AGE and SEX DISTRIBUTIONS of WINTERING SURF SCOTERS: IMPLICATIONS for the USE of AGE RATIOS AS an INDEX of RECRUITMENT Author(S): Samuel A

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AGE AND SEX DISTRIBUTIONS OF WINTERING SURF SCOTERS: IMPLICATIONS FOR THE USE OF AGE RATIOS AS AN INDEX OF RECRUITMENT Author(s): Samuel A. Iverson, Barry D. Smith, Fred Cooke Source: The Condor, 106(2):252-262. Published By: Cooper Ornithological Society DOI: http://dx.doi.org/10.1650/7386 URL: http://www.bioone.org/doi/full/10.1650/7386

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AGE AND SEX DISTRIBUTIONS OF WINTERING SURF SCOTERS: IMPLICATIONS FOR THE USE OF AGE RATIOS AS AN INDEX OF RECRUITMENT

SAMUEL A. IVERSON1,4,BARRY D. SMITH2 AND FRED COOKE3 1CWS/NSERC Centre for Wildlife Ecology, Simon Fraser University, 5421 Robertson Road, Delta, BC, V4K 3N2, Canada 2Paci®c Wildlife Research Centre, Canadian Wildlife Service, 5421 Robertson Road, Delta, BC V4K 3N2, Canada 3Larkins Cottage, 6 Lynn Road, Castle Rising, Norfolk PE31 6AB, UK

Abstract. We assessed age- and sex-speci®c distribution patterns of Surf Scoters (Me- lanitta perspicillata) wintering in southern coastal British Columbia, Canada, and evaluated potential biases associated with the use of male age ratios as an index of their recruitment. For surveys conducted during 2000 through 2002, annual variations in male age ratios were evident, with estimates ranging from 0.07 Ϯ 0.02 to 0.13 Ϯ 0.03 (SE; ®rst-year males:total males). Flock composition patterns indicated ®rst-year males did not distribute independently, but tended to associate with other ®rst-year males. With respect to habitat, male age- class proportions did not vary among shoreline substrate types, but higher proportions of ®rst-year males were found in sites with low exposure to wind and waves (Ͻ50 km fetch). To determine the ef®cacy of male age ratios for indexing recruitment, we used a power analysis, which incorporated overdispersion in age-class segregation and determined the sample sizes required for precise estimates of the proportion of ®rst-year male Surf Scoters. Samples of approximately 600±1000 total males were required to obtain 95% con®dence limits within 5% of the estimated mean, with sampling accuracy leveling off at about 2% when 6000 or more males were aged. Recruitment among waterfowl species is typically modeled using the ratio of female recruits to breeding-age females. Based on the sex and male age-ratio estimates obtained in this study, we calculated a female age ratio of 0.23 (®rst-year females:adult females). Key words: age-ratio estimation, ¯ock, Melanitta perspicillata, overdispersion, recruitment, seaduck, Surf Scoter.

Distribuciones Invernales de Edad y Sexo en Melanitta perspicillata: Implicancias del Uso de Cocientes de Edad como unÂ Indice de Reclutamiento

Resumen. Estimamos los patrones de distribucioÂn especõÂ®cos de edad y sexo de Mela- nitta perspicillata invernando en la zona costera sur de la Columbia BritaÂnica, CanadaÂ, y evaluamos los sesgos potenciales asociados con el uso de cocientes de edad de machos como unÂndice õ de su reclutamiento. Durante los muestreos realizados entre el 2000 y el 2002 las variaciones anuales en el cociente de edad de los machos fueron evidentes, con estimaciones que variaron entre 0.07 Ϯ 0.02 (EE) y 0.13 Ϯ 0.03 (machos del primer anÄo: total de machos). Los patrones de composicioÂn de las bandadas indicaron que los machos del primer anÄo no se distribuyeron independientemente, sino que tendieron a asociarse con otros machos del primer anÄo. Con relacioÂn al haÂbitat, las proporciones de clases de edad de los machos no variaron entre los tipos de substrato de la lõÂnea de costa, pero se encontraron mayores proporciones de machos del primer anÄo en sitios con baja exposicioÂn al viento y a las olas (sitios de mar abierto Ͻ50 km de ancho). Para determinar la e®cacia de los cocientes de edad de los machos comoÂndice õ de reclutamiento, usamos un anaÂlisis de poder, el cual incorporoÂ sobre-dispersioÂn en la segregacioÂn de las clases de edad y determinoÂ los tamanÄos de muestreo necesarios para estimaciones precisas de la proporcioÂn de machos del primer anÄo de M. perspicillata. Muestras totales de aproximadamente 600±1000 machos fueron necesarias para obtener lõÂmites de con®anza del 95% dentro del 5% de la media estimada, con la exactitud de muestreo nivelaÂndose cerca del 2% luego de estimarse la edad de 6000 o maÂs machos. El reclutamiento entre las especies de Anseriformes es modelado

Manuscript received 26 June 2003; accepted 14 January 2004. 4 E-mail: [email protected]

[252] SURF SCOTER WINTER AGE RATIOS 253

tõÂpicamente usando el cociente entre nuevas hembras y hembras en edad reproductiva. Ba- sados en las estimaciones de cocientes de sexo y edad de los machos obtenidas en este estudio, calculamos un cociente de edad de las hembras de 0.23 (hembras del primer anÄo: hembras adultas).

