NBER WORKING PAPER SERIES

THE SHORT-AND LONG-TERM CAREER EFFECTS OF GRADUATING IN A

RECESSION: HYSTERESIS AND HETEROGENEITY IN THE

MARKET FOR COLLEGE GRADUATES

Philip Oreopoulos

Till von Wachter

Andrew Heisz

Working Paper 12159

http://www.nber.org/papers/w12159

NATIONAL BUREAU OF ECONOMIC RESEARCH

1050 Massachusetts Avenue

Cambridge, MA 02138

April 2006

We would like to thank David Card, Ken Chay, Pierre-André Chiappori, Damon Clark, John DiNardo,

David Figlio, David Lee, Justin McCrary, Bentley McLeod, Daniel Parent, Mike Riordan, Eric

Verhoogen, and participants at the NBER Summer Institute 2005 and at seminars in UC Berkeley,

Cornell, UCLA, Stanford, Columbia University, University of British Columbia, University of Maryland,

University of Michigan, University of Florida, the Bank of Italy, and Tor’ Vergata for helpful comments.

All remaining errors are our own. Corresponding author: Till von Wachter vw2112@columbia.edu.

The views expressed herein are those of the author(s) and do not necessarily reflect the views of the National

Bureau of Economic Research.

©2006 by Philip Oreopoulos, Till von Wachter and Andrew Heisz. All rights reserved. Short sections of

text, not to exceed two paragraphs, may be quoted without explicit permission provided that full credit,

including © notice, is given to the source.
The Short- and Long-Term Career Effects of Graduating in a Recession: Hysteresis and

Heterogeneity in the Market for College Graduates

Philip Oreopoulos, Till von Wachter and Andrew Heisz

NBER Working Paper No. 12159

April 2006

JEL No. J0, J3, J6, E3

ABSTRACT

The standard neo-classical model of wage setting predicts short-term effects of temporary labor

market shocks on careers and low costs of recessions for both more and less advantaged workers.

In contrast, a vast range of alternative career models based on frictions in the labor market suggests

that labor market shocks can have persistent effects on the entire earnings profile. This paper

analyzes the long-term effects of graduating in a recession on earnings, job mobility, and employer

characteristics for a large sample of Canadian college graduates with different predicted earnings

using matched university-employer-employee data from 1982 to 1999, and uses its results to assess

the importance of alternative career models. We find that young graduates entering the labor market

in a recession suffer significant initial earnings losses that eventually fade, but after 8 to 10 years.

We also document substantial heterogeneity in the costs of recessions and important effects on job

mobility and employer characteristics, but small effects on time worked. These adjustment patterns

are neither consistent with a neo-classical spot market nor a complete scarring effect, but could be

explained by a combination of time intensive search for better employers and long-term wage

contracting. Allresults are robust to an extensive sensitivityanalysisincluding controlsfor correlated

business cycle shocks after labor market entry, endogenous timing of graduation, permanent cohort

differences, and selective labor force participation.

Philip Oreopoulos

Department of Economics

University of Toronto

150 St. George Street

Toronto, ON M5S 3G7

Canada

and NBER

oreo@economics.utoronto.ca

Till von Wachter

Columbia University

1022 International Affairs Building

420 West 118th Street

New York, NY 10027

www.columbia.edu/~vw2112

Andrew Heisz

Statistics Canada - Analytical Studies Branch

120 Parkdale Avenue

24th Floor, R.H. Coats Building

Ottawa, Ontario K1A 0T6

Canada

andrew.Heisz@statcan.ca
2 

1. Introduction 

In the standard neo-classical view, labor markets function like perfectly competitive spot markets and 

any effects of bad luck – such as from a labor demand shock or a recession – are temporary and 

inconsequential for wages and career development (Friedman 1953). In such a setting, lasting wage 

differences among similar workers arise from leisure preferences or from choices over human capital 

acquisition, rather than luck (Mincer 1974). Within this framework, typical estimates imply small welfare costs 

of recessions (Lucas 1987). On the other extreme, models based on institutional wage setting or segmented 

labor markets predict that temporary shocks can have permanent consequences, especially for less advantaged 

workers (Doeringer and Piore 1971, Okun 1973). More recent theoretical models of wage determination 

based on job search (Burdett 1978, Topel and Ward 1992), contracting (Harris and Holmstrom 1982, 

Macleod and Malcolmson 1994), or efficiency wage considerations (Akerlof and Yellen 1985, Acemoglu 

2001) also emphasize the role of frictions in the labor market, but predict persistent, not permanent, negative 

effects on career outcomes from temporary labor market shocks. 

