannon, 2005) to Canadian legislative documents, some of which were published as early as 1980 (e.g., in British Columbia). The intention is to look for specific descriptions of occupational licensing in legislation across the 10 Canadian provinces. In this type of study, the unit of the analysis is the act that establishes the occupation in a particular year in a particular province.

The main sources for this jurisdictional review are the revised statutes and the corresponding regulations for professions and occupations in all 10 Canadian provinces from 1990 to 2016, the latest year of publication. These were accessed from each province’s Queen’s Printer, in either online or print format. I also conducted a review of the annual (point in time) statutes that were published by the 10 provincial legislatures, so as to identify whether a specific legislative clause had been amended between the two cycles of the revised statutes. This search yielded a total of 1,537 relevant acts and regulations.

The content analysis method in this research is built on one developed by Adams (2010). The author conducted a thorough investigation of the legislation of professional regulations in British Columbia, Saskatchewan, Ontario, Quebec, and Nova Scotia between 1867 and 1961. Focusing on the professions, she documented the years in which 38 professions and 18 occupations were first regulated in the five specified provinces, based on a holistic review of the self-regulation features addressed in the legislative language. My approach is more direct than that of Adams (2010). Since there is no clear legislative definition for licensing and certification, descriptions of exclusive right-to-practice and exclusive right-to-title, respectively, are used as proxies. Table 1 of the Appendix 5A presents examples of the language used in the statutes. For instance, an occupation is licensed if there is a clear clause that states that there is exclusive right-to-practice; a profession is certified only if a statute grants exclusive right-to-title. When both clauses exist in an occupational legislation, licensing status shall prevail. 143

5.5 Results

5.5.1 Regulated occupations are not all licensed

In Canada, the federal database and provincial statutes and regulations do not consistently define regulated occupations, regulated professions, or compulsory skilled trades. According to the federal definition, regulated occupations are those that require a license to practise. However, Ontario’s fair access commissioner defines regulated occupations as “professionals who need the authority of the regulatory body to practise their profession in Ontario, to use a professional designation, or both.”93 No definitions specifically stated in licensure laws are found in other provinces.

Occupational licensing falls solely in the purview of the provincial legislatures.94 A search through the legislative documents related to licensure laws reveals very weak support for the establishment of clear guidance for the regulatory processes in Canada. Although acts that regulate occupations have been introduced and revised, until recently there have been no laws that describe how occupations should be regulated. The only legislative documents found to do so are in the Statutes of Nova Scotia, 2012, which define regulatory processes as follows:

“Regulatory processes” means those processes and matters generally prescribed by a regulated health profession’s governing statute to: (i) establish the scope of practice of the profession, (ii) govern the registration or licensing of members of the profession, (iii) establish investigative and hearing processes for complaints

93 http://regulatorsforaccess.ca/resources/regontario.aspx 94 Provincial occupational licensing laws are enacted in different forms. There are two types of acts – public and private. Public acts apply to all persons within a jurisdiction. Private acts only affect particular groups or areas. Occupational regulation legislation exists in public acts in six Canadian provinces. However, only Manitoba has a private act respecting a specific occupation; this one act, established in 1990, covers social workers. In New Brunswick, almost all occupational acts are categorized as private acts; the New Brunswick statutes cover such occupations as agrology, architecture, engineering, occupational therapy, and veterinary medicine. Private occupational regulation acts are found in Nova Scotia as far back as 1968, however the more recent pieces of legislation, from 1990 onward, have been public acts. In Ontario, most of the licensing laws are legislated as public acts, while several titling laws (i.e., certification laws) are private acts, among which two were revised as public acts (i.e., dieticians (1991) and human resources professionals (2013)). 144

involving members of the profession, (iv) address quality assurance matters for the profession, (v) govern appeals or reviews from any decisions made pursuant to the profession’s governing statute, or (vi) regulate any other aspect of the profession; (Regulated Health Professions Network Act, SNS, 2012, c 48).