INTRODUCTION siderations. First, age ratios are subject to mis- Few reliable population indices or estimates of interpretation if factors such as geographic lo- annual productivity are available for seaducks cation, differential use of habitats, or variable (Anatidae: Mergini) in North America. Analyses group composition prevent representative sam- of existing survey data indicate declining pop- pling of age classes. Further, because waterfowl ulation trends among 10 of the continent's 15 populations typically have male-biased sex ra- seaduck species (USFWS 1993, Goudie et al. tios (Bellrose et al. 1961), estimating the number 1994). These declines are poorly understood, of potentially breeding females is of primary im- and there is a need for new techniques to mon- portance in models (Johnson et al. 1992). For itor populations. Typically, monitoring efforts seaducks, female age classes cannot be distin- focus on estimating total abundance. However, guished on the basis of plumage alone and must incorporating demographic data into monitoring be calculated from male age ratios. Calculation programs can lend important insight into mech- of female age ratios is straightforward, so long anisms underlying population change and help as secondary sex ratios are equal (Blums and identify life stages at which changes are occur- Mednis 1996) and prebreeding mortality rates ring. are the same for subadult males and females; Recruitment, the process by which young however, any sex-related distributional biases birds are added to the breeding population, is a must be known. crucial component of avian demographics. Thus, The purpose of this study was to determine monitoring variation in recruitment rates is im- the age and sex composition of Surf Scoter (Me- portant from a management standpoint. Tech- lanitta perspicillata) populations wintering in niques traditionally used to assess waterfowl re- southern coastal British Columbia, investigate cruitment include pair:brood ratios (Kirby 1980, age- and sex-speci®c distribution patterns, and Rumble and Flake 1982), nesting success rates infer the reliability of male age ratio as an index (Milne and Reed 1974, Klett et al. 1988) and of recruitment. Until recently, the Surf Scoter age ratios based on wings turned in by hunters has been among the least studied North Ameri- (Bellrose et al. 1961, Munro and Kimball 1982). can waterfowl (Savard et al. 1998). Current data Unfortunately, these techniques have been of indicate population numbers are declining in limited value for many seaduck species, as most western North America (Goudie et al. 1994, breed in low densities in remote portions of the Hodges et al. 1996; Nysewander et al., unpubl. continent (Bellrose 1980) and rarely appear in data), and aerial surveys in Alaska suggest long- hunter bags (Bartonek 1994). term (1957±2001) breeding population declines An alternative method for indexing recruit- (Conant and Groves, unpubl. data). Surf Scoter ment by seaducks is the use of age ratios cal- breeding populations have been dif®cult to study culated from direct counts on nonbreeding areas. because nesting occurs in low densities across This technique relies upon delayed plumage remote boreal forest regions (Bellrose 1980). maturation to distinguish cohorts and estimate The demographic data that are available indicate population age structure. Age-related plumage variable, but generally low, reproductive rates variation has been described among males of all and high adult survival (Savard and Lamothe 15 North American seaduck species (Palmer 1991, Morrier et al. 1997). However, these data 1976), and winter age-ratio estimation has re- are from a few localized studies and systematic cently been used to assess the age structure of techniques for assessing breeding populations Common Goldeneye (Bucephala clangula; Dun- have not been implemented. can and Marquiss 1993) and Harlequin Duck For most of the year, Surf Scoters frequent (Histrionicus histrionicus; Smith et al. 2001, shallow marine habitats on wintering areas, Rodway et al; 2003, Robertson, in press) popu- where scoter ¯ocks are a conspicuous part of the lations. However, use of male age ratios in pop- avifauna in many coastal areas. On the Paci®c ulation models requires several important con- coast a vast majority of individuals occur within 254 SAMUEL A. IVERSON ET AL.