 Recent empirical work uses exogenous variation in economic conditions to test directly for the shortand long-term effects of luck (Bertrand and Mullainathan (2001) for CEOs; Baker, Gibbs, and Holmstrom 

(1994) for workers at a large firm; Jacobson, Lalonde, and Sullivan (1993) for workers displaced by mass 

layoffs; Welch (1979) and Freeman (1979) for workers in large cohorts). Although these studies find that 

certain groups of workers experience highly persistent effects from random luck, it may be difficult to 

generalize these findings to the effects caused by a commonly recurring labor market shock – such as a 

recession – and for a more representative group of workers. However, another strand of recent literature, 

based on more representative samples of the labor market, suggests that frictions play an important role in 

the wider labor market as well, with job mobility (Topel and Ward 1992, Neal 1999), wage differentials 

(Dickens and Katz 1987, Krueger and Summers 1988, Abowd, Creecy, and Kramarz 2002) and long-term job 

attachment (Farber 1999, Gibbons and Waldman 1999) all playing an important role. Alternative models 

implied by these stylized facts have starkly different predictions on the long-term effect of short term labor 

market shocks. However, few papers have used variation from recessions to learn about the role of frictions 

in the labor market, and whether such shocks and frictions affect some workers more than others.2

 In this paper, we analyze the short- and long-term effects of graduating in a recession on college 

graduates with different educational backgrounds using an unusual match between administrative universityemployee-employer data from Canada. We thereby fill three important gaps in the literature. First, we analyze 

the effect of a typical recurring shock – economic downturns – affecting a large group of workers, while 

 

2

 Important exceptions are Beaudry and DiNardo (1991) and some of the literature on ‘scarring,’ further discussed 

below. Although a long literature assesses the current effects of recessions for low-income workers, little is known 

about the long-term effects (e.g., Hines, Hoynes, and Krueger 2002, Cutler and Katz 1991). 
3 

minimizing the confounding effects of correlated shocks in the labor market or selection. Second, we use the 

detailed longitudinal information on employment, job mobility, and employer characteristics covering a 

horizon of ten years to assess the empirical relevance of competing models that predict persistent effects of 

early labor market conditions. Third, unique information about an individual’s school and academic program 

allows us to investigate whether workers predicted to earn less suffer larger long-term losses and adjust 

differently to an initial shock. 

Young workers are an ideal group to study the importance of ‘luck’, since they may be particularly 

affected by temporary labor market conditions (Katz and Autor 1999, Freeman 1979, Okun 1973).3 Since 

labor market entrants have similar prior labor market experience, they are affected by luck in a comparable 

moment of their career and can be more easily categorized into more and less advantaged groups based on 

their university background.4 In addition, the unique data set at our disposition allows us to address several 

important empirical challenges inherent in estimating the long-term impact from initial labor market shocks. 

First, the distinction between temporary and persistent shocks is crucial to isolate the role of luck. The broad 

range of cohorts in our sample yields sufficient variation to isolate the effect of the very first labor market 

conditions. Second, typically there are too few recurring labor market shocks affecting large groups of similar 

workers. Our data cover almost the universe of 20 years of graduating cohorts encompassing two large 

recessions with differential strength across ten regions.5 Third, detailed information on longitudinal 

employment patterns and the timing of college entry and exit allows us minimize the confounding effects 

from selective participation or graduation. 

This is the first paper to use matched university-employer-employee data to systematically analyze the 

effects of initial luck on the entire process of dynamic adjustment for a large number of cohorts and groups 

of college graduates with different earnings capacities. Our results suggest three main findings. First, luck 

matters – graduating in a recession leads to large initial earnings losses that do fade, but slowly and over an 

eight to ten year period after graduation. A typical recession -- a rise in unemployment rates by five 

 