Based on the description of provincial statutes, it appears that the federal definition of regulated occupations is, in fact, misleading. Regulated occupations could be either licensed or certified, based on provincial legislation. It remains uncertain as to why both federal and provincial governments do not clearly identify what constitutes a licensed or certified profession.

5.5.2 Licensing legislation: Exclusive right-to-practice

The licensed professions typically have their own acts. Table 5.2 showcases six occupations that have an independent act: accountants, foresters, architects, dietitians, lawyers, and psychologists.95 These six professions are representative of the array of occupations that have been regulated in at least one province. ------Insert Table 5.2 about here ------

Several observations can be drawn from the review of the corresponding regulatory acts. First, the years in which legislative assemblies approved acts concerning the regulation of a particular profession differ from province to province. Architects gained exclusive-right-to- practice between 1960 and 1987, while accountants achieved exclusive-to-title between the 1960s and 1990s. Table 5.2 comes with a caveat: the year listed is not the very first time the relevant clauses were established. Rather, it is the first time such a clause appeared in the legislation during the period of my jurisdictional review (i.e., 1990–2016). As a result, while the table may seem to indicate that there has been policy diffusion across provinces, that conclusion cannot be drawn—although, for a relatively newly established profession, such as

95 Full list of occupational licensure law can be provided upon request. 145

dietitians, I speculate this might be true.

Second, occupational regulation also varies by profession. For example, dietitians are licensed in Ontario, Prince Edwards Island, and British Columbia because of their exclusive right-to- practice, whereas in other provinces they have only exclusive right-to-title privileges. Accountants achieved exclusive-right-to-practice clauses in the three Maritime Provinces (i.e., Nova Scotia, Prince Edward Island, and Newfoundland and Labrador), while I found only exclusive-right-to-title clauses in other provinces.

Third, most of the regulatory acts of licensed professions have clauses granting both the right- to-practice and the right-to-title; the exception is for well-established professions such as medicine, law, and teaching. It is unclear why both are granted, and I speculate this reflects the progression of occupational licensing in the legislative process. For long-standing occupations with the right-to-practice, society recognizes that these professions have the right to impose standards of practice and to restrict entry; title protection seems redundant. There are a few cases where right-to-title and right-to-practice were enacted at different times. For example, professional engineers in New Brunswick achieved right-to-title first, in 1973, then right-to- practice in 1986.96 However, I will not extrapolate that title protection is the stepping stone to licensure.

The majority of licensed professions had already gained legislative recognition by 1990. In a few cases, professions licensed by newly established laws had already been licensed in one or more other provinces before the 1990s. For example, British Columbia and Quebec established licensing power for the forester profession in 1970 and 1973, respectively, while this profession was licensed in Alberta and Ontario around 2000 (See Table 5.2). Other examples are: respiratory therapy, which gained licensing status in Quebec in 1973, Manitoba in 1987, Ontario in 1991, Saskatchewan in 1999, and New Brunswick and Alberta in 2009; and medical

96 A couple cases in which a right-to-practice clause was enacted earlier than a right-to-title clause were also found in this review. For example, dental hygienists in New Brunswick were granted the right-to-practice in 1976 and reserved title in 1985. Further review suggests that these occupations had experienced changes in their professional title over time, and the new reserved title reflects such changes. 146

laboratory technology, which was licensed in Saskatchewan in 1989-1990; New Brunswick and Ontario in 1991; and Nova Scotia, Alberta, and Manitoba after 2000. Only a few professions have been able to gain licensing status since the 1990s (e.g., early childhood education in Newfoundland and Labrador, and biological and medical laboratory technology in Alberta). ------Insert Table 5.3 about here ------

A review of the legislation shows that exclusive right-to-practice is the dominant form of occupational regulation. The legislation for most occupations has an exclusive-right-to- practice clause, which means that these occupations are licensed. In contrast to Table 5.1, which shows the number of regulated occupations across provinces based on the federal database, Table 5.3 indicates the numbers of licensed professions according to provincial laws of exclusive-right-to-practice as of 2016. In total, 106 professions are licensed in at least one of the Canadian provinces. Ontario, with 63, has the largest number of licensed professions, while Newfoundland and Labrador, with 40, has the lowest.