or continuous section of coastline). An additional seven survey areas, each composed of 2±4 sample stations, for a grand total of 57 sample stations, were added in 2001 and 2002. All counts were made from shore during daylight hours (08:00±16:00), using a 20±60ϫ spotting scope, when Beaufort scale sea conditions were rated as 3 (small scattered whitecaps, gentle breeze, wind speed 12±19 knots) or less. Each replicate required 3±5 days to complete. At each sample station, the number of Surf Scoters within 400 m of shore was counted. Af- ter estimating abundance, a scan was performed FIGURE 1. Map of the study area in the Strait of Georgia, British Columbia, Canada. Circles indicate to determine age and sex composition. This scan areas used to survey Surf Scoters during 2000, 2001, proceeded slowly over each ¯ock while a min- and 2002; squares indicate survey areas used during imum of birds were in motion or diving. For 2001 and 2002; triangles indicate survey areas used Surf Scoter ¯ocks with more than 50 individu- during 2000. als, age and sex determinations often could not be made for every individual due to birds being underwater or obstructed from view by other 1 km of shore (Vermeer 1981), making winter birds. For these ¯ocks, the age and sex deter- populations readily accessible to researchers. mination scan was considered a random sample Iverson et al. (2003) determined ®rst-year males of ¯ock composition, and the proportions ob- in these ¯ocks can be reliably distinguished tained were treated as representative of the ¯ock from adult males and females on the basis of as a whole. Sample station estimates were av- plumage, raising the possibility that winter age eraged to produce a mean for each survey area. ratios could be used as an index of recruitment. Individual Surf Scoters were categorized by For this paper, our four primary research objec- sex (M ϭ male; F ϭ female), and males were tives were to (1) estimate Surf Scoter winter age categorized by age (ADU ϭ adult, Ն12 months and sex ratios in the Strait of Georgia, British old; 1Y ϭ ®rst-year, Ͻ12 months old). These de- Columbia, Canada, and document annual varia- terminations were made on the basis of plumage tion in population composition; (2) investigate and bill appearance (Iverson et al. 2003). During the potential effects that age and sex-related dif- winter, ®rst-year males (M1Y) are characterized ferences in ¯ock composition, habitat use, and by dull multicolored bills, light belly plumage, latitudinal distribution patterns might have on mottled black and brown feathering on the head age-ratio estimates; (3) determine the sample and back, poorly de®ned nape patches, and the sizes required to detect changes in age ratios absence of forecrown patches. Adult males among years; and (4) discuss the advantages and (MADU) are distinguished by their bright, bul- disadvantages of using winter age ratio as an bous, multicolored bills, lustrous black plumage, index of recruitment by Surf Scoters and per- and large, clearly de®ned nape and forecrown haps other migratory bird species. patches. Both ®rst-year and older females exhib- METHODS it dark bills, brown plumage, and variable amounts of lighter feathering on the breast and STUDY DESIGN belly. Female age is not distinguishable on the The age and sex composition of Surf Scoter pop- basis of appearance alone; therefore, we grouped ulations was investigated in the Strait of Geor- them into a single category (FTOTAL; Iverson et gia, British Columbia (Fig. 1), during the win- al. 2003). ters of 2000, 2001, and 2002. Three replicate surveys were conducted per year, at 10±14 day STATISTICAL ANALYSES intervals, commencing the third week of January Age and sex ratio estimation. The proportion each year. During 2000, counts were made at 36 ®rst-year males (M1Y/MTOTAL), and proportion of sample stations, encompassing seven survey ar- females (FTOTAL/[MTOTAL ϩ FTOTAL]) were estimat- eas (survey area de®ned as a bay, inlet, sound, ed for each year (Ϯ SE) using mean values for SURF SCOTER WINTER AGE RATIOS 255 each survey area weighted according to sample upon the analyses being performed) for ¯ock i size. Recruitment among waterfowl is common- of n with Fi birds; and a, b, ␭1, and ␭2 are pa- ly indexed as the ratio of ®rst-year females to rameters to be estimated. Variables Y2001 and adult females (Cowardin and Blohm 1992). To Y2002 are categorical (0 or 1), indicating the year. facilitate comparison with literature values, sex When all parameters except a equal zero,pFi is ratio (MTOTAL/FTOTAL) and male age ratio (M1Y/ constant for all ¯ock sizes. We chose the statis-

