ASR76410.1177/0003122411414817Weste 414817rn and RosenfeldAmerican Sociological Review

American Sociological Review 76(4) 513­–537 Unions, Norms, and the Rise © American Sociological Association 2011 DOI: 10.1177/0003122411414817 in U.S. Wage Inequality http://asr.sagepub.com

Bruce Westerna and Jake Rosenfeldb

Abstract From 1973 to 2007, private sector union membership in the United States declined from 34 to 8 percent for men and from 16 to 6 percent for women. During this period, inequality in hourly wages increased by over 40 percent. We report a decomposition, relating rising inequality to the union wage distribution’s shrinking weight. We argue that unions helped institutionalize norms of equity, reducing the dispersion of nonunion wages in highly unionized regions and industries. Accounting for unions’ effect on union and nonunion wages suggests that the decline of organized labor explains a fifth to a third of the growth in inequality—an effect comparable to the growing stratification of wages by education.

Keywords wages, inequality, unions, labor markets, norms

The decline of organized labor in the United DiNardo, Fortin, and Lemieux 1996); and States coincided with a large increase in wage limited, accounting for only a small fraction inequality. From 1973 to 2007, union member- of rising inequality and only among men ship in the private sector declined from 34 to 8 (Card, Lemieux, and Riddell 2004). percent for men and from 16 to 6 percent for We revisit the effects of union decline on women. During this time, wage inequality in inequality and offer two extensions to earlier the private sector increased by over 40 percent. research. First, we study the effects of union Union decline forms part of an institutional decline while controlling for education and account of rising inequality that is often con- other factors. Analyzing education alongside trasted with a market explanation. In the mar- unions allows a comparison of market and ket explanation, technological change, immi- institutional effects on rising inequality. Sec- gration, and foreign trade increased demand ond, we examine union effects on nonunion for highly skilled workers, raising the premium wages, considering whether wage inequality paid to college graduates (for reviews, see is lower among nonunion workers in highly Autor, Katz, and Kearney 2008; Gottschalk unionized regions and industries. and Danziger 2005; Lemieux 2008). Compared to market forces, union decline is often seen as a modest source of rising aHarvard University inequality (Autor et al. 2008). Scholars view bUniversity of Washington unions’ effects as indirect, mediating the Corresponding Author: influence of technological change (Acemoglu Bruce Western, Harvard University, 33 Kirkland 2002); secondary to other institutions like the Street, Cambridge, MA 02138 minimum wage (Card and DiNardo 2002; E-mail: [email protected]

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Union effects on nonunion workers can even more, rising from .20 in 1973 to .30 in work in several ways. Nonunion employers 2007. Overall trends were driven by move- may raise wages to avert the threat of union ments at the top and the bottom of the wage organization (Leicht 1989). We argue that distribution. Increasing inequality in the late unions also contribute to a moral economy 1970s and 1980s reflected falling wages at that institutionalizes norms for fair pay, even the bottom and rising wages at the top of the for nonunion workers. In the early 1970s, distribution. Since the late 1980s, the growth when 1 in 3 male workers were organized, in wage inequality has been propelled by unions were often prominent voices for wage increases for the highest-paid workers equity, not just for their members, but for all (Lemieux 2008). workers. Union decline marks an erosion of Figure 1 divides the total variance in log the moral economy and its underlying distri- wages into between- and within-group com- butional norms. Wage inequality in the non- ponents. We obtained these components from union sector increased as a result. a regression of log wages with age, race, eth- Our analysis estimates union effects on nicity, education, region, union membership, wage inequality by decomposing the growth in and industry-region unionization rates as pre- hourly wage inequality for full-time workers in dictors. Between-group inequality, measured the private sector. Analyses of the Current by the variance of predicted wages, describes Population Survey (CPS) show that union the dispersion of average wages across the decline explains a fifth of the increase in groups defined by the predictors. Within- inequality among men and none of the increase group inequality, measured by the residual among women if only union wages are consid- variance, describes the spread of wages ered. The effect of union decline grows when among workers in each of these groups. Ris- we account for the link between unionization ing inequality between and within groups and nonunion wages. In this case, deunioniza- increased total wage inequality in the 1980s. tion explains a fifth of the inequality increase Since the 1990s, rising inequality is mostly for women and a third for men. The decline of due to increased within-group inequality. organized labor among men contributes as Falling unionization accompanied rising much to rising wage inequality as does the wage inequality. CPS tabulations indicate growing stratification of pay by education. union decline was especially large among men, falling from 35 percent in 1973 to less than 10 percent by 2007. Employer surveys Trends in Wage show that CPS survey respondents mistak- Inequality and enly report their union status 2 to 3 percent of Unionization the time (Card 1996). When unionization is very low, a relatively large number of non- We analyze trends in the private sector— union respondents incorrectly report being about 85 percent of nonfarm employment union members. Figure 2 reports the observed (U.S. Bureau of the Census 2007)—where proportion of unionized workers and an growth in inequality and decline in union adjusted series that corrects for error in density was largest. Using data from the May reported union status in the CPS. Adjusting and Merged Outgoing Rotation Group files of for reporting error, only 8 percent of private the CPS, we measure inequality by the vari- sector men and less than 4 percent of private ance in log hourly wages for men and women sector women were union members by 2007. working full-time in private sector jobs. From Scholars often attribute union decline in the 1973 to 2007, men’s wage inequality increased United States to changes in the economy and by 40 percent, from .25 to .35, with most of intensified political conflict in the workplace the rise unfolding from 1978 to 2000 (see (Farber and Western 2001; Freeman and Medoff Figure 1). Women’s wage inequality increased 1984; Goldfield 1987). In this account, union

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(a) Men 0.40

Total variance 0.35 Between group

0.30 Within group

0.25

0.20

0.15 Variance of Log Wages

0.10

0.05

0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

(b) Women 0.40

Total variance 0.35 Between group

0.30 Within group

0.25

0.20

0.15 Variance of Log Wages 0.10

0.05

0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 1. Inequality in Hourly Wages among Full-Time, Private Sector Men and Women, 1973 to 2007 Note: Figures are calculated from the May and Merged Outgoing Rotation Group files of the CPS.

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40

Observed unionization 35 Adjusted for misclassification

30

25

20

Men

Percent Unionized 15

10 Women

5

0 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 2. Union Membership Rates among Full-Time, Private Sector Men and Women, 1973 to 2007 Note: Rates are unadjusted and adjusted for reporting error in union status in the CPS. firms could not respond to 1970s stagflation, Board’s rightward shift under Republican industry deregulation, and economic globaliza- administrations made organizing more diffi- tion. The biggest driver of decline in the per- cult in the 1980s. Political defeats in the centage unionized was employment growth 1970s and 1980s yielded a set of “enervated” outside the traditional union strongholds of labor laws that enabled employers to block manufacturing, construction, and transportation, organizing campaigns and weaken existing utilities, and communications. Faced with a unions (Cowie 2010:288). newly competitive economic environment, employers in unionized industries intensified their opposition, and union employment and Unions and Inequality new organizing—at least through union elec- Union decline and rising inequality has moti- tions—plunged through the 1980s (Hirsch vated research on the share and the shape of 2008; Tope and Jacobs 2009). the union wage distribution. We extend this Employer opposition unfolded in an research by linking unionization to wage increasingly adverse political context for inequality among nonunion workers. labor. Hacker and Pierson (2010) report that an influx of corporate donations influenced Wages in the Union Sector policymakers to oppose pro-union reforms of labor law in the 1970s. Union reform efforts Research on unions and inequality has were defeated, and the National Labor Relations focused on two effects. First, unions raise

