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Immigrant employment and earnings growth in Canada and the USA: evidence from longitudinal data

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Abstract

We study the short-term trajectories of employment, hours worked, and real wages of immigrants in Canada and the USA using nationally representative longitudinal datasets covering 1996–2008. Models with person fixed effects show that, on average, immigrant men in Canada do not experience any relative growth in these three outcomes compared to men born in Canada. Immigrant men in the USA, on the other hand, experience positive annual growth in all three domains relative to US-born men. This difference is largely on account of low-educated immigrant men, who experience faster or longer periods of relative growth in employment and wages in the USA than in Canada. We further compare longitudinal and cross-sectional trajectories and find that the latter over-estimate wage growth of earlier arrivals, presumably reflecting selective return migration.

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Notes

  1. Figures are from: International Migration Statistics, Migration Policy Institute. http://www.migrationpolicy.org/programs/data-hub/international-migration-statistics. In proportion to overall population, the foreign born constitute a larger share of the Canadian population (21 % as of 2011) compared to the USA (13 % as of 2011) (Statistics Canada 2011; Census Bureau 2015a).

  2. In 1992, the Canadian government allocated 12 points (out of 100) to high education and 15 to French or English proficiency. By 2006, prospective immigrants with a bachelor’s degree received 20 points (out of 100), and those with a masters or Ph.D. received 25 points. The points allocated to language proficiency were raised to 24. Despite the focus on the point system, in 2006, only around 20 % of Canadian immigrants were adjudicated under the point system as principal applicants and another 25 % as their family members (CIC 2013). Research shows that family members’ credentials in the skilled worker category are correlated with those of the principal applicant (Sweetman and Warman 2010).

  3. All temporary workers on non-immigrant visas can, and often do, subsequently adjust their status to permanent residents. A growing proportion of new permanent residents in the USA in recent years have been temporary migrants already residing in the country. In 1986, 37 % of the foreign born receiving permanent residency were temporary residents; by 2009, the proportion had increased to 59 % (USDHS, 2013).

  4. Net inflows of undocumented migrants have been negligible since 2007 (Passel and Cohn 2012b).

  5. Researchers attribute the decline in entry earnings of successive immigrant cohorts in Canada to compositional shifts in language ability and region of birth, deterioration in returns to foreign labor market experience, and non-random sorting of immigrants across establishments in Canada’s major cities and geographic regions (Aydemir and Skuterud 2005, 2008; Green and Worswick 2009).

  6. See Chiswick (1978) and Borjas (1985, 1994) for cross-sectional research on US immigrants, and see Baker and Benjamin (1994), Bloom et al. (1995), Frenette and Morissette (2005), Warman (2007), and Warman and Worswick (2004) for research on Canadian immigrants. For longitudinal studies of immigrant earnings assimilation, see Borjas (1989), Duleep and Dowhan (2002), Hall and Farkas (2008), Hu (2000), Lubotsky (2007), and Kaushal (2011) for the USA and Banerjee (2009), Beenstock (2006), Li (2003), and Picot and Piraino (2013) for Canada.

  7. Observational studies in Canada indicate that immigrants face a considerable degree of occupational mismatch, implying presence of obstacles that may limit immigrants in achieving their full potential (Reitz 2001).

  8. About 65 % of permanent immigration to the USA and between 20 and 26 % of the permanent migration to Canada was family migration during our study period 1996–2008 (Government of Canada 2014; US Department of Homeland Security 2014). However, in recent years, a large proportion of foreigners who acquire permanent residency in the USA via family unification enter the country on temporary visas (Massey and Malone 2002; Hayes and Hill 2008).

  9. Unfortunately, it is not possible to directly test this because our data do not include information on class of entry. However, fixed effects models do help account for differences in earnings across entry classes within a country.

  10. Regional differences in unemployment insurance in Canada, however, may result in unemployment benefits that immigrants receive in major urban areas with low unemployment rates being lower than the benefits that immigrants receive in major urban areas in the USA.

  11. McDonald et al. (2012) find that foreign-trained doctors in Canada faced a lower probability of working as a physician than foreign-trained doctors in the USA, which they attribute to differences in selection policies.

  12. The 1996 and 2004 panels span 48 months and the 2001 is 36 months.

  13. We also conducted analysis keeping all 6 years of data for SLID, and the results were similar.

  14. Our analysis includes the self-employed to offer a more complete picture. We conducted sensitivity analysis, dropping self-employed, and found similar results for both men and women.

  15. Together, excluding those who arrived before 17, those in school, and immigrants from Canada in the USA (and vice versa) drops about 10 % of each the SLID and SIPP samples. We also did the analysis including immigrants from Canada to the USA and immigrants from the USA in Canada. The results were similar to those reported.

  16. We replace non-response months with average monthly hours worked in that year.

  17. Roughly 1 % of each sample is excluded in hours worked models due to this restriction.

  18. We also conducted similar analysis with hours worked as dependent variable assigning 0 h worked to non-workers. Estimates were similar and can be obtained upon request.

  19. Total annual earnings in SIPP are derived by summing the monthly earnings in each year. We replace non-response months with the average of non-missing monthly earnings in that year.

  20. Roughly 1 % of each sample is excluded in hourly wage models due to the wage restriction. We also did the analysis by dropping the wage restriction, and the results were similar to those with the restriction.