MADU) were also calculated. Female age ratio tical error of this model (␧i) to follow a beta- (M1Y/[FTOTAL Ϫ M1Y]) was derived, based upon binomial distribution, the assumption of a 1:1 ratio among ®rst-year p[s ] ϭ BB[s ; F , p , ␪2 ] males and females. Annual variation in these pa- FFiFFiiii rameters was assessed using ANOVA, with ␣ϭ where p[SFi ] is the probability of having ob- 0.05. Speci®cally, we used a general linearized served SF ®rst-year males (or females) in a ¯ock model ANOVA appropriate for our binomial 2 of size Fi, and where 0 Յ ␪ Fi Յ 1 is the over- data, with the Wald-statistic, W (StatSoft 1995). dispersion parameter of the beta-binomial distri- Flock composition patterns. A typical as- bution. To model the degree of overdispersion sumption of a random sampling protocol to es- we used timate age and sex ratios would be that the clas- 1 ses assort themselves randomly within ¯ocks in 2 ␪ϭF the sampled population. To determine if this was i {1 ϩ eϪ(cϩd ln[Fi])ϩ␥1Y 2001ϩ␥ 2Y 2002} the case with Surf Scoters, we challenged a null which allows ␪2 to vary with ¯ock size, where hypothesis that aged and sexed individuals were Fi c, d, , and are parameters to be estimated independently distributed (i.e., no overdisper- ␥1 ␥2 sion) and that there was no trend in age and sex and variables Y2001 and Y2002 are as above. When 2 composition with respect to ¯ock size. These hy- all parameters except c equal zero, ␪ Fi is con- potheses were tested by regressing the propor- stant for all ¯ock sizes. tion ®rst-year male (®rst-year males:total males), Interpretation of the overdispersion parameter, ␪2 , is fundamental to understanding the degree and the proportion female (total females:total Fi birds), against ¯ock size, in separate analyses. to which cohorts group together into ¯ocks. At one extreme, when ␪2 ϭ 0, all individuals are We wished to know (1) whether members of ei- Fi ther age class (®rst-year males versus adult randomly distributed; that is, the beta-binomial males) or sex class (females versus males) tend- distribution limits to the binomial distribution. 2 ed to occur more frequently in ¯ocks of a par- At the other extreme, when ␪ Fi ϭ 1 individuals ticular size, and (2) whether members of either associate only with their kind. For example, with age or sex class tended to occur in ¯ocks com- respect to the segregation of ®rst-year and adult 2 posed predominately of like individuals. males, ␪ Fi ϭ 1 would indicate that each ¯ock To effect our analyses we treated each ¯ock consisted entirely of either ®rst-year males or as an independent sample unit, yielding a sample adult males. To elucidate the magnitude of over- size of n ϭ 408 for all years and replicates com- dispersion, we calculated the effective indepen- bined. These regressions used AICc and likeli- dent unit, mathematically de®ned as the ratio of hood ratios to rank and test competing paramet- the beta-binomial variance to the equivalent bi- ric hypotheses (Burnham and Anderson 2002) nomial variance with no overdispersion, i.e., and were designed to respect the binomial error 1 ϩ␪2 (F Ϫ 1). structure of the data by using a beta-binomial Fii error distribution (Mood et al. 1985, McCullagh Hence, an effective independent unit equal to and Nelder 1989). The linear equations for the one would indicate individuals were distributing predicted proportions at each ¯ock size were lo- themselves at random among ¯ocks. A ratio git-transformed to retain the predictions within greater than one would indicate individuals were the range 0±1. Thus our statistical model was not acting independently; rather, they could be 1 interpreted as associating in numbers equal in pFiϭϩ␧size to the estimated effective independent unit. i {1 ϩ eϪ(aϩb ln[Fi])ϩ␭1Y 2001ϩ␭ 2Y 2002} One key consideration was the possible in¯u- where 0 ՅՅpFi 1 is the predicted proportion ence of random measurement error on the de-

®rst-year male (or proportion female, depending gree of overdispersion inpFi , since erroneous 256 SAMUEL A. IVERSON ET AL. counts of ®rst-year or adult males would con- Simple linear regression was used to evaluate tribute to observed overdispersion. Since we had latitudinal trends in age and sex proportions for no independent assessment of measurement er- wintering aggregations ranging from 35±58ЊN. 2 ror, we used our estimates ofpFFii and ␪ to cal- Power analysis. To determine our power to culate the standard deviation of extrabinomial estimate the proportion of ®rst-year male Surf variance associated with each estimate of pFi. Scoters with a speci®ed con®dence and make For these calculations we used recommendations for future study design, we used a bootstrap resampling scheme to generate SD[p ] ϭ ͙␪2 p (1 Ϫ p ), FiFFiii con®dence intervals for various sampling inten- derived from the ratio of variance of the beta- sities. Our motivation was to determine the sam- binomial to the binomial distribution. We then ple size required to con®dently estimate this pro- made an a posteriori judgment if measurement portion, given the overdispersion observed in error was likely to be a signi®cant contributor to our data. Such overdispersion increases the size the value of SD[].pFi of the effective independent unit, thus propor- Habitat segregation. To determine whether tionately increasing the sample size required to ®rst-year males or females segregated into dif- achieve a desired con®dence interval. In our ferent habitats, midwinter survey counts were power analysis, ¯ock composition data from all related to habitat attribute data. We used the three years and replicate samples were treated British Columbia Marine Ecosystem Classi®ca- both individually and pooled into a single data- tion (BCMEC) for the Paci®c Coast of Canada set composed of 408 ¯ock observations, con- (Zacharias et al. 1998) to characterize the sub- taining 6422 total males. strate and exposure for each sample station. For Our ®rst step was to perform a jackknife pro- the BCMEC ecounit themes, substrate categori- cedure (Efron and Tibshirani 1998) to estimate zations were taken from Geological Survey of the mean proportion of ®rst-year males pÅ F, and 2 Canada sediment distribution maps, with sites ␪¯ Fpi, as well as their standard errors (sÅF and classi®ed as rocky: bedrock, boulders, or cobble, S 2 , respectively) using the model equations de- ␪¯ F with gravel/sand interspersed; sandy: sand or scribedi above, but with all parameters except a sand/gravel; or muddy: mud or sand/mud. Wind and c equal to zero. Our choice to use a jack- and wave exposure were categorized as low: knife procedure to calculate these moments was protected area, fetch Ͻ50 km; or moderate: open motivated by the variable ¯ock sizes and over- sound or strait, fetch 50±500 km. Variations in dispersion attributes related to ¯ock size and age and sex composition were investigated using habitat in our data, features not captured when the same ANOVA procedures described above these moments are estimated using sample sta- for binomial data, where mean values were cal- tistics. Given estimates for these population mo- culated for each survey area weighted according ments, which we treated as known values, we to sample size. could randomly sample ®rst-year and adult