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 Western and Rosenfeld 517 wages among less-educated and blue-collar spillover but increase nonunion wages with workers. This between-group effect of unions threat. Empirical studies tend to support the reduces educational and occupational inequal- threat effect, showing that nonunion wages ity. Second, collective bargaining standard- are higher in highly unionized industries, izes wages within firms and industries. This localities, and firms (Farber 2005; Leicht within-group effect of unions on inequality 1989; Neumark and Wachter 1995). reduces the spread of wages among union mem- The theory of union threat has distribu- bers with similar characteristics. Comparing tional implications. If unions threaten to these effects, Freeman (1980) found that organize low-wage workers, employers may unions reduce wage inequality more within raise wages, thereby equalizing the wage than between educational and occupational distribution. Testing this theory, Kahn and groups. Because nonunion women are concen- Curme (1987) estimated the effect of indus- trated in occupations with low and relatively try unionization on the variance of nonunion equal wages, unions’ effect on within-group wages for detailed industries and occupa- inequality is largest among men (MacPherson tions in the 1979 CPS. Consistent with the and Stewart 1987). equalizing effect of union threat, they found Unions’ within- and between-group effects less inequality among nonunion workers in suggest union decline is associated with ris- highly unionized industries (cf. Belman and ing wage inequality. Scholars estimate that Heywood 1990). declining unionization explains 10 to 20 per- cent of the growth in men’s wage inequality Unions and the Moral Economy from the late 1970s through the late 1980s (Card 1992; Freeman 1993). DiNardo and The theory of union threat takes a rationalist colleagues (1996) studied the top and bottom view of employers and a minimalist view of tails of the wage distribution separately and labor market institutions. Employers mini- found that deunionization is chiefly associ- mize labor costs and only raise wages when ated with top-tail inequality; union decline threatened with even greater pay increases explains nearly a third of the growth in the through unionization. Institutions are con- gap between the median wage and the 90th ceived minimally in the sense that unions are percentile. In their analysis, deunionization the key distortion in an otherwise competitive accompanied the declining middle of the pay labor market. distribution. We relax these assumptions, arguing that the labor market is embedded in a moral economy in which norms of equity reduce Unions and Nonunion Wages inequality in pay. The moral economy con- Union wages have been the main focus of sists of norms prescribing fair distribution research on inequality, but organized labor that are institutionalized in the market’s for- also affects nonunion workers. Economists mal rules and customs. In a robust moral often contrast the effects of spillover and economy, violation of distributional norms threat. When unions raise wages for their inspires condemnation and charges of injus- members, employers may cut union employ- tice. We often think of the moral economy ment, forcing unemployed workers to find historically—determining, for example, fair jobs in the nonunion sector. Spillover of prices for bread and flour under the British workers into the nonunion labor market Corn Laws (Thompson 1971) or the relative causes wages to fall. The threat effect results rank and standing of English workers in the from nonunion employers raising wages to nineteenth century (Polanyi [1944] 1957).1 the union level to avert the threat of unioniza- Unions are pillars of the moral economy in tion. The two theories yield opposing predic- modern labor markets. Across countries and tions: unions reduce nonunion wages with over time, unions widely promoted norms of

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 518 American Sociological Review 76(4) equity that claimed the fairness of a standard and Rubin 1978).2 Union political pressure rate for low-pay workers and the injustice of also reaches beyond wages to social legisla- unchecked earnings for managers and owners tion. For example, major unions regularly (Hyman and Brough 1976; Webb and Webb backed proposals for universal healthcare and 1911). Comparative researchers emphasize supported the creation of Medicare in the the role of distributional norms governing mid-1960s (Derickson 1994). In the 1970s, European industrial relations (Elster 1989b; unions joined with states in litigation oppos- Swenson 1989). The U.S. labor movement ing cuts to the federal food stamp program, never exerted the broad influence of the Euro- and they threatened to sue again in 1980 to pean unions, but U.S. unions often supported keep the program solvent. norms of equity that extended beyond their Institutionally, U.S. industrial relations own membership. In our theory of the moral often extend union conditions to nonunion economy, unions help materialize labor mar- workers. When a third of the male labor force ket norms of equity (1) culturally, through was organized, unions were national eco- public speech about economic inequality, (2) nomic actors who shaped centralized wage politically, by influencing social policy, and policy. During World War Two, government (3) institutionally, through rules governing boards with business and labor leaders helped the labor market. set wage standards to control inflation and Culturally, industrial unions often use a assist wartime production. Centralized wage language of social solidarity in public dis- policy continued during the Korean War. The course and within firms. Walter Reuther’s tripartite Wage Standardization Board moni- postwar leadership of the United Auto Work- tored wage increases among key firms and ers (UAW) provides a key example. Labor aimed to narrow inter-firm wage differentials historian Nelson Lichtenstein (1995:300) and reduce wage inequality in the wider econ- writes that Reuther aimed to “reshape the omy (Ross and Rothbaum 1954). In the consciousness of millions of industrial work- 1960s, unions influenced national pay policy ers, making them disciplined trade unionists, when the Kennedy and Johnson administra- militant social democrats, and racial egalitar- tions set wage and price targets to stabilize ians.” The UAW sought to develop a network prices and promote the “distributional equity” of union-based community organizations, of wages (Ulman 1998:170). The Nixon published hundreds of newspapers, and administration also adopted a tripartite wage pressed Presidents Kennedy and Johnson on policy. Concluding that “no program works civil rights legislation (Boyle 1995; Lichten- without labor cooperation,” (Matusow stein 1995). Within firms, unions have been 1998:160), Nixon’s national pay board urged voices for equity, protesting the pay of upper wage restraint in major contract negotiations management. Consistent with the egalitarian but also examined executive pay levels, sup- effect of union advocacy, studies of data from ported raises for low-wage workers, and mon- the mid-1970s through the early 2000s find itored merit pay increases (Mitchell and that managerial compensation is lower in Weber 1998). In the 1970s, President Carter unionized firms, and managerial employment pursued a national wage policy with union is lower in highly unionized industries representation in a Pay Advisory Committee (DiNardo, Hallock, and Pischke 1997; Gomez that set industry wage and price guidelines. and Tzioumis 2006). Unions also helped establish pay norms in Politically, U.S. unions have been frequent local labor markets. In some industries, union advocates for redistributive public policy. influence was amplified by law. The federal Highly unionized states have higher mini- Davis-Bacon Act and its state-level variants mum wages, and their congressional repre- require public construction projects to pay at sentatives are more likely to support minimum least the locally prevailing wages and fringe wage increases (Cox and Oaxaca 1982; Kau benefits. Studies from the 1970s show that

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Davis-Bacon raised construction wages in Consensus that followed. The era of the nonunion firms and reduced the difference Treaty of Detroit was named for the landmark between union and nonunion wages (Goldfarb wage agreement of 1948 between General and Morrall 1981). Motors and the UAW that provided an annual Beyond federal contractors, major union wage increase of 2 percent plus cost-of-liv- agreements also set the pattern for industry- ing. Wages across manufacturing industries wide wage increases. In the 1960s and 1970s, moved broadly according to this formula until employers used industry wage surveys to 1980. The Treaty of Detroit was succeeded by reduce wage differences between union and the Washington Consensus, an era of deregu- nonunion firms (Dunlop 1977; Foulkes 1980; lation and eroded pay norms in which earn- Jacoby 1997). Of course, adoption of union ings inequality increased as managers’ and standards in nonunion firms was often professionals’ compensation rose. intended to prevent unionization. Still, non- Because union decline varies across regions union companies in the 1970s closely moni- and industries, we view the transformation of tored union contracts even in lightly unionized the moral economy not as a discrete turning industries where the threat of unionization point but as a gradual process unfolding un- was remote (Foulkes 1980). Predominantly evenly across the labor market through the nonunion firms with small union workforces 1980s and 1990s. Our research design exploits abandoned merit pay, and norms of equal this variation to study how deunionization is treatment governed distribution of fringe ben- associated with rising wage inequality among efits. Summarizing his survey of pay practices nonunion workers in regional labor markets. in large nonunion firms, Foulkes (1980:153) writes: Limitations and Rival Explanations In many environments, providing and dem- We can directly observe the union wage dis- onstrating equity generally means that a tribution’s contribution to wage inequality, company’s pay rates favorably compare but union threat and norms of equity are with those of unionized companies. It would unobserved mechanisms. In our approach, be acceptable to say that the activities of union threat and egalitarianism in the moral many unions in the United States are bene- economy are indicated by industry-level fiting many nonmembers; in other words, unionization in a given region. This is an indi- unions are doing much good for people who rect measure compared to direct observation do not pay them any dues. of the cultural, political, and institutional channels of union influence on nonunion The slow postwar decline of unionization wages. Still, the industry-region unionization rates accompanied the erosion of the labor rate captures in a single index organized market’s moral economy. The cultural, politi- labor’s salience in the surrounding labor mar- cal, and institutional indicators of equitable ket. We can measure the index across the pay norms declined with union membership. national labor market over decades, and it Many researchers see 1981 as a watershed usefully captures the uneven decline of U.S. year, when the Reagan administration defeated labor as a voice for equity in the economy as the air traffic controllers’ strike by hiring per- a whole. manent replacements. Voss and Sherman Despite these virtues, industry-region union- (2000:311) characterize the turning point as a ization is likely correlated with economic and change in norms when “corporate leaders social conditions that are also associated with stopped playing by the rules.” Levy and Temin rising inequality. In particular, increased (2011) similarly divide the U.S. labor market’s demand for highly skilled workers may raise postwar history into the eras of the Treaty inequality and reduce unionization. Scholars of Detroit until 1980 and the Washington regularly interpret the rising demand for skilled