5.5.3 Exclusive right-to-title

The exclusive right-to-title clause usually requires practitioners who are not members of the relevant professional associations not to use the title or imply that they are part of their profession’s association. There are fewer professions with sole title protection (i.e., certified professions) than there are licensed occupations, according to provincial occupational- regulation legislation. The number of certified professions varies slightly among provinces except for Prince Edward Island, and Newfoundland and Labrador, which have only two certified professions. Notably, the numbers presented in Panel C of Table 5.3 reflect the number of occupations with right-to-title and not the number of reserved designations that these occupations are entitled to. For example, accountants in Ontario had three distinctive reserved designations before 2014. These designations have been amalgamated into one since the proclamation of Chartered Professional Accountants of Ontario Act in 2017 (Chartered Professional Accountants of Ontario Act, S.O., 2017, c 8).

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Similar to licensed professions, various professions have been certified at different times by different provincial legislatures. For example, engineering technologists or technicians became certified in Ontario, New Brunswick, and Nova Scotia in the late 1970s and early 1980s. Translators received exclusive right-to-title in Ontario, Quebec, and New Brunswick in the late 1980s and early 1990s.

5.5.4 Compulsory skilled trades

The skilled trades are regulated in each province’s apprenticeship and trades act. Two types of market restrictions can be found in the apprenticeship acts—restrictions on practice and restrictions on employment. In the Revised Statutes of Ontario 1970, where a certified trade is defined based on the statutes established in 1968-1969, the right-to-practice is described as follows:

No person, other than an apprentice or a person of a class that is exempt from this section or a person referred to in subsection (4), shall work or be employed in a certified trade unless he or she holds a subsisting certificate of qualification in the certified trade

No person shall employ any person, other than an apprentice or a person of a class that is exempt from this section (4), in a certified trade unless the person employed holds a subsisting certificate of qualification in the certified trade” (Apprenticeship and Tradesmen's Qualification Act, R.S.O., 1970, c 24, s.10).

The language used to describe the licensing restriction for the trades covered in the acts is very similar across jurisdictions,97 despite minor variations when describing who may obtain an exemption and, especially, the requirements associated with obtaining temporary permits. Most acts limit both the supply and demand side of the trades at the same time. The exception is Alberta, whose relevant act has a section on the prohibition of employment (implemented in 1978), which was implemented before the definition of a trade as a licensed practice (in 1980).

97 Right-to-practice clause in the trade and apprenticeship acts is typically described as follows, “[a] person shall not work in a compulsory certification trade unless that person (a) holds a trade certificate in that trade, …” (RSA 2000, A-42, s.21); and prohibition of the employment of unauthorized practitioners is defined accordingly, “[a]n employer shall not employ a person to work in a designated trade if the employer knows, or would reasonably be expected to know, that the person who is to carry out that work is not permitted under this Act to carry out the work in that trade. 1991 cA-42.3 s26” (RSA 2000, A-42, s.21). 148

In Panel B of Table 5.3, I present the number of compulsory trades, as defined by either restriction on practice or employment, by provinces. Quebec has the highest number of compulsory trades, 28; while British Columbia and Newfoundland and Labrador have no compulsory trades based on the corresponding apprenticeship act.

The rationale for regulating skilled trades was to establish appropriate apprenticeship programs. Over time, a few skilled trades become “compulsory” because the apprentices are required to acquire the restricted skill sets through registered apprenticeship programs. Table 5.4 shows the year in which prohibitions against unauthorized practice and employment appeared in the legislative language of the respective provinces’ apprenticeship acts. Table 5.4 also shows the years in which the trades gained compulsory status. The year the legislation was enacted varies across provinces: Ontario was the first to implement this legal restriction, followed by New Brunswick, Prince Edward Island, Alberta, and Saskatchewan, all of which introduced a restrictive clause in the 1970s; British Columbia, Manitoba, and Nova Scotia implemented the exclusive right-to-practice in the late 1990s or early 2000s. In the period considered, only Newfoundland and Labrador ever imposed a limit on minors working in the trades. The most interesting case is British Columbia’s disappearing exclusive-right-to-practice clause; while other provincial governments have tended to tighten the occupational regulation oversight in recent years, the British Columbia government took an opposite turn. Here, the exclusive- right-to-practice clause appeared in the Industry Training and Apprenticeship Act (1996), but disappeared six years later, in the Industry Training Authority Act (2003), when British Columbia decided to pursue a market-driven approach to trades training and abolished the compulsory trades system (e.g., Heinrichs, 2015). ------Insert Table 5.4 about here ------