Latitudinal variation. Additional surveys males from ¯ock Fi, according to a beta-bino- were conducted during 2001 at locations along mial probability distribution. This bootstrapping the Paci®c coast to assess latitudinal variation in protocol was carried out for different sampling age and sex proportions. These surveys coincid- intensities, with the number of ¯ock observa- ed with survey dates used in the Strait of Geor- tions in each subsample increasing by sequential gia and followed identical sampling protocols. powers of 2 up to the maximum sample size of They were conducted by volunteers with previ- 408. For each subsample we used the model Ã 2 ous survey experience, whose assistance was so- equations to estimate pÅ F and␪ˆ Fi , and repeated licited through birding organizations throughout the procedure 100 times to generate stable esti- Ã the United States and Canada. Volunteers were mates ofsÃ Å andS 2 . These estimates were then pFF␪ i trained using a website and written materials, used to infer the minimum number of sample which provided detailed instructions on identi- stations and total males that would be required ®cation of age classes and survey guidelines. Of to estimate the proportion ®rst-year male within 40 original respondents, 27 completed the three 2% and 5% of the overall mean, using 95% con- replicate surveys. Seven surveys were dropped ®dence intervals for the upper and lower due to insuf®cient sample size (Ͻ40 total birds bounds. Note that for this procedure we pre- per survey) leaving an overall sample size of 20. sumed no knowledge of any relationship be- SURF SCOTER WINTER AGE RATIOS 257

TABLE 1. Number of Surf Scoters observed in the Strait of Georgia, British Columbia, Canada, 2000±2002, by sex and age class. Values are means Ϯ SE for three replicate midwinter surveys.

2000 2001 2002 Total Total Surf Scoters 1221 Ϯ 218 1316 Ϯ 238 1080 Ϯ 36 1206 Ϯ 56 Total females 442 Ϯ 66 422 Ϯ 77 379 Ϯ 23 414 Ϯ 15 Total males 779 Ϯ 152 894 Ϯ 162 701 Ϯ 12 791 Ϯ 46 Adult males 680 Ϯ 132 833 Ϯ 150 629 Ϯ 26 714 Ϯ 50 First-year males 100 Ϯ 21 62 Ϯ 16 72 Ϯ 14 78 Ϯ 9 Proportion ®rst-year malesa 0.13 0.07 0.11 0.10 Sex ratiob 1.76 2.12 1.87 1.91 Female age ratioc 0.29 0.17 0.23 0.23 a The number of ®rst-year males/the total number of males. b The number of males/the number of females. c The number of ®rst-year males/(the number of females Ϫ the number of ®rst-year males). tween these moments and ¯ock size (i.e., only statistical signi®cance using likelihood-ratio parameters a and c were estimated), contrary to tests, our best model could not support a slope what we will report in this manuscript. The rea- different from zero for the proportion ®rst-year son for this is that (1) these relationships would males,pÃÅ F , (b ϭ 0, P ϭ 0.78), suggesting no not necessarily exist in other data collections, strong age-related differences in ¯ock composi- and (2) if they exist but are not identi®ed, our tion with respect to ¯ock size. However, the cor- 2 power analysis will provide a conservative es- responding estimates of␪ˆ Fi were signi®cantly timate of the sample size required to overcome greater than zero (P Ͻ 0.001), and changed with the variance added by any such relationship. ¯ock size (d ϭϪ0.46, P Ͻ 0.001), indicating Means are presented Ϯ SE. that ®rst-year males did not independently sort into ¯ocks (Fig. 2). As in our age-ratio analysis RESULTS above, we also identi®ed interannual differences Age and sex ratios. Similar numbers of Surf in the estimated mean proportion of ®rst-year Scoters were counted each year, with survey males,pÃÅ F , in this analysis. Our best model in- means ranging from 1080 Ϯ 36 birds in 2002 to cluding year effects was 378 times more likely 1316 Ϯ 238 in 2001. Estimates for the propor- than its direct competitor lacking them, and tion ®rst-year males differed among years (W2 thousands of times more likely than the null Ã 2 ϭ 41.6, P Ͻ 0.001), with annual estimates of model (constantpÅ FF and␪ˆ i ). The effective in- 0.13 Ϯ 0.03 in 2000, 0.07 Ϯ 0.02 in 2001, and dependent unit for males (averaged over all 0.11 Ϯ 0.02 in 2002. For males, these values years 2000±2002) increased from about 3.4 in yielded an overall age composition estimate of ¯ocks of 10 males, to 8.4 in ¯ocks of 50, and 0.10 ®rst-year males:total males. With respect to 12.6 in ¯ocks of 100, indicating a tendency for sex, the population composition was strongly ®rst-year males to increasingly associate with male biased, with an overall estimate of the pro- other ®rst-year males as ¯ock size increased. portion female of 0.35 total females:total birds. The standard deviation of extrabinomial vari-