  21. The initial response rate in the SIPP sample is 91.6 % in the 1996 panel, 87 % in the 2001 panel, and 85 % in the 2004 panel. Including initial nonresponse and attrition, the cumulative sample loss was 35 % (over 12 waves) in 1996 panel, 31 % (over 9 waves) in 2001 panel, and 37 % (over 12 waves) in 2004 panel (National Research Council 2009; US Census Bureau 2008). By wave 8, the rates of sample loss for the 1996, 2001, and 2004 panels were 31, 30, and 34 %, respectively (US Census Bureau 2008). SLID has comparable response rates to SIPP. In panel 2 (beginning in 1996), the initial response rate was 89.5 %, which had fallen to 82.7 % by the fourth year of the panel. Over time, the representativeness of each panel has diminished, such that by panel 5 (starting in 2005), the initial response rate was 78.8 %, which fell to 72.8 % by the fourth year of the panel. For details, see http://www23.statcan.gc.ca/imdb-bmdi/document/3889_D13_T2_V2-eng.pdf. Table 5.2 p. 16.

  22. The distinction between sample attrition due to return migration, onward migration, and attrition due to other reasons is important (see Aydemir and Robinson 2008). Unfortunately, our data do not provide information that would allow us to separate the cause of attrition.

  23. We did the analysis including age controls as dummy variables. The results were similar to those reported. This is expected. Our models include individual fixed effects. Therefore, only a small proportion of the samples report variation in age dummies across years.

  24. Studies document many non-economic reasons for return migration (see Klinthäll 2006, 2007; Masferrer and Roberts 2012 and Maron and Connell 2008).

  25. We evaluated how sample attrition may affect the results by comparing models based on respondents present in at least two waves and those present across all waves of the survey. The results are qualitatively similar (see Appendix Table 9), suggesting that attrition is not likely to seriously bias the results.

  26. The last year of observation in the Canadian data is 2008—the beginning of a recessionary period (the US data covers through 2007 only). We conduct analysis excluding 2008 and find similar results for employment status and hourly wage for both the men’s and women’s analyses. The increase in hours worked for the 0–5-year male cohort remains similar in magnitude but is reduced to non-significance.

  27. We further explored the decline in employment among earlier cohort of immigrants in Canada by stratifying the data based on age in the base year of the survey in two groups: men aged 25–42 and men aged 43–59. Our estimates indicate that the decline in employment is significant only for immigrant men aged 43–59.

  28. The results in Tables 2 and 3 show that employment levels and hours worked (the coefficient on the trend variable) are decreasing throughout the sample period in both countries. This means that the composition of those that are working is also changing in these countries. To provide a complete picture, we present trajectories of employment, hours worked, and wage. Note that our longitudinal analysis, unbalanced panel is based on respondents who are in the sample for at least two waves and balanced panel is based on respondents who are in the sample in all waves. Thus, our analysis is less affected by the compositional changes than a corresponding analysis based on cross-sectional data.

  29. Immigrant and native outcomes may be differently affected by national policies due to a number of factors including immigrant eligibility and knowledge of these policies (or programs). Immigrants may also be culturally averse to utilizing welfare programs.

  30. We did the cross-sectional analysis with two samples: (i) all respondents in the first year and (ii) all respondents in year 1 who are present in at least 2 years. The results from these analyses were qualitatively similar.

  31. The issue of collinearity is also resolved as we include cohort of arrival and years since immigration as a set of dummy variables as indicated above rather than linear variables.

  32. One potential cause of difference between Picot and Piraino and our study is that we examine the trajectories of all Canadian immigrants who were 17 or more at the time of entry, whereas Picot and Piraino focus on Canadian immigrants who were between 25 and 44 years of age at entry. We also study the short-term labor market trajectories whereas Picot and Piraino study long-term trajectories.

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Acknowledgments

The authors are grateful for support by the National Science Foundation (SES 1226546), the Russell Sage Foundation, and the Columbia Population Research Center with funding from the Eunice Kennedy Shriver National Institute of Child Health and Human Development (R24 HD058486). The study also benefited from the SIPP data workshop and conference travel grant funded by the National Science Foundation (SES 1131897). The analysis presented in this paper was conducted at the Quebec Interuniversity Centre for Social Statistics, which is part of the Canadian Research Data Centre Network (CRDCN). The services and activities provided by the QICSS are made possible by the financial or in-kind support of the SSHRC, the CIHR, the CFI, Statistics Canada, the FRQSC, and the Quebec universities. The views expressed in this paper are those of the authors and not necessarily those of the CRDCN or its partners. The authors acknowledge the generous help and guidance of three anonymous referees and the editor that tremendously improved the paper.

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Authors and Affiliations

  1. School of Social Work, Columbia University, 1255 Amsterdam Avenue Room 917, New York, NY, 10027, USA

    Neeraj Kaushal & Julia Shu-Huah Wang

  2. Department of Sociology, Columbia University, 606 West 122nd Street, 614 Knox, New York, NY, 10027, USA

    Yao Lu

  3. Department of Sociology, McGill University, Leacock Building, 855 Sherbrooke Street West, Montreal, Quebec, H3A 2T7, Canada

    Nicole Denier

  4. Department of Social Work and Social Administration, The University of Hong Kong, Pok Fu Lam, Hong Kong

    Julia Shu-Huah Wang

  5. Department of Economics, University of Texas, 2225 Speedway Stop C3100, Austin, TX, 78712-1690, USA

    Stephen J. Trejo

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  1. Neeraj Kaushal
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Correspondence to Neeraj Kaushal.

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Appendix

Appendix

Table 7 Descriptive statistics
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Table 8 Estimates of labor market outcomes of native-born and immigrant men in Canada and the USA (models using longitudinal data with longitudinal weights and individual fixed effects)
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Table 9 Estimates of labor market outcomes of native-born and immigrant men in Canada and the USA (models using longitudinal data with individual fixed effects: sample: individuals present in all years of a panel)
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Kaushal, N., Lu, Y., Denier, N. et al. Immigrant employment and earnings growth in Canada and the USA: evidence from longitudinal data. J Popul Econ 29, 1249–1277 (2016). https://doi.org/10.1007/s00148-016-0600-5

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