A review of the statutes on skilled trades also shows that the right-to-practice was enacted earlier than the term compulsory trades was entered into the statutes. Prior to this, provincial lawmakers assigned various other names to this group of professions, such as designated trades, certified trades, or compulsory certified trades.

149

For statutes on compulsory trades, no clauses on title protection could be found in any of the 10 provinces, despite the fact that in some jurisdictions, the trades are called designations. The exception is in a 2012 revision of the Prince Edward Island Apprenticeship and Trades Qualification Act. In this act, the legislature spent a significant amount of effort in defining title protection for practitioners of the compulsory trades (Apprenticeship and Trades Qualification Act, RSPEI 1988, c A-15.2, s.22).98

5.5.5 Other market closure clauses

There are other ways in which legislative power can restrict workers from practising in a certain occupation. Early acts include the right to advertise. Legislative language was found that specifically prohibits professionals from using their professional titles or the name of their professional associations in advertising. These clauses were all established before 1990. Most disappeared after recent amendments.

Another type of restriction seen in the legislation is the prohibition of nonregistered members from collecting fees for services rendered. For example, the Prince Edward Island Engineers Act states,

No person, partnership, association of persons, or corporation shall be entitled to the payment or recovery of any fees or charges in any court, or otherwise, for any service performed within the practice of professional engineering unless, at the time the services were performed, the person was registered or licensed under this Act or the partnership association of persons, or corporation was the holder of a valid certification of authorization. 1990, c.12, s.27. (Engineering Profession Act, RSPEI 1988, c E-8.1)

Brockman (1996) also documents other limited restrictions that are found in the professional acts including granting power to sanction authorized practice through criminal charges or civil injections, tightened government oversight, and prohibition of delegation.

In several cases other than the skilled trades, prohibition of employment is observed in acts

98 “(7) No person shall use any title, name, abbreviation or description implying that he or she holds a permit in a compulsory certified trade unless he or she holds, or is deemed to hold, a permit in that compulsory certified trade” (Apprenticeship and Trades Qualification Act, RSPEI 1988, c A-15.2, s.22). 150

that regulate particular occupations. For example, licensing laws for real estate brokers also explicitly legislate that unauthorized practitioners cannot be employed in this profession. This trend, coupled with the disappearance of restrictions on advertising, is in line with the trend in the transition from self-employed to salaried professions.

5.6 Discussion and Conclusion

This jurisdictional review finds that exclusive-right-to-practice is the dominant form for occupational regulation in the Canadian context. The majority of provincial occupational statutes and regulations are licensure laws, many of which also include a right-to-title clause. The regulated occupations with title protection are either relatively new occupations or those for which it is becoming harder to make a public-interest argument for licensing status.

In Canada, lawmakers are not particularly concerned about the distinction between licensing and certification despite the fact that legislative discussion has been concentrated around whether or not it is appropriate to regulate certain professions. There are several signs that lawmakers have attempted to restrict occupational regulations. First, the number of newly licensed or certified occupations has significantly decreased since the 1970s. My review of Canadian legislation indicates that, since the 1990s, no occupation has been completely licensed or certified without either a previous private legislative effort or an attempt at jurisprudence in other jurisdictions.