Annual variations were signi®cant (W2 ϭ 11.3, ance SD[pFi ] was estimated to range from 0.1 to P ϭ 0.04), with estimates of 0.36 Ϯ 0.02 in 0.2 for ¯ock sizes of 1 to 100 birds. 2000, 0.32 Ϯ 0.02 in 2001, and 0.35 Ϯ 0.02 in With respect to sex, our best model indicated

2002. This male-biased estimate for population that the proportion female,pFi , tended to be composition contributed to calculated female higher in larger ¯ocks (b ϭ 0.08, P Ͻ 0.001, 2 age ratios that were more than twice as high as Fig. 2). The estimate of ␪ Fi ϭ 0.02 was signif- male age ratios (female age ratio ϭ 0.23 ®rst- icantly different from zero (P Ͻ 0.001), but 2 year females:adult females, range 0.17±0.29; there was no trend in ␪ Fi with ¯ock size (d ϭ male age ratio 0.11 ®rst-year males:adult males, 0, P ϭ 0.74). As in our sex ratio analysis above, range 0.07±0.15; Table 1). we also identi®ed interannual differences in the

Flock composition patterns. On the basis of mean proportion of females, pÃÅ F, in this analysis. model selection using AICc, and judgment of Our best model including year effects was 41 258 SAMUEL A. IVERSON ET AL.

FIGURE 2. Relationship between Surf Scoter ¯ock FIGURE 3. Geographic variation in age and sex size and (a) proportion ®rst-year male or (b) proportion composition of Surf Scoter ¯ocks on the Paci®c coast female. Points indicate observed proportions, solid line of North America. Data were collected during January indicates predicted proportions averaged over all years, and February 2001 by volunteers at 20 coastal loca- and dashed line indicates the effective independent tions between Alaska and California. unit, a measure of the degree to which individuals in a given age or sex cohort tend to associate, and thus behave as a unit for statistical purposes. Interannual total birds) in the three habitats were estimated differences in the trends portrayed were statistically detected, but to retain clarity only the mean values are as 0.38 Ϯ 0.01 in rocky habitats, 0.34 Ϯ 0.01 in presented. muddy habitats, and 0.29 Ϯ 0.01 in sandy habitats. First-year male proportions were higher in sites with low exposure to wind and waves, 0.11 times more likely than its direct competitor lack- Ϯ 0.01, than in sites with moderate exposure, ing them, and thousands of times more likely 0.08 Ϯ 0.01 (W1 ϭ 16.3, P Ͻ 0.001). With re- Ã 2 than the null model (constantpÅ FF and␪ˆ i ). The spect to sex, there was no difference in propor- effective independent units differed only slightly tion female between low and moderate exposure from unity (complete independence) in the sites (W1 ϭ 1.2, P ϭ 0.27). smallest ¯ocks (averaged over all years 2000± Latitudinal variation. Results of the volunteer 2002), increasing to about 1.2 in ¯ocks of 10 surveys suggested that latitudinal variation in birds, then approximately 3 in ¯ocks of 100. The the age composition of Surf Scoter populations standard deviation of extrabinomial variance may exist (Fig. 3). The proportion of ®rst-year

SD[pFi ] was estimated to range from 0.05 to 0.06 males was negatively related to latitude (M1Y/ 2 for ¯ock sizes of 1 to 100 birds. MTOTAL ϭ 0.38 Ϫ 0.01 latitude, F1,18 ϭ 13.8, r Habitat segregation. When relating the com- ϭ 0.43, P Ͻ 0.01). Sex ratios did not exhibit a position of ¯ocks to substrate type, age-related latitudinal trend (FTOTAL/(MTOTAL ϩ FTOTAL) ϭ 0.41 2 segregation was not evident among males (W2 ϭ latitude, F1,18 ϭ 0.4, r ϭ 0.02, P ϭ 0.56). These 1.8, P ϭ 0.41). However, sex-related habitat seg- results, however, should be treated with caution, regation was evident, with females being under- as the survey was conducted only for a single represented in sandy substrate and overrepre- year and the number of samples at both the sented in rocky and muddy substrate (W2 ϭ 61.2, northern (Ͼ55ЊN) and southern (Ͻ45ЊN) ends of P Ͻ 0.001). Female proportions (total females: the range was small. SURF SCOTER WINTER AGE RATIOS 259