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 520 American Sociological Review 76(4) workers, indicated by the increasing college likely remains and this affects our interpreta- wage premium, as resulting from technologi- tion of the results. cal changes in which computerization replaces routine work tasks (Autor and Dorn 2009; Autor, Levy, and Murnane 2003). In some Data and Methods accounts, firms adapt to technological change We link union decline to rising inequality by by introducing performance pay and other decomposing the variance of log wages. The workplace incentives, weakening internal decomposition uses a variance function regres- labor markets and increasing inequality within sion in which the mean and the variance of an firms (Bloom and Reenen 2010; Cappelli outcome depend on independent variables, 2001). Although skill-biased technological providing a model for between- and within- change has been scholars’ central focus, econ- group inequality (Western and Bloome 2009). omists also argue that immigration and trade The decomposition is based on a regres- have shifted demand away from low-skill sion on the log hourly wage, y , for respond- i workers, further contributing to college gradu- ent i (i = 1,...,N) for a given year of the CPS. ates’ wage premium (Card 2009; Cline 1997). Our key predictors are an indicator for union Union decline may have enabled or influ- membership, u , and a continuous variable, i enced technological and organizational change ū , that records for each respondent the i and economic globalization. Merit pay, for unionization rate for the industry and region example, is less common in union firms in which they work. Covariates, including (Lemieux, MacLeod, and Parent 2009). Evi- schooling, age, race, ethnicity, and region, dence from a plant restructuring indicates that are collected in the vector, x . The model i unions shape the introduction of new tech- includes equations for the conditional mean nologies to moderate pay inequality and limit of log wages, layoffs (Fernandez 2001). Deunionization ^ ′ may thus have indirect or mediated effects on yui =+xiiα12αα+ ui 3 inequality beyond the effects of union threat or normative influence. and the conditional variance, Alternatively, unions flourish in concen- 2 ′ trated, protected, smokestack industries logσβii=+x β12uuii+ β3 . employing relatively homogeneous work- forces. Technological change, human resource We expect union membership and industry- management, and economic globalization region unionization to have positive effects may undermine workplace solidarity, fueling on average wages, but negative effects on the deunionization and wage inequality. Declin- variance of log wages. Economy-wide changes ing unionization and rising inequality may in average wages and within-group inequal- share common roots in shifting market condi- ity, perhaps due to general shifts in norms or tions. Effects we attribute to union decline the macroeconomy, are captured through the may really be due to other changes in the regression intercepts. economy. We measure between-group wage inequal- We account for some rival explanations in ity by the variance of the conditional means, a regression that controls for education, N demographic characteristics, and region. The 2 Bw=−()yy^ , (1) increasing education gradient accompanying ∑ i i i =1 skill-biased technological change, immigra- tion, and trade is directly incorporated into the where w is the sample weight for respondent i analysis. The regression also allows a com- i, normed to sum to 1, and y– is the grand parison of effects of deunionization versus mean of log wages. We measure within-group effects of education. Still, omitted variable bias inequality by the residual variance,

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N 2 union decline and educational inequality on Ww= ∑ iiσ , (2) the other. Adjusted variances might be inter- i =1 preted as counterfactual measures of inequal- That each respondent has a variance, σ 2, may ity, prevailing if unionization rates or i be counterintuitive, but it is simply estimated regression coefficients remained fixed at by the squared residual from the regression on 1973 levels. The counterfactual assumes, y . Because unions are mostly associated with however, that regression coefficients accu- i within-group inequality, we expect deunioni- rately estimate causal effects, that all other zation to be closely associated with an predictors and their coefficients change as increase in within-group variance, W. Sum- observed, and that there are no broader effects ming within- and between-group components on employment of fixing unionization rates or yields the total variance, a measure of overall other quantities. inequality: Counterfactual interpretation is likely implausible because of omitted variable bias. V = B + W. Because we analyze the change in inequality rather than its level at a point in time, biases We decompose the rise in inequality with of constant magnitude will not affect our three adjusted variances that fix in a baseline analysis of the trend. However, if union mem- year either the coefficients or the distribution bers are increasingly positively selected, of some predictors. First, focusing on individ- upward bias in wage effects will grow because ual union membership, we calculate the level of union workers’ rising productivity. In the of inequality assuming unionization had context of rising inequality, shifts in technol- remained at its 1973 level. We estimate this ogy, industry regulation, and the use of per- compositional effect by reweighting the data to formance pay, for example, might be preserve the 1973 unionization rate across all correlated with industry-region unionization years, from 1973 to 2007. We then use the and wages. Industry and region fixed effects adjusted weights in Equations 1 and 2 to calcu- might reduce omitted variable bias, although late adjusted variances for all years. Second, fixed effects account for nearly all the varia- we add the effects of union threat and equitable tion in industry-region unionization, leaving wage norms on nonunion workers by calculat- little variance for the effects of interest. We ing wage inequality assuming industry-region try to reduce bias with region fixed effects. unionization and its coefficients remain con- Given a clear trade-off between population stant at the 1973 level. Fixing industry-region description and causal analysis, we prefer this unionization coefficients, and reweighting to simple model because our interest centers on hold union density constant, yields adjusted the population trend in rising wage inequality. values of ŷ and σ 2 that are plugged into Equa- We aim to show how the wage distribution i i tions 1 and 2 to obtain adjusted measures of moves with shifts in unionization and other between- and within-group inequality. Finally, labor market conditions, rather than provid- we assess education’s contribution to rising ing a causal analysis of wage determination. inequality by fixing education coefficients for Measurement error in union membership the mean and variance equations at their 1973 also adds bias. Using data from a 1977 employer values. Fixed education coefficients adjust the survey, Card (1996) found that 2.5 percent of values of ŷ and σ 2 and the resulting measures CPS respondents misreport their union status.3 i i of between- and within-group inequality (see With this estimate of measurement error, at 50 Appendix A for more details). percent unionization, the number of nonunion Adjusted variances quantify the growth in workers misclassified as union equals the num- inequality statistically attributable to changes ber of union members misclassified as nonun- in unionization rates or changes in coeffi- ion. With observed unionization at 10 percent, a cients. Adjustments describe the association large number of nonunion workers (2.5 percent between rising inequality, on one side, and of 90 percent) report they are union members.