In addition, the methods and formats of occupational regulations have tended to shift from a homogeneous approach within a province to one that converges across provinces for the same occupation. This trend could also support the tightening of legislative oversight between provinces. A professional association in a single province may not have sufficient power to push further legalization of the licensure laws. Ontario provides an example of the early approach to regulating occupations. Its legislative language and format were identical for all professional acts in both the 1980 and 1990 revised statutes. Other provinces exhibited more or less similar patterns during the same period. Recently, the amalgamation of three professional accountancy designations, in several provinces, within two years (i.e., Quebec (2012), Alberta (2014), Saskatchewan (2014), New Brunswick, (2014), Newfoundland and 151

Labrador (2014), British Columbia (2015), Manitoba (2015), Prince Edward Islands(2015), Nova Scotia (2015), and Ontario (2017)), has demonstrated single-occupation convergence. Two possible explanations for this move toward consolidation can be inferred based on similar legislative language found in those acts: aggregating professional influence to push the licensing status, or combining power to resist the attempt to deregulate the profession. Either way, these professional associations recognize the disadvantage of having competing licensing bodies. Not only do they now push toward a unified self-regulatory body within their jurisdictions, but also they are actively working toward achieving interprovincial consistency.

In conclusion, there has been a lack of identification of occupational licensing and certification in the Canadian context, which has led to an underutilization of research findings from other countries such as the United States and the United Kingdom. After reviewing legislative documents from the 10 Canadian provinces, I constructed both licensing and certification indicators based, respectively, on the exclusive-right-to-practice and exclusive-right-to-title clauses enacted in the provincial statutes. This jurisdictional review also finds that the majority of the licensed occupations enjoy title protection in the acts, which further demonstrates that the federal definition of regulated occupation does not accurately represent the nature of occupational regulation in Canadian provinces.

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Adams, T. L. (2004). Inter-professional conflict and professionalization: Dentistry and dental hygiene in Ontario. Social Science & Medicine, 58(11), 2243–2252.

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Adams, T. L. (2010). Profession: A useful concept for sociological analysis? Canadian Review of Sociology/Revue Canadienne de sociologie, 47(1), 49–70.

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Becker, G. (1986). The public interest hypothesis revisited: A new test of Peltzman's theory of regulation. Public Choice, 49(3), 223–234.

Begun, J. W., Crowe, E. W., & Feldman, R. (1981). Occupational regulation in the states: a causal model. Journal of Health Politics, Policy and Law, 6(2), 229–254.

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Begun, J. W., & Lippincott, R. C. (1987). The origins and resolution of inter-occupational conflict. Work and Occupations, 14(3), 368–386.

Boulier, B. L. (1980). An empirical examination of the influence of licensure and licensure reform on the geographical distribution of dentist. In Rottenberg, S. (Ed.). Occupational Licensure and Regulation, 73–79. Washington, DC: American Enterprise Institute.

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Bourgeault, I. L., & Fynes, M. (1997). Integration lay and nurse-midwifery into the US and Canadian health care systems. Social Science & Medicine, 44(7), 1051–1063.

Brockman, J. (1996). Dismantling or fortifying professional monopolies? On regulating professions and occupations. Manitoba Law Journal, 24(2), 301.

Byington, J. R., Sutton, S., & Munter, P. (1990). A professional monopoly's response: Internal and external threats to self-regulation. Journal of Corporate Accounting & Finance, 1(4), 307–316.

Byington, J. R., & Sutton, S. G. (1991). The self-regulating profession: An analysis of the political monopoly tendencies of the audit profession. Critical Perspectives on Accounting, 2(4), 315–330.

Carpenter, D. M. (2012). Testing the utility of licensing: Evidence from a field experiment on occupational regulation. The Journal of Applied Business and Economics, 13(2), 28.

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Girard, M., & Smith, M. (2013). Working in a regulated occupation in Canada: An immigrant– native born comparison. Journal of International Migration and Integration, 14(2), 219– 244.

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Hall, J. C., & Pokharel, S. B. (2016). Barber licensure and the supply of barber shops: Evidence from US states. Cato Journal, 36, 647.

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Kleiner, M. M. (2006). Licensing occupations: Ensuring quality or restricting competition? Kalamazoo, MI: W. E. Upjohn Institute.

Kleiner, M. M. (2013). Stages of occupational regulation: Analysis of case studies. Kalamazoo, MI: W. E. Upjohn Institute.