3 During the first 10 years of work, individuals experience 70% of overall wage growth, cha nge jobs frequently, and 

find a career occupation and employer (Murphy and Welch 1990, Topel and Ward 1992). In this formative period young 

workers are particularly at risk of earnings losses (Blanchflower and Oswald 1994), job losses (Farber 2003), and 

unemployment (Ryan 2001) from a recession. Similarly, a long literature documents how disadvantaged workers are hurt 

most in recessions and may gain most in expansions (e.g., Hines, Hoynes, and Krueger 2002, Freeman 2001). 4 For more mature workers, hard-to-observe differences in on-the-job experience, training, and job levels may not be 

approximated well by potential labor experience. Thus, a labor market shock at a given year of potential experience may 

affect them at different points in their actual career development and at different skill levels. Similarly, college 

background is a worse predictor of potential earnings as workers accumulate experience. 5 Several previous studies on the persistent effects of aggregate labor market conditions have used the Panel Study of 

Income Dynamics (Devereux 2003) and the National Longitudinal Studies of Youth (Gardecki and Neumark 1998, 

Neumark 2002, Kahn 2005). While providing detailed survey information on careers and worker demographics, the 

small samples of these data sets do not allow controlling for cohort, state, and year effects, controlling for persistent 

correlated labor market conditions, or studying other career outcomes than wages with sufficient degree of precision. 

Moreover, often the range of cohorts studied covers a limited range of business cycles. For example, Kahn’s (2005) 

careful study of white male college graduates can exploit longitudinal information on somewhat more than 500 workers 

graduating around one major recession (1982). 
4 

percentage points in our context -- implies an initial loss in earnings of about 9 percent that halves within 5 

years and finally fades to zero by 10 years. This result is robust to the use of both national and regional 

unemployment rates, does not appear to arise due to correlation with labor market shocks occurring later in 

workers’ careers, and does not seem to be due to selective employment and graduation decisions. 

Second, initial random shocks affect the entire career profile, including systematic changes in job 

mobility and firm characteristics. We find that temporary reductions in the quality of the first employer can 

explain 30-40% of initial earnings losses. The dynamic adjustment process is characterized by increased 

mobility across employers and industries and improvements in average firm characteristics, but little effects 

on employment or time worked. Only workers with the lowest predicted earnings are permanently downranked to firms paying lower wages. We also find an important degree of persistence of unemployment rate 

shocks within firms, especially for very large and high paying employers. 

Third, some workers are more affected by luck than others. In particular, earnings losses from 

temporarily high unemployment rates are strongest for labor market entrants and small for workers with two 

or more years of labor market experience. Among graduates, those at the bottom of the wage distribution and 

those with the lowest predicted earnings suffer significantly larger and much more persistent earnings losses 

than those at the top. Thus, we see a temporary increase in inequality from recessions lasting up to ten years 

and lasting increases in inequality in the present discounted value of earnings. 

For all but the most disadvantaged workers, these findings appear inconsistent with the “long-term 

scarring” view of the labor market. Yet, the “perfect spot market” model of labor market may only be a good 

approximation in the very long run (a time horizon of 10 years). In the short- to-medium run the observed 

patterns of job mobility and firm characteristics imply a process of time-intensive job search in the presence 

of wage differentials (e.g., Okun 1973, Krause and Lubik 2005). In such an environment, least able workers 

are hurt most in recessions as firms raise their hiring standards (e.g., Hall 1974), consistent with what we 

find.6 Yet, job search alone cannot explain differences in the effects between labor market entrants and more 

mature workers, and the finding that cyclical effects are more persistent for workers starting to work at large, 

high-paying employers. These differences may be due to the presence of long-term implicit contracts (Harris 

and Holmstrom 1982), or due to cyclical job assignment and differences in on-the-job training (Gibbons and 

Waldman 2004). Overall, it seems that a combination of search and contracting models is necessary to fully 

explain patterns of recovery from temporary demand shocks seen in our data. 

Our paper contributes to several recent strands of literature. By examining the long-term effects of a 

typical labor market shock free of confounding effects from persistent shocks and selection for a 

representative number of cohorts, a wide range of career outcomes, and workers of different experience and 

ability, it extends the existing literature documenting that luck can matter for single occupations (Betrand and 

 

6 However, also consistent with our results is a process of sequential sorting with employer learning as in Gibbons, Katz, 

Lemieux, and Parent (2005) in which workers are sorted into high-wage firms (Fox 2004). 
5 

Mullainathan 2001, Oyer 2005), for job losers (Jacobson, Lalonde, and Sullivan 1993, von Wachter and 

Bender 2005), or for single cohorts of workers (Freeman 1975, Welch 1979, Kahn 2005). 