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Given a misclassification rate, λ, and observed and the proportion top-coded varies over sur- unionization, ū, the true unionization rate is (σ 2 veys, biasing estimates of inequality. For each i – λ)/(1 − 2λ). With observed unionization at .10 year, we impute the top 2 percent of wages (10 percent) and a misclassification rate of from a Pareto distribution. Other common λ = .025 (2.5 percent), the true unionization rate methods for top-codes yield similar results to is .079, a measurement error of about 20 per- those reported here. Incomes greatly increased cent. Measurement error biases estimates of in the top percentile since the 1990s, a trend union effects and increases with declining measured by administrative rather than sur- unionization. vey data (Piketty and Saez 2003). Similar to We account for measurement error in union other research on wage inequality (Autor status by augmenting the likelihood for the et al. 2008; Mouw and Kalleberg 2010), we wage model with an extra term for the probabil- cannot analyze the top fractions of a percen- ity of misclassification (for other approaches, tile of the income distribution with these data see Card 1996; Hirsch 2004). Because the mis- (for coding details, see the online supplement classification rate may vary across surveys, we [http://asr.sagepub.com/supplemental]). specify an average rate of 2.5 percent but allow Table 1 reports descriptive statistics for the it to vary from 2 to 3 percent with a Bayesian CPS data. Wage stagnation is indicated by prior distribution. Correcting for misclassifica- trends in average pay for union and nonunion tion significantly reduces bias in the estimated men. Average pay increased for women, reflect- union effects (see Appendix B). ing women’s entry into professional and other We compiled data from the annual May skilled occupations. Inequality trends are simi- files of the CPS from 1973 to 1981 and the lar for men and women, increasing everywhere annual Merged Outgoing Rotation Group and most rapidly among union workers. Rising files of the CPS from 1983 to 2007 (Unicon wage inequality among union workers tends to Research Corporation various years; U.S. reduce the effect of union decline on inequality, Bureau of the Census various years). We as the union wage distribution comes to resem- exclude 1982, when union questions were ble the nonunion distribution. Educational omitted from the survey, and 1994 and three- attainment also increased across the labor mar- quarters of 1995, when allocation flags for ket, as the proportion of high school dropouts in wages were missing. The analysis includes men the workforce dropped from around 1 in 4 and women working full-time (i.e., 30 hours a workers in the 1970s to about 1 in 10 by the week or more) in the private sector. 2000s. Fewer than 20 percent of private sector The dependent variable is log hourly workers were college graduates in the 1970s, wages adjusted for inflation to 2001 dollars. but more than a quarter had college degrees Several adjustments improve the quality and three decades later. The college graduation gap continuity of the wage data. Nonresponse to between union and nonunion workers also nar- wage and income questions increases over rowed from the 1970s to the 2000s. time, and by 2007, about a third of CPS work- Other covariates might be analyzed, but ers did not report wages. The CPS imputes coding inconsistency and missing data pre- wages to nonrespondents, but regression vent us from greatly augmenting the model. coefficients for nonmatched criteria are atten- In additional analyses, we explored the effects uated and the residual variance for wages is of marital status, Hispanics, urban residence, sensitive to the imputed data (Hirsch 2004; and occupations. Results are largely unchanged Mouw and Kalleberg 2010). We thus omit for men. There is greater confounding for imputed wages from the analysis.4 Earnings women, but their union effects are relatively imputation flags change across surveys, and small in any case. we follow Hirsch and Schumacher (2004) to To estimate the effect of unions on non- create a consistent series of nonimputed earn- union workers, earlier research measures union- ers. A few respondents with wages less than ization in industries and localities (e.g., Freeman one dollar are excluded. Top-codes for wages and Medoff 1981; Neumark and Wachter 1995).

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Table 1. Descriptive Statistics for Analysis of Unionization and Hourly Wage Inequality among Private Sector Men and Women, Working Full-Time, CPS, 1973 to 2007

1973 to 1984 1985 to 1995 1996 to 2007

NUNUNU

Male Workers Log Hourly Wages Mean 2.84 3.05 2.81 3.04 2.87 3.07 Variance .29 .12 .32 .14 .35 .19 Level of Schooling (proportion) Less than high school .23 .31 .14 .17 .13 .10 High school graduate .33 .47 .35 .51 .31 .46 Some college .23 .17 .24 .24 .26 .33 BA or more .21 .04 .27 .07 .29 .11 Demographic Characteristics (proportion) White .86 .83 .83 .81 .76 .77 Black .07 .09 .06 .09 .06 .08 Other .08 .08 .11 .10 .18 .15 Age (years) 36.11 39.04 36.39 40.56 38.79 42.43 Female Workers Log Hourly Wages Mean 2.46 2.66 2.56 2.74 2.67 2.82 Variance .19 .12 .24 .18 .29 .23 Level of Schooling (proportion) Less than high school .20 .33 .10 .17 .08 .10 High school graduate .47 .48 .40 .48 .32 .36 Some college .22 .14 .29 .22 .33 .30 BA or more .11 .05 .20 .13 .27 .23 Demographic Characteristics (proportion) White .84 .74 .81 .68 .76 .64 Black .09 .16 .09 .18 .09 .16 Other .07 .10 .10 .14 .15 .20 Age (years) 35.59 38.83 36.41 40.05 39.20 42.50

Note: N = nonunion workers, U = union workers.

Indeed, regions and industries are two key west, and the West). Regional disaggregation dimensions along which workers and employ- helps account for time-varying changes in ers make wage comparisons (Foulkes 1980; regional pay scales and allows union strength to Hyman and Brough 1976). To capture the influ- be spread unevenly across the country. Smaller ence of unions on the nonunion sector, we spatial units, like states, might correspond better measure unionization in 18 industry categories to local labor markets, although a full set of state combining broad (one-digit) sectors with sev- identifiers are available only since 1977 and eral detailed (two-digit) industries. We recode region codes are available for the whole series. the several revisions of CPS industry classifica- In large capital-intensive industries where tions to produce a relatively consistent series unions are concentrated, pay norms, like union- whose distribution changes smoothly across ization, are also likely to stretch across state survey redesigns. We disaggregate further by lines. measuring industry unionization in four census Dividing the national labor market into 18 regions (i.e., the Northeast, the South, the Mid- industries by four regions yields 72 annual

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Industry Transportation Manufacturing G Transportation G Utilities and Sanitation G Retail Grocery G Communications G Non−durable Manufacturing G Construction. G Durable Manufacturing G Mining G Printing and Publishing G Entertainment and Recreation G Repair Services G Wholesale Trade G Professional Services G Business Services G Retail Trade G Finance, Insurance, and Real Estate G Agric., Forestries, Fisheries G

Region Midwest G G Men Northeast G Women West G South G

010203040

Percent Unionized

Figure 3. Average Unionization Rates for Full-Time, Private Sector, Male and Female Workers, in 18 Industries and Four Regions, 1973 to 2007 industry-region unionization rates. On aver- threat, workers’ skills in different industries, age, unionization rates were highest in utili- or industry rents. Figure 5 shows preliminary ties, transportation, and communications, support for unions’ waning normative influ- particularly in the Northeast and the Midwest ence on nonunion wages. For men and (see Figure 3). Unionization was lowest in women, nonunion wages are more com- agriculture, finance, and retail trade, and pressed in local labor markets that are highly throughout the South. Throughout the coun- unionized, and highly unionized local labor try, women’s unionization rates in the private markets become less common over time. sector were lower than men’s. We find evidence for the relationship between industry-region unionization and Results wages by plotting unionization for the 72 To study the effects of declining union member- industry-regions’ nonunion wages for each ship on wage inequality, we fix the unionization year from 1973 to 2007. Average nonunion rate at 1973 levels. Similar to earlier research, wages are higher in regions and industries we find the largest effect of declining union where unionization is higher (see Figure 4). membership on men’s within-group inequal- Higher wages may reflect the effect of union ity (see Figure 6, panel a). The observed