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Table 5.1 Interprovincial Differences in Regulated Occupations, 2015

Province BC AB SK MB ON QC NB NS PEI NFL Panel A Number of regulated occupations 67 92 67 71 80 98 65 69 55 49

Panel B Number of regulated professions 54 64 53 63 63 61 54 53 44 40 Panel C Number of compulsory trades 6 21 6 9 9 29 6 9 6 2 Source: Government of Canada Job Bank data accessed at: https://www.jobbank.gc.ca/content_pieces- eng.do?cid=24#Regulated_Occupations

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Table 5.2 Examples of Interprovincial Differences in Right-to-practice and Right-to-Title Clauses

Province BC AB SK MB ON QC NB NS PEI NFL Accountants Right-to-title 1960 1970 1978 1987 1991 1973 1986 1967 1988 1984 Right-to-practice 1967 1988 1984

Foresters Right-to-title 1970 1999 2006 2000 1973 2001 1999 Right-to-practice 1970 2000 1973 2011

Architects Right-to-title 1960 1970 1978 1987 1973 1987 1968 1978 Right-to-practice 1960 1970 1978 1984 1973 1987 1968 1988 1978

Dietitians Right-to-title 1970 1978 1987 1991 1973 1988 1973 1994 1990 Right-to-practice 2008 1991 1994

Lawyers and Quebec Notaries Right-to-title 1990-91 Right-to-practice 1987 1970 1990-91 1987 1990 1973 1973 1970 1988 1990

Psychologists Right-to-title 1977-87 1970 1978 1987 1991 1973 1980 1980 1990 1985 Right-to-practice 1977-87 1970 1991 1990 Note: The information was collected based on the author’s own jurisdictional reviews of legislative documents across 10 Canadian provinces.

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Table 5.3 Interprovincial Differences in Licensed and Certified Professions, as of 2016

Province BC AB SK MB ON QC NB NS PEI NFL Panel A Number of licensed occupations (exclusive right-to-practice) 60 92 67 71 80 98 65 69 55 47 Panel B Number of licensed professions (exclusive right-to-practice) 54 64 53 56 63 61 54 53 44 40 Panel C Number of compulsory trades (exclusive right-to-practice) 0 21 6 9 9 29 6 9 6 0

Panel D Number of certified professions (exclusive right-to-title) 13 12 18 12 8 28 8 13 4 3 Note: The information was collected based on the author’s own jurisdictional reviews of legislative documents across 10 Canadian provinces. Some licensed professions also have right-to-title clauses in their professional acts. Hence, the certified occupations here refer to only those that have right-to-title clauses.

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Table 5.4 Years of Individual Trades Designated as Exclusive Right-to-practice by Provincial Apprenticeship Acts Province BC AB SK MB ON QC NB NS PEI NFL Year established as compulsory trade 1996/97- 2000 1997 1998 1990 1996 2003 n.a. 2003

Year listed with exclusive right-to-practice (mostly referred to as certified trades) 1968/69 1969 1995 Note: The information was collected based on author’s own jurisdictional reviews of legislative documents across 10 Canadian provinces.

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Appendix 5A

Table 5A.1 Examples of the Legislative Language Used in the Occupational Regulation Statutes Year Act Clauses Regulation Status 1991 Ontario Regulated Controlled acts restricted Exclusive right-to- Health 27 (1) No person shall perform a controlled act set out in subsection (2) in the course of practice providing health care services to an individual unless, Professions Act, (a) the person is a member authorized by a health profession Act to perform the controlled act; 1991 or Licensing (b) the performance of the controlled act has been delegated to the person by a member described in clause (a). 1991, c. 18, s. 27 (1); 1998, c. 18, Sched. G, s. 6

2000 Professional Practice Exclusive right-to- Geoscientists of 3. (1) An individual shall not practise professional geoscience unless he or she is a member of practice & the Association and practises in accordance with the terms, conditions and limitations imposed Ontario Act, 2000 on his or her membership. 2000, c. 13, s. 3 (1). exclusive right-to- Use of designations title 5. (1) An individual shall not use the designation “professional geoscientist” or the abbreviation “P.Geo.”, or the corresponding French expression or abbreviation, unless he or she is a member of the Association. 2000, c. 13, s. 5 (1). Licensing