year females:adult females. These relatively low age-ratio estimates are consistent with expecta- tions for a seaduck species, whose life history strategies are characterized by high annual adult survival and variable, but generally low, annual productivity (Goudie et al. 1994). Flock composition and habitat use patterns suggest Surf Scoter winter distributions are not entirely independent with respect to age and sex class. There was a tendency for ®rst-year males to cluster into ¯ocks with other ®rst-year males, particularly within larger ¯ocks. The grouping of male age classes is not surprising, given the strong ¯ocking behavior of Surf Scoters in general (Savard et al. 1998) and the postulated role FIGURE 4. Number of sample stations and total of dominance interactions within waterfowl number of male Surf Scoters necessary to estimate the ¯ocks (Nichols and Haramis 1980, Hepp and proportion ®rst-year male Surf Scoters (pFi) within Hair 1984, Alexander 1987). Relatively high speci®ed con®dence limits. values for the estimated extrabinomial variance,

SD[pFi ], which ranged from about 0.1 to 0.2, Power analysis. Our power analysis, designed suggest that our measure of overdispersion for to calculate con®dence levels for various sam- ®rst-year males is not due entirely to measure- pling intensities, allowed us to assess the ef®- ment error. That is, it seems unlikely that our cacy of winter age-ratio estimation. In our da- standard error of measurement would be equal taset that included samples from all three years to or double the estimated proportion ®rst-year combined, a jackknifed mean proportion ®rst- male (pÃÅ F ϭ 0.10). year male,pÅ F, of approximately 0.10 was esti- In contrast to ®rst-year males, there was little 2 mated, with spÅF ϭ 0.01. The value for ␪ Fi was evidence that females were joining ¯ocks com- estimated as 0.22, with S 2 ϭ 0.04. Bootstrap posed primarily of other females. Our estimates ␪ Fi simulation results suggest that in order to narrow of the effective independent unit (Fig. 2) for fe-