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(a) Men 3.5 GG G G G G G G G G GGG G G G GG G G G G G G G G G G G G G G G G G GG GG G G G G GGG G GG GG G GG GG G GGG G GG GG GG G G G G G GG G G G G G G G G G G GG GG GG G G G GGG GG G GGG G G GG G G GG GGG GG G G G GGGG G GG G G G G GG G G GGG G G G G G G G G GGG GGGGG GG G G G GGGG G G G GG GG GGG G GG G GG GG GG G GGG G GGGG G G G GGGGG G G G GG G GG G G G G GG G G G GGGG G G G GG G GG GG GGGGG G GGGG G G GG G G GG G G GG G GGG GGGGGG GG G G G G G G GGGG GGGG G GG GG G GG G GGGGGGGG GGG GGGG GG GG G GG G GG GGG G G G G G G G GGGGGGG G GG GGGGGGGGG G G G GG G GG GGGGG G G G G G G GGGGGGGGG G GG GG GGGGG GGGGGGGGG GGGG G G G GGG GGG GG G G G GG G GGGGGGGG G GG G G G GG G GG G GGG G G G G G GGGGGGGGG GGGG GGGGGG GGG GGGG G G GGGGG G G G G GG G G G G G G G GGGG G GGG G GGGGGGGG GGG G G GG GGGG G G G GGGG GG G G GGGGG GG G G G GG GGGG GGGGGGGG G GGGGG G G GG GGG G G GGGGG G G G G G GG G G G GG G G GG G G GGGGGG GGGGGG GGGG G G GG GGG G G G G G G GG G G G G GG GGGGGG GGG GGGGGGGG GG GGGGG G G GGG GG G G G G GGG G GGG GGGGG G G G GGGG GGGGG GGG G GGGGG G GGG GGG G GGG GGG G G GG GG GG G GG G GGGGGGG GGGGGGGGGGGGGGG G GGG GGGG GGGG GG GGGGG G G G GG G G GGG G G G G G 3.0 G GG G GGGG GGG G G G GG G GG G G G G GG GG GGGGGGG GGGGGGGG GGG GG GGGG GGG GGG G G GG GG GGGGGG GGG G GGGGG GG GG GGGGGGGGGG G G GGGG GG GGGGG G G GG GG G G G G GG G GG G G GGGGGG GGG GGGGGGGGG GGGGGG GGG G GGG G GGGGG G GG G G G GG G G G GG GGG GGGGGGGGGGG G GG GGGGGGG GGGG GGGGGGGGGG G G G G GG G G GGG G GGGGGG GGGGGGGGGGGGGGGG G GGG G G G G G G G G GG G G G G ge G GG GGGGGGG GGG GGG GGGG G GGGGG G GGG G G G GG G GGGGGGGGGGG GG GGGG GGG GGGGG GGGGGG G GGGGGGGG GG GGGG GGG G G G GGGGGGGGGGGGGGG GGGGGGGGGGG GGGGGG G G GGGGG GGGGG G G GG G G G GGG GGGGGGGGGGGGGGGGGGGG GGGG G G GG GGGGGGGG GGGG G G G G G GG GGGGGGGGGG G GGGGG GG GG GGGGGG GG GGGG GGGGGG G GGG G G G G G G GG GGGGGGGGGGGG GGGGG G GGGGGG GGGGG G GGG GGG GG GGG GGG G G G G GGGG GG G GG G GG G G GGGGGG G G G G GGG G G G GGGG G GG GG G GG GG GGGG G G GGG GG GG G GGGGG G GGGG G GG G G G GG Wa GG G G G GGGG GG G GGGGGG G GGGGGGGG GG G G GG G G GGGGGGG GGGGGGGGG GGGG GG G GGG G GG GG G G G GGGGGGGGGGGG G GG GG GGGG G G GG GG GGGG GGGGGGGGGGGG GG GG G G G G GG G G GG G GG G G GGG GG GGGGGG G GGG G G G G GGGGGGGGG GG G G G G G G GGG G G GG G G GGGGGGGGGG GGG GG GG G G G GG G G GG G G G GGGGGGGGGGGG GG GGGG G G G G G G G G G G GGGGGGG GGG GG G G G GG G GG G G G G GGGGGGGGGG GGG G GGG GG G GG G G GG GGGGG GGG GGGGGG GG G G G GG GGGGGGGG G GGG GGG G G G GG G GG G G G G GGGG GGGGGGGG G GGGGG G G GGGG GGGGG G G G GGG G G G GGG GG GGG GGGG GG G G GG G G G GG GGGG GGGG G G G G G GGGG GGG G G GGGGG GGGG G G GGG G GGGGGGGGGG G G GG G GGG GGGGGGG G G GGG G G G G GGGGG GGG G G GG G GG G GGG G G GG GG G Mean Log GG GGG G GG GG GGGG G G G GGGGGG G GGG G G GGGGG GG G G GGGGGGG G GG GGG G GGG GG GG G G GG G G 2.0 2.5

01020305040 60 70

Unionization Rate

(b) Women

G 3.5

G G G G G G G GG GG G G G G G G G G G G G G GG G G G G G GG G GG G G G G GG G G G G G GG G G G G GGG G G G G G G G G G G G G GG GG GG GG G GG G G G G G G G G G GG G GG GG G G GGGG GG GG GG GGGGGGG GG G G G GG GG G G G GG GGG GG GG GG G GG G GG G GG G G G G G G G G GG G GG G GGG G G G G 3.0 G G G G G G G G GG G GGG G G G G G G GGG G GG GG GG G G G GGG GG G G GGGG G G G G G G GGG G G G G GGGG G G GG GGGGG GGGG G GGGG G G GGG GGGG G G G GGGG GGG GGG G GG G GG G G GG GG G GGGGG G GGGGGGG GG GG GG G GG G GG GGGG GG G GG G ge G GG GGG GG GG G GG GG G GG GG GGG GGG G GGGGGGGGG G G G G G GG G G G G GG GG G G G GG GGGGGGG G GGG GG GGGGGG GG G GG GG G GG G G G G GG G G GGG G G GGGGGG G G GGGG G GG GG GG G GG GG G G G GG GG GGGGGGGGG GGG GGGG G GG G G G GG GGGG G GGGG G GGGGGGGGG G GG GG G GG G G GG G G G G G G G G GGGGGGGGG GGGGG GGGG GGGGGGG G GGGG G G GGGGG GGG G GG GG GG G GG G G GGGGGGGG GGGGGGGGGGG GGGGGG GG GGG GG GGG G GG GG G G G G G G G G Wa GGGGGGG GG GGGGGGGG GG G GGGGGG G GGG G GGG GG G G G G G GG G GGG G G GGGGGGG GGGGGGGGGGGG GGG GGGGGG GGG GG GGGGGGGGG G GG G GG GG GG G GG GGGGGGG GGGGGGGG GGGGG GGGGGGGG GGGGG GG GGGGG GGG G GG GG GGG G G G GG GG GG G G GGGGGGGGGGGGGGGGGGGG GGG GGGG G GGGG GGG GGGGGGGG GGG GGGG G G G G G G G GG G G GGG GGGGG GGGGGGGGGGGGG GGGGGGGGG G G GGGG GG GG GG GG G G GG G G G GG G G GGGGGGGG G GGGGGGGGGGGGGGG G GG GG GG G GGGGGG GGG G G G G GGGGGGG GGGGGGG GGG GGGGGGGGGGG G GGGGG GGGG GGGG GG GG GGG G G GG GGG GG G G G G G GGGGGGGGGGGGGGGGGGGG GGGGGG GG GG G G GGGGGGG G GG GGGG G GGG GGGG G GG G G G G G GGGGGGGGGGGGGGG GGG G GGGGGG GGG GG GGG GGGGG GG GGGGGGGGGGGGG G GG GG GG G G G GGGGGGGGGGGGGGGGGGGGGG GGGGG G GGGGG GG GGG GG G GG GG G GG G G G G G G GGGGGGGGGG G GGGGGGGGGGGGG GG GGGG GG G GGG G GGGG G GG GG G GG G GGGGGGGGGGGGGGGG G G GGGG GGG GG GG G GG G GGG G G GGGG G GGG G GGGGGGGGGGGGGG GG GGGGGGGGGGGGG GG G G G GG G GGGGG G G GGGGGGGGGG G GGGGGG GGG GG G G G GG GGG GG GGG G G G G G G G GGGGGGGGG GGGG GGG G G G G G GG GGG G G G G GGG GGG G GG G GGG GG G GGG G G GG G GG GGGGGGGGGGGG G G G G GGG G G GG G GGGGGGGGG GGG G G G GG G GG G G G G G GGGGGGGGG GG GG G GG GG G G GGG G G GG GG G G G G GGGGGGGG GGG G G G G G GGG G GG G G GG GG G G G GGG GGG GG G G G G GG GGG G GGG GG G G G G G G GGGGGGGGGGG G GGGGGG GGGG G G GGG G G GGG GG GGGG GGG GGG GGGG G G G G GGGGG G G G G GGG GGGGGGGGGGG G G G GGG G GG GG GGGGGGGGGGGG GGGGGGGGG G G GG GG G G G GG GG G G GGGGGGGGGGGG GGGGGG G G G G G GG Mean Log G G GG G G G G G GGG GGGGGG GGGGGGGGG G G G G G GG G G G G GGGGGGGGG G G GG G G G GG G GGGGGGGGG G GGG G GGGGG G G G G G GG G GGGG G G G G G G GGGGGGG GG GGGGGG G GGG G GG G G GGG G G G G GGG GG G G G G G G GGGG G GGGG GGG GG G G G GGGG G G GGG GG G GGGGGGG GG G G G G GG GG G GG G G G GG G