2010 Certified Prohibition, individuals Exclusive right-to- Management 26 (1) No individual, other than a member of the Corporation, shall, through an entity or title otherwise, Accountants Act, (a) take or use the designation “Certified Management Accountant”, “comptable en 2010 [Repealed management accrédité”, “Registered Industrial Accountant” or “comptable en administration on May, 2017] industrielle”, or the initials “C.M.A.”, “CMA”, “F.C.M.A.”, “FCMA”, “R.I.A.” or “RIA”, alone Certification or in combination with other words or abbreviations; (b) take or use any term, title, initials, designation or description implying that the individual is a Certified Management Accountant or a Registered Industrial Accountant; (c) practise as a Certified Management Accountant or Registered Industrial Accountant; or (d) otherwise hold himself or herself out as a Certified Management Accountant or a Registered Industrial Accountant, regardless of whether he or she provides services as a Certified Management Accountant or Registered Industrial Accountant to any individual or entity. 2010, c. 6, Sched. B, s. 26 (1).

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2017 Chartered Designations Exclusive right-to- Professional 17 (1) Subject to the by-laws, a member of CPA Ontario is entitled to use the following practice & designations: Accountants of 1. “Chartered Professional Accountant” and “comptable professionnel agréé”. exclusive right-to- Ontario Act, 2017 2. Any other designation provided for by the by-laws. title Prohibitions, individuals 29 (1) No individual, other than a member of CPA Ontario, shall, through an entity or otherwise, Licensing (a) take or use a designation referred to in section 17 or initials referred to in section 18, whether alone or combined or intermixed in any manner with any other words or abbreviations; (b) take or use any term, title, initials, designation or description implying that the individual is a Chartered Professional Accountant, Chartered Accountant, Certified Management Accountant, Registered Industrial Accountant or Certified General Accountant; (c) otherwise hold himself or herself out as a Chartered Professional Accountant, Chartered Accountant, Certified Management Accountant, Registered Industrial Accountant or Certified General Accountant; or (d) practise as a Chartered Professional Accountant.

2002 An Act to Amend 4 Section 12 of the Act is amended (a) in subsection (1) by striking out “Any person” and Exclusive right-to- the Nurses Act of substituting “Subject to subsection (1.1), any person”; practice & (b) by adding after subsection (1) the following: New Brunswick 12(1.1) Only a person whose name is endorsed in the register as a nurse practitioner is entitled exclusive right-to- to engage in the practice of a nurse practitioner and, subject to any conditions, limitations or title restrictions set out in the by-laws or rules is entitled to hold herself out as a nurse practitioner and use the designation “nurse practitioner”, “NP”, “N.P.” or any other words, letters or figures indicating that she is a nurse practitioner." Licensing

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2010 An Act Title and Designation Exclusive right-to- Respecting the 17(1) No person other than a member of the College shall be entitled to engage in the practice title & exclusive of physiotherapy or use the title “Physiotherapist”, or any words or letters indicative of such College of designation. right-to-practice Physiotherapists 17(2) A member of the College who is entitled to engage in the practice of physiotherapy shall of New use the title “Physiotherapist” or “Physical Therapist” and may use such titles in association Brunswick with the designation “PT”. Licensing Right to Practise 18(1) No person shall practise physiotherapy in New Brunswick unless registered to practise under the provisions of this Act and the regulations. 18(2) No employer shall knowingly cause or permit any person to practise physiotherapy in the Province unless that person is a member of the College.

2005 Registered 18 Except as provided in this Act or the by-laws, no person, other than a Registered Exclusive right-to- Professional Professional Planner, shall assume or use any title, name, designation, initials or description, title including, but not limited to, “Registered Professional Planner” or “RPP” alone, or in Planners Act of combination with any other designation that does, or could, lead the public to believe that New Brunswick person is a Registered Professional Planner. Certification

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