This is also one of the few papers showing that job mobility and employer characteristics respond 

systematically to exogenous labor market shocks (von Wachter and Bender 2006). It thereby extends earlier 

studies on the role of job search without access to exogenous variation or detailed information on employers 

(Topel and Ward 1992); studies on the effect of unemployment shocks within firms (Beaudry and DiNardo 

1991, Baker, Gibbs, and Holmstrom 1994); and studies suggesting an important role of industry and firm 

characteristics in wage determination (e.g., Dickens and Katz 1987, Krueger and Summers 1988, Abowd, 

Creecy, and Kramarz 2002).7

Our paper also contributes to two strands of literature in macroeconomics. First, while 

macroeconomists typically measure the cost of recessions for a representative worker (Lucas 1987),8 a large 

literature suggests that less able workers may be hurt most in recessions (e.g., Hines, Hoynes, and Krueger 

2002, Freeman 2001). Our data allows us to provide actual estimates of the change in the present discounted 

value of earnings due to a recession for more and less advantaged workers. Second, our evidence on the 

response of wage setting and job mobility to recessions suggests an important role of search frictions in 

generating cyclical labor market dynamics (Hall 2005, Shimer 2005), but also suggests an important role for 

mobility across firms and sectors (Krause and Lubik 2005) and long-term wage contracts (Beaudry and 

DiNardo 1991). 

The rest of the paper is organized as follows. Section 2 gives an overview of relevant models of 

career determination. Section 3 outlines the empirical strategy we use to address these questions and our 

approach to deal with selective labor market entry, selective labor force participation, and the presence of 

correlated unemployment shocks. The main results on earnings, employment, job mobility, and firm 

characteristics are discussed in Section 4, followed by an analysis of heterogeneity. A detailed sensitivity 

analysis addressing among others the question of selection and omitted variable bias due to recurring 

aggregate unemployment conditions is summarized in the Appendix. The sixth section discusses our results in 

light of different models of dynamic adjustment. The last section concludes. 

2. Models of Career Determination 

The following section briefly reviews the main models of career determination that have guided the 

interpretation of our results. Although most features of these models are relevant for the full labor market, we 

focus the discussion on young workers who may be hurt most by conditions in the external labor market 

 

7 By analyzing workers’ adjustment patterns to a recession shock, it also contributes to the recent macroeconomic 

literature on job and earnings dynamics over the business cycle (e.g., Hall 2005, Shimer 2005, Krause and Lubik 2005). 8 Lucas’ original measure asked how much additional consumption would make workers indifferent between economies 

with and without consumption risk based. More recent papers distinguish between individuals with and without wealth 

holdings, or between workers on the job and those laid-off (see, e.g., Barlevy 2005 for an excellent survey). 
6 

(e.g., Freeman 1975, Katz, Loveman, and Blanchflower 1995, Katz and Autor 1999).9 Existing results indeed 

suggest young workers may be hurt most by recessions, but little is known about the degree of persistence of 

these losses.10

 As a common benchmark, the neo-classical model of wage determination with or without human 

capital accumulation does not predict long-term effects from short term labor market shocks. Without human 

capital accumulation, the basic neo-classical model predicts earnings should only depend on current labor 

market conditions in similar fashion for all age groups, and long-term effects of early labor market conditions 

only arise due to correlated persistent labor market shocks.11 If human capital accumulation occurs on the job 

(e.g., Mincer 1974), long spells of non-employment or reduced working hours initially lower earnings growth 

for younger workers. However, the standard model has difficulty explaining lasting effects of recessions that 

involve short spells of unemployment as typically observed for young college graduates. Extensions of the 

standard model assign a role to the accumulation of firm specific capital (e.g., Becker 1964) or industry 

specific capital (e.g., Neal 1995). If persistent early labor market conditions lead to increased job mobility, 

workers may initially accumulate less specific human capital, and catch up by finding stable employment in a 

firm or industry.12

Some more recent models of career determination imply that even temporarily adverse labor market 

conditions can have persistent effects, particularly for workers who just entered the labor market. As opposed 

to the ‘standard human capital’ view, these could be broadly termed the ‘contracting’ or ‘within firm’ view, 

and the ‘search’ or ‘between firm’ view. The full implications from these models for the pattern of adjustment 

to labor market shocks have not been fully exploited in the literature. 