95% con®dence intervals to within 5% of pÅ F, males differed only slightly from unity in small- 40±60 sample stations or 630±960 male Surf er ¯ocks, and the magnitude of overdispersion Scoters would need to be surveyed. For an es- was not suf®cient to distinguish it from potential timate within 2%, sampling intensity would have measurement error in larger ¯ocks. Although to increase to 400±600 sample stations or 6280± there was no trend in the overdispersion param- 9420 males (Fig. 4). eter, a positive relationship between proportion female and ¯ock size was evident. We speculate DISCUSSION that this relationship is related principally to At a regional scale, such as the Strait of Georgia, habitat preferences, which appear to differ be- Surf Scoter winter population composition can tween the sexes. Female proportions were high- be estimated with con®dence. However, in order est over rocky substrates, where Surf Scoters to avoid potential biases resulting from age- and forage primarily on mussels (Vermeer 1981). sex-related differences in ¯ock composition and Flocks in these rocky habitats tended to be larger habitat use patterns, large samples drawn rep- and more densely packed than in sandy areas, resentatively from available habitats must be ob- where clams are the principal prey item and tained. In this study, we estimated an overall ¯ocks were typically smaller and more dispersed proportion ®rst-year male of 0.10 ®rst-year (Iverson 2002). First-year males did not segre- males:total males, with annual estimates ranging gate from adult males according to substrate between 0.07 and 0.13. The population was also type, but they were underrepresented relative to strongly male biased, with a sex ratio of 1.9 adults in more exposed sites. The most obvious males per female. From these estimates we cal- explanation for the small number of ®rst-year culated a male age ratio of 0.11 ®rst-year males: males in exposed sites would be body size dif- adult males and a female age ratio of 0.23 ®rst- ferences; however females, which are smaller 260 SAMUEL A. IVERSON ET AL. and 10±20% lighter than males (Palmer 1976), However, sample sizes in this part of the study did not exhibit the same pattern. It should be were small, and there is a danger of committing noted, however, that we sampled only under rel- a Type I error if too much credence is given to atively calm weather conditions, and it is pos- an extrapolation from a single year's observa- sible that exposure-related distributional differ- tion. Future research will be necessary to eval- ences are more pronounced under extreme wind uate the trend, determine if it is authentic, and and wave conditions. if so investigate whether distributional differenc- Our power analyses, which incorporated the es are related primarily to site preferences by tendency for individuals of various age and sex young or competitive exclusion by adults. classes to associate with like individuals, and The potential for differential distribution of thus behave as a group for statistical purposes, the age classes at a latitudinal scale during win- were designed to assess the level of precision ter also raises the issue of how migratory pop- attainable at various sampling intensities. These ulations are linked through the annual cycle, as analyses indicated that approximately 600±1000 migratory movements often confound efforts to total males were needed to obtain estimates delineate distinct population units for manage- within 5% of the population mean, and that pre- ment purposes (Esler 2000, Webster et al. 2002). cision could be further improved with larger Population models most often focus on breeding samples, reaching roughly 2% from a sample of aggregations, in part due to the direct implica- 6000 or more total males. While sampling inten- tions for reproductive dynamics (Lindberg et al. sities of such a magnitude would be logistically 1998). Use of demographic data derived from feasible in many areas, and could act as a guide wintering areas, however, requires that addition- for other researchers when designing monitoring al consideration be given to seasonal movements programs, we caution that demographic param- and potential linkages between breeding and eters may differ among populations. Our analy- nonbreeding areas. ses indicated that the sample size required to Seaduck species exhibit a wide range of as- achieve a chosen level of con®dence depends sociation patterns between breeding and non- upon the mean age proportion, mean ¯ock size, breeding areas. Common Eider (Somateria mo- and overdispersion characteristics, all of which llissima) populations, for example, are among may vary from location to location. the most geographically structured, with birds While the potential estimation complications breeding in speci®c areas in arctic and western resulting from age-related differences in ¯ock Alaska known to have separate wintering ranges composition, overdispersion, and habitat asso- (Petersen and Flint 2002). In such cases, age- ciation patterns can be remedied by sampling ratio estimates made on wintering areas should design, the data from volunteer surveys con- correspond directly with productivity estimates ducted at sites along the Paci®c coast may pre- from speci®c breeding locations. However, when sent a greater challenge for the application of migratory movements produce wintering aggre- winter age ratios in population monitoring gations which include multiple breeding popu- schemes and demographic models. This study lations, interpretation is more complex. For ex- was the ®rst to investigate the possibility of dif- ample, Paci®c populations of Steller's Eiders ferential migration according to age class among (Polysticta stelleri) assort into three distinct wintering seaducks. Latitudinal variation in Surf breeding aggregations, but combine to form a Scoter age ratios was evident, with a higher pro- single wintering population (Dau et al. 2000). portion of ®rst-year male Surf Scoters wintering Winter age ratios in such a population would farther south. Differential migration is common represent a pooled estimate of productivity from among migratory birds, with prey and habitat multiple breeding areas, and would have to be differences, competitive exclusion by adults, and treated as such in population models. Finally, for lower cold tolerance among young all potential- species such as Long-tailed Ducks (Clangula ly underlying geographic distribution patterns hyemalis), where little to no population structure during winter (Cristol et al. 1999). If latitudinal has been identi®ed (M. Petersen and P. Flint, un- variation in Surf Scoter winter age ratios exists, publ. data), winter age ratios can serve only as then estimates taken at a regional scale would a rough measure of age composition, with ®ne- be greatly complicated by interannual ¯uctua- scale linkages between speci®c breeding and tions in productivity in the larger population. wintering areas not being possible. To date, mi- SURF SCOTER WINTER AGE RATIOS 261 gratory movements and potential linkages be- LITERATURE CITED tween breeding and nonbreeding areas have not ALEXANDER, W. C. 1987. Aggressive behavior of win- been elucidated for Paci®c Surf Scoter popula- tering diving ducks (Aythyini). Wilson Bulletin tions. Therefore, caution must be applied when 99:38±47. BARTONEK, J. C. 1994. Waterfowl harvests and status, attempting to relate winter age-ratio estimates to hunter participation and success, and certain hunt- productivity and recruitment estimates at specif- ing regulations in the Paci®c Flyway and United ic breeding locations, and when extrapolating re- States. USDI Fish and Wildlife Service, Paci®c gional estimates to larger geographic scales dur- Flyway Data Book, Portland, OR. BELLROSE, F. C. 1980. Ducks, geese and swans of ing winter. North America. Stackpole Books, Harrisburg, PA. At present there is a clear need for new tech- BELLROSE, F. C., T. G. SCOTT,A.S.HAWKINS, AND J. niques to assess and monitor seaduck popula- B. LOW. 1961. Sex ratios and age ratios in North tions. Demographic data can be particularly use- American Ducks. Illinois Natural History Survey Bulletin 27:391±474. ful for understanding the mechanisms underly- BLUMS,P.,AND A. MEDNIS. 1996. Secondary sex ratio ing population changes. For species such as the in Anatinae. Auk 113:505±511. Surf Scoter, which breed in low densities in re- BURNHAM,K.P.,AND D. R. ANDERSON. 2002. Model mote portions of the continent, estimation of re- selection and multimodel inference: a practical in- formation-theoretic approach. 2nd edition. Spring- cruitment from winter age ratios offers an at- er, New York. tractive alternative to traditional methods. Fur- COWARDIN, L. M., AND R. J. BLOHM. 1992. Breeding ther, because distribution patterns can reveal im- population inventories and measures of recruitment, p. 432±445. In B. D. Batt, A. D. Afton, M. portant relationships to food resources, as well G. Anderson, C. D. Ankney, D. H. Johnson, J. A. as adaptive social organization important to sur- Kadlec, and G. L. Krapu [EDS.], Ecology and man- vival, understanding age- and sex-related differ- agement of breeding waterfowl. 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