2.0 2.5 GG G G G GGGG G

G

01020305040 60 70

Unionization Rate

Figure 4. Unionization Rates and Mean Log Hourly Wages for Full-Time, Private Sector, Male and Female Nonunion Workers, by 72 Industry-Regions, 1973 to 2007 Note: Points go from light gray to black, from 1973 to 2007.

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(a) Men 1.21.00.80.60.40.20.0 ge Wa

G

G

G G G G G G GG G G G G G GG G G G G ance of Log GGG G G G GGGGGGGGG GG G GGGGG G GGGGGGGG GGGGGG G G G GGGGGGGGGGGGGGGGG G GGG G G ri G G G GG G G GGGGGG GG GGGGG GG GGG G G GGGGGGGGG GGG GGGGGGGGGGGGGGGG GG G G G G G GGGGGGGGGGGG GGGG GGG GGG GGG G G G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG G G G GG G GGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGG G GGGG G G G G GGGGGGGGGGGGGGGGGGGGGGG GGGGG G G GGGG G GG G G G G G Va GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG GGG G GG G G G G G G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GG GGGG GGGGGGGGGG G G G G G G G GG GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG G G G GGGGGGGGGG G G GG GG G GG GG G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG GGGG G GGGGGGGG GGG GGG GGGGGGGG GG GGG GG GGG G GG GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGG G GGGG GGGGGGGGGGGGG G GGG GG GG GGGGGG GG G G GGGG GG G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGG GGGGGG G G GGGGG GGGG GGGGG GGG GG GGG G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG GGGGGGGGGGGGG GGGGG GGGGGGGGGG GGGG G G G G G G G GGGGGGGGGGGGGGG GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG GGGGGGGGGGGG GGGGGGGG GG GGGGGG GGG GGGGG GGG G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGG G GGGGGGGGGGG GGG G GG G GG GGGGGGG GGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGGGGGG GGG GGGGGGGGGGGGGGGGGGG GGGGGGG G GGG G GG GG GGG GG GG G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG GGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGG GGGGGGGG GGGGGGGGGGG GG GGGGG G GGGG G G GGGG G G GGGGGGG GG GGGGGGG GGGGGGG GGG GGGGGGGGGGGGGGGG GGGGGGG GGGG G GGGGGG G G GG G G G GG GGGG G GGGGGGG GG G G GGGG GGGG GGGGGGG GG GG G GGGGG GGGGG GGGG GGGGGGGGGGG G G GG G GG G G GGGGGGGGGGG G G G GGGG G GGG GGGG GG G GGG GGG GGG GG G GGG G G GG G G GGG G GG GG GG G GG GGG G G G GGG GG G GG GG G GG GGG G G GGGGGGG G G GG G G G G G GG GG G GGG G GG G G G G G G GGG G G GGG G GG G G G G GG G G G G G G G GG G G GG G G G G G G G G G G GG G G

01020305040 60 70

Unionization Rate

(b) Women

G 1.21.00.80.60.40.20.0 ge Wa

G G

G G G G G G G G

G G G G G G G G ance of Log G G G ri G G G G G G G G G G G G G GGGG G G G GG G G G G G GG GG G GG G GG G G GGGGGGG GG G GG G G G GG G GG GGGGGG G G GG G GG G G Va GGG GG GG G G G G G G GGGGGGGGG G G GGGGGGGGGG G GG GG G G GGGGGGGGGGGGGGGG GGG GG GGGGG G G G GG G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGG G GG GGG GGG G G G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GG G G G G G G G GGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGGGGG G G GG G G G GGG G GG G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGG GGG G G GGGGGGGG GGG G G G G G G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGG GGG GG GG G G G G G G GGG GG GG GGG GGGGGGGGGGG GGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGGGGGGGG GGGGGG G G G G G GG GGGG G G GGG G GG G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGG G GG G GGGGGGGGGGGGGG GGGGGGGGGGGGGGGGGGGGG GGGGGGG G GGG GGG G G G GG GG GG G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGG GGGGG G GGGGGGGGG GGG GG GGGGG GG G GG GGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGGGGG GGGGGGGGGG GGGG GGGGG GGGGGGGGGG GG GGGGG GG GGGGGGGGG G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGGGGG GGGGGGGGGGGG GG GGGGGGGGGGG G GG GG G GG G G G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG GGGGGGGGGG GGG GGGGGGGGGGGG GGGGGG GGGG GG GGGGGG G G GGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGGG G G GGGGG GGGG GGG GGG GG GGGGGGGGGGGGGGG GGG GGGGGGGG G GGGGGGGGGGGG GG GG G G G GGGGGGGGGGGGGGGGGGGGGGGG G GGGG GGGGGG GGGGGGG G GGGGGGGGGG GGGGGGGGGGG GGGGG G GGG GG G G GG G G GGGGG GG GGGGGGGGGGGGGGG G GG G GGGGGGGGG GG GGGGG GGGGGGGG GGGG G G GG GGG GG GGGGG G G G GG G GGGGGGGGGG G GGGGGGGGGGGGGGGGGGGGGGG G GGGGGGG G GGG G GGG GG GGGGGG G GGGGGGGGG GGGGG GGGGG GG GG G G G GGGG GGGGGGGGG GGGGGG GGGG G G GG G GG GGGGGGG GG G G GGG GGGGG GG G GGG G G GG GGGGGGGG GGGG G G GG GGGG GGGGGGGG G GG GG GGGGG G GGG GGGGG GGGGG G GG G GG G G GGG GG GG G G GGGGGGGG G GGGGGGGGG GG GGG G GG G GG G GGGG G G GG G G G G GG G GGG G G G GG GG GGG GG GG GG G GGGGGGGG GGGG G G G GG G G GGG G GGG G G GGG G G G GG G G GG G G G G G G G GGG G G G G G G G G G GG G G G G G G G G G G G G G G GG G G G

01020305040 60 70

Unionization Rate

Figure 5. Unionization Rates and the Variance of Log Hourly Wages for Full-Time, Private Sector, Male and Female Nonunion Workers, by 72 Industry-Regions, 1973 to 2007 Note: Points go from light gray to black, from 1973 to 2007.

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0.25 (a) Men

0.20

0.15

0.10

Variance of Log Wages Between-group Within-group 0.05 Adjusted between Adjusted within

0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

0.25 (b) Women

0.20

0.15

0.10 Variance of Log Wages

Between-group 0.05 Within-group Adjusted between Adjusted within 0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 6. Observed and Adjusted Within- and Between-Group Variances of Log Hourly Wages, Full-Time, Private Sector Men and Women, 1973 to 2007; Adjusted Variances Fix Union Membership at the 1973 Level

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 528 American Sociological Review 76(4) within-group variance increases by .046 points, a third of the rise in wage inequality. This but the adjusted variance (with the 1973 union- accounting, which includes unions’ effects on ization rate) increases by only .028 points. Over nonunion wages, is 50 to 100 percent larger a third of the increase in within-group inequality than the union effects on inequality reported is associated with declining union membership in other decompositions (cf. Card 2001; ([.046 − .028]/.046 = .40). Between-group DiNardo et al. 1996). inequality increases by .055 variance points Among women, the association between and the adjusted series increases by almost as union decline and wage inequality is smaller, much, by .053, indicating that union decline but greater than zero, once nonunion wages explains little of the rise in men’s between- are taken into account (see Figure 7, panel b). group inequality. Summing the between- and Declining industry-region unionization accounts within-group effects, the decline in unioniza- for over half of the increase in women’s tion from 34 to 8 percent explains about a fifth within-group inequality. Adding between- of the rise in inequality in hourly wages among and within-group effects together, union decline full-time, private sector men. is associated with about a fifth of the rise in Among women, the effect of declining union wage inequality among women. membership on wage inequality is smaller (see How do the effects of union decline compare Figure 6, panel b). Inequality grew slightly to the growing inequality of wages by educa- more for women than for men, but increasing tion? Counting union and nonunion wage women’s wage inequality is unrelated to union effects, deunionization explains about a third of membership. Within-group inequality increased the rise in men’s earnings inequality (see Figure by about .047 points among women but the 8, panel a). Increasing returns to education and adjusted series, with 1973 unionization held increasing wage inequality among highly edu- constant, increases nearly as much, by .043 cated workers explain a similar share of the rise points. Similarly, between-group inequality in wage inequality. Among women, union increased by .051 points and the adjusted decline explains about a fifth of the rise in wage between-group variance, with union member- inequality (see Figure 8, panel b); rising educa- ship fixed, increased by .054 points. Consistent tional inequality in pay explains nearly twice as with other research, holding the unionization much. In short, deunionization’s effects on ine- rate constant explains almost none of the rise in quality are only half as large as education’s women’s wage inequality. effects for women, but union and education Union decline explains more of the rise in effects are equally large for men. wage inequality once we account for the link Finally, the combined effects of union between unions and nonunion wages. The decline and education on wage inequality can link between organized labor and nonunion be calculated by holding constant the unioni- workers is captured by the effects of industry- zation rate, industry-region unionization and region unionization on the mean and variance its coefficients, and education coefficients of wages. The total effect of deunionization (see Figure 9). From 1973 to 2007, the can be measured by fixing the 1973 unioniza- adjusted variance for men’s wage inequality tion rate as before, and also fixing industry- increased only 10 percent, compared to a 40 region unionization rates and their coefficients percent increase in the observed variance. at 1973 values (see Figure 7). This adjustment Three-quarters of the rise in men’s hourly indicates the strong relationship between wage inequality is thus associated with union union decline and rising within-group in- decline and increasing inequality by educa- equality. Among men, adjusted within-group tion. Women’s wage inequality increased by inequality does not increase between 1973 nearly 50 percent, but adjusted inequality— and 2007, suggesting effects of union threat accounting for inequality by education and and eroding norms of equity on the wage dis- unionization—increased 17 percent. Decom- tribution. Deunionization’s effect on union position of women’s wage inequality indi- and nonunion wages is associated with about cates that unions and returns to schooling

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(a) Men 0.25

0.20

0.15

0.10 Between-group Variance of Log Wages Within-group 0.05 Adjusted between Adjusted within

0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

(b) Women 0.25

0.20

0.15

0.10 Variance of Log Wages

Between-group 0.05 Within-group Adjusted between Adjusted within 0.00 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 7. Observed and Adjusted Within- and Between-Group Variances of Log Hourly Wages, Full-Time, Private Sector Men and Women, 1973 to 2007; Adjusted Variances Fixed at the 1973 Level: Union Membership, Industry-Region Unionization Rates, and Industry- Region Unionization Effects

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1.6 (a) Men

1.5

1.4

1.3

1.2

1.1

1.0

Observed variance Variance of Log Wages 0.9 Adjusting for unionization 0.8 Adjusting for education effects 0.7

0.6 1970 1975 1980 1985 1990 1995 2000 2005 2010

(b) Women 1.6

1.5

1.4

1.3

1.2

1.1

1.0

Observed variance Variance of Log Wages 0.9 Adjusting for unionization 0.8 Adjusting for education effects 0.7

0.6 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 8. Total Variance of Log Hourly Wages, Full-Time, Private Sector Men and Women, 1973 to 2007; Variances Adjust for Unionization and Education; Variances Are Set to Equal One in 1973

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1.6 (a) Men

1.5

1.4

1.3

1.2

1.1

1.0

Variance of Log Wages 0.9

0.8 Observed variance 0.7 Adjusting for union and education effects 0.6 1970 1975 1980 1985 1990 1995 2000 2005 2010

(b) Women 1.6

1.5

1.4

1.3

1.2

1.1

1.0

Variance of Log Wages 0.9

0.8 Observed variance 0.7 Adjusting for union and education effects 0.6 1970 1975 1980 1985 1990 1995 2000 2005 2010

Figure 9. Total Variance of Log Hourly Wages, Full-Time, Private Sector Men and Women, 1973 to 2007; Adjusted Variances Fixed at the 1973 Level: Unionization Rate, Industry- Region Unionization Rate and its Effects, and Education Effects; Variances Are Set to Equal One in 1973

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Table 2. Summary of the Decomposition of the Change in Variance of Log Hourly Wages for Full-Time, Private Sector Men and Women, 1973 to 2007

Between Within Total

Male Workers Observed change in variance of log wages .055 .046 .102 Change in variance fixing: Union membership rate .053 .028 .081 + Industry-region unionization .070 −.003 .067 + Education effects .030 −.005 .025 Percentage of change explained: Union membership rate 3.2 40.3 20.2 + Industry-region unionization −26.9 106.3 33.9 + Education effects 45.8 110.1 75.2 Female Workers  Observed change in variance of log wages .051 .047 .098 Change in variance fixing: Union membership rate .054 .043 .097 + Industry-region unionization .059 .019 .078 + Education effects .018 .017 .035 Percentage of change explained: Union membership rate −5.2 9.2 1.7 + Industry-region unionization −16.1 59.8 20.4 + Education effects 64.9 63.6 64.3

Note: Adjusted variances are obtained by successively adding the listed effects. For example, adjusted variances in line 2 are obtained by fixing union membership, adjusted variances in line 3 are obtained fixing union membership and regional unionization rates, and so on.

together account for nearly two-thirds of the also equalize the nonunion wage distribution by rise in wage inequality. threatening union organization and buttressing Table 2 summarizes results of the variance norms for fair pay. We found strong evidence decomposition. Two findings stand out. First, that unionization rates in detailed industries for the effect of declining unionization on private geographic regions are positively associated sector wage inequality is much larger for men with wage equality among nonunion workers. than for women. This result is consistent with As unionization rates fell in the national labor the large fall in private sector unionization market, industry-region unionization rates also among men. Second, most of the rise in declined, and within-group inequality increased men’s wage inequality can be explained by among union and nonunion workers. deunionization and increasing returns to school- A variance decomposition analysis esti- ing. Educational inequality in wages explains a mated the effect of union membership decline large fraction of the increase in between-group and the effect of declining industry-region inequality. Increasing inequality in highly unionization rates. When individual union unionized industries and regions explains a membership is considered, union decline large fraction of rising within-group inequality. accounts for a fifth of the growth in men’s earnings inequality. Adding normative and threat effects of unions on nonunion Discussion pay increases the effect of union decline on We revisited the effect of declining union mem- wage inequality from a fifth to a third. By bership on wage inequality, arguing that unions this measure, the decline of the U.S. labor not only equalize union members’ wages, they movement has added as much to men’s wage

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 Western and Rosenfeld 533 inequality as has the relative increase in pay unions likely slowing or cushioning the for college graduates. Among women, union impact on wages. decline and inequality are only related through Union decline, too, captures much of the the link between industry-region unionization growth in inequality, with just two explana- and nonunion wage dispersion. Union decline tory variables: union membership and industry- contributes just half as much as education to region unionization rate. Given the parsimony the overall rise in women’s wage inequality. and empirical power of union effects com- These results suggest unions are a normative pared to a multiplex and hard-to-measure presence that help sustain the labor market as process of economic change, we view the a social institution, in which norms of equity empirical analysis as supporting unions’ broad shape the allocation of wages outside the influence on U.S. labor market inequality. union sector. Having established the population-level Of course, not all unions of the 1970s were association, more research is needed to better in the vanguard of egalitarianism. In skilled understand the mechanisms connecting de- trades and construction, unions often rein- unionization to nonunion wages in key indus- forced racial and ethnic inequalities. Com- tries and labor markets. pared with the American Federation of More generally, the analysis contributes Labor’s craft unions, industrial unions origi- to a political account of rising economic nating in the Congress of Industrial Organiza- inequality in the United States. The analysis tions were more supportive of redistributive suggests that unions helped shape the alloca- public policies, including civil rights and the tion of wages not just for their members, but welfare state (Draper 1989; Quadagno 1994). across the labor market. The decline of U.S. Nor were all nonunion workplaces of the labor and the associated increase in wage 1970s dominated by norms of fairness and inequality signaled the deterioration of the limited markets. Nonunion employers were labor market as a political institution. Work- often virulently anti-union (Jacoby 1997), ers became less connected to each other in and wage inequality was clearly higher in the their organizational lives and less connected nonunion sector. in their economic fortunes. The de-politiciza- Could the union–inequality connection we tion of the U.S. labor market appears self- observed spuriously result from economic reinforcing: as organized labor’s political changes that helped eradicate organized labor, power dissipates, economic interests in the increase inequality, and perhaps transform labor market are dispersed and policymakers labor market norms? Computerization, firm have fewer incentives to strengthen unions or and industry deregulation, and economic glo- otherwise equalize economic rewards. balization may have fostered a more dynamic Although industry and regional variation and unequal U.S. capitalism to which unions play important roles, the key comparison were poorly adapted (Hirsch 2008). In this implied by our analysis is fundamentally his- case, deunionization is not directly implicated torical, from the early 1970s to the 2000s. In in the process of growing inequality; it is just the earlier period, unions offered an alterna- the visible byproduct of new kinds of employ- tive to an unbridled market logic, and this ment relations in new, nonunion industries. institutional alternative employed over a third At least some of the association between of all male private sector workers. The social industry-region unionization and nonunion experience of organized labor bled into non- wages is likely related to technological, union sectors, contributing to greater equality organizational, and institutional changes that overall. As unions declined, not only did the eliminated rents and fueled deunionization. logic of the market encroach on what had Still, changes in the economy are themselves been the union sector, but the logic of the unlikely to be purely exogenous; the process market deepened in the nonunion sector, too, of economic change spread unevenly, with contributing to the rise in wage inequality.

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Appendix The third adjusted variance fixes education effects in α and β at their 1973 values. The 1t 1t modified coefficient vectors yield an adjusted A. Calculating adjusted set of conditional means, which are plugged into variances the between- and within-group equations.

Between- and within-group variances in year t can be written in matrix form: B. Correcting for 2 Misclassification of B wy′ ^ y tt=−()tt (A.1) Union Status and, ′ 2 Card (1996) analyzed validation data from an Wtt= w σt (A.2) employer survey that indicated the misclassi- where w is a vector of survey weights, y^ is a fication rate of true union status in the 1977 t t – vector of conditional means of log wages, y t is May CPS is about 2.5 percent (λ = .025). If 10 2 the grand mean, and σt is the vector of resid- percent of respondents are union members, ual variances. The conditional means and 2.5 percent of the 10 percent misreport that variances are given by the variance function they are nonunion, and 2.5 percent of the 90 regressions, in matrix form, percent who are nonmembers misreport that they are union members. Formally, if an indi- yX^ =+α u αα+ u t tt12tt tt3 cator for true union status, u* = 1, then and, ** pu()ii==01||up==()uuii 10==λ. 2 log.σβtt=+Xu12ttttββ+ u 3t Given the observed unionization rate, ū, The first adjusted variance is based on the and the true unionization rate, ū*, reweighted data: uu()11−+λλ**()−u .

* wuti bt/uuif ti =1 wti = Rearranging terms yields an adjustment to the {wuti ()11−−bt/( u ),otherwise observed rate, providing the true rate of where b is a baseline year set here to b = 1973. unionization: The adjusted weights, w* , are then plugged t * u −λ into the between- and within-group equations, u = . 12− λ A.1 and A.2. The second adjusted variance adds to the Evidence for misclassification of union status union effect by fixing industry-region unioni- is based on just one year and measurement zation terms at their 1973 values: error in other years may be larger or smaller. To accommodate this uncertainty, we place a yX^* α uuαα t =+tt12tt+ bb3 prior distribution on λ. True union status, u , i* and, is also unobserved, and uncertainty about union status should be incorporated in the *2 log.σβtt=+Xu12ttu ββtb+ 3b final results. To describe this uncertainty, we can write a probability distribution for u , i* The adjusted means and variances, with weights given the observed union status. adjusted to 1973 unionization rates, are then Rearranging terms we can find the proba- plugged into the equations for between- and bility distribution for true union status, u* , 2 i **′ ^ ** within-group variance, wytt− yt and * uu−1 **′ 2 () pu(,|u==01λπ)(00−π ), wttσ .

Downloaded from asr.sagepub.com at Mina Rees Library/CUNY Graduate Center on May 15, 2015 Western and Rosenfeld 535 where, anonymous reviewers who commented on earlier versions of this article. Jessica Simes and Jennifer Laird provided λu* π = excellent research assistance. 0 1−u and

** Notes pu(,*|u 11)(uu),−1 ==λπ11−π 1. Thompson (1971) coined the term “moral economy.” Elster (1989a) argued that distributional norms are where, * ()1− λ u departures from strict rationality, eliciting strong π = . 1 u emotions in response to violation. 2. Union self-interest may drive political support, however, To incorporate uncertainty about the mis- where minimum wage increases reduce competition classification parameter, we write from nonunion workers. 3. Card (1996) studied a sample of 1,718 men with valid pu()**||up= (,uuλλ)(pd),λ wage data and nonmissing employer and employee ∫ union membership responses. 4. Analysis of earnings nonresponse reveals modest posi- where p(λ) is a prior distribution. In our analy- tive selection, but bias in regression coefficients is small sis, we specify this distribution to be uniform if imputed data are excluded (Bollinger and Hirsch on the interval, [.02, .03]. 2009). If unobserved high earners are nonunion, union If log wages, y , are normal, the variance wage effects and the union decline effect on between- i group inequality will be overestimated. But if high-pay function model, conditional on true union unobserved nonunion workers are drawn from low- status, u , can be written, i* union industries and regions, inequality and the union effect on within-group inequality will be underesti- ^ 2 yii∼Ny(,σi ), mated. In short, effects of nonresponse are likely modest where and small biases in union effects are offsetting.

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