Among the several variants of the ‘within firm’ view (e.g., Prendergast 1999, Gibbons and Waldman 

1999), two are particularly relevant. First, models of long-term contracting (e.g., Harris and Holmstrom 1982, 

 

9 Katz, Loveman, and Blanchflower (1995) and Katz and Autor (1999) suggest that demand shifts towards high-skilled 

labor may have hurt younger unskilled workers more than mature unskilled workers. The same may hold for changes in 

relative supply (Freeman 1975). A large theoretical literature has invoked the presence of implicit contracts for mature 

workers to explain the weak cyclicality of real wages for mature workers (e.g., Prendergast 1999, Malcolmson 1999). 10 E.g., Clark and Summers (1981), Blanchflower and Oswald (1994), and Farber (2003) document the responsiveness of 

young workers’ employment, wages, and job stability to cyclical shocks. We know that job movers have more procyclical wages than workers on the job (e.g., Devereux 2001, Bils 1985), and that job finding rates are more pro-cyclical 

than job separation rates (e.g., Shimer 2005b). Since labor market entrants are most likely to be among those searching 

for jobs (Topel and Ward 1992) and most likely to be unemployed (Clark and Summers 1981), this further underscores 

their vulnerability to cyclical shocks 

11 This could be modeled as temporary adjustment processes, as in a standard neo-classical model with flexible (but 

possibly slow) wage adjustment. It could also arise within a neo-classical model augmented with an explicit equilibrium 

relationship between wages and unemployment. A complete model of regional wage and employment adjustment with 

and without unemployment is outlined in Blanchard and Katz (1992). 12 If workers pay for training by wage cuts (Mincer 1974) and jobs offer different degrees of learning (Rosen 1972), then 

earnings trajectories may also be affected by changes in the availability of jobs in the economy. However, if the amount 

of ‘learning’ jobs declines, then wage trajectory should be characterized by higher entry wages and lower growth rates. 

On the other hand, if workers forsake more earnings in recessions to have faster catch-up, we may underestimate initial 

earnings losses. 
7 

MacLeod and Malcolmson 1993) have been used by Beaudry and DiNardo (1991) to explain persistent effects 

of labor market conditions when workers enter a firm. Second, models of job assignment (e.g., Prendergast 

1993) can also imply persistent recession effects within firms. For example, Gibbons and Waldman (2004) 

rationalize persistent conditions found by Baker, Gibbs, and Holmstrom (1994) by the presence of task 

specific skills. Since workers should move employers to avoid persistent losses, firms-specific effects of 

recessions require some form of barrier to job mobility. This is more likely to occur in large firms that may 

provide rents, insurance, career-options, or other non-pecuniary benefits not captured in the wage, something 

we will analyze explicitly below. 

 Neither the standard approach nor the ‘within firm’ approach explicitly model job change. Yet, 

beneficial job mobility as predicted by models of search appears an integral part of young workers’ careers 

(e.g., Topel and Ward 1992). For workers entering the labor market in a recession, search models predict that 

a temporary worsening of the wage offer distribution leads to a time intensive catch-up process. This process 

involves a high degree of job mobility with wage gains concentrated at job changes rather than accruing on 

the job (e.g., Burdett 1978). Given the prevalence of job mobility for young workers job search is likely to be 

an important channel of adjustment in the face of early unemployment shocks. However, although the job 

search model is featured in many empirical and theoretical analyses of the labor market, as of yet few papers 

analyze the causal determinants of job search, the role of search in the adjustment to labor market shocks, or 

the role of firm characteristics in the search process.13

The analysis of employer characteristics is important for understanding how workers advance in the 

labor market and respond to shocks. Recently, a number of studies suggest firms pay rents that cannot be 

explained by worker characteristics (e.g., Acemoglu 2001, Abowd, Kramarz, and Margolis 1999, Abowd, 

Creecy, and Kramarz 2003). Clearly, if high wage jobs are concentrated in firms that pay more on average, job 

search leads workers to switch towards high wage employers. Similarly, a longstanding hypothesis in labor 

economics suggests some firms and sectors pay rents, that high wage firms have more pro-cyclical job 

creation, and that young and less able workers are down ranked as employment at high wage firms and 

sectors contracts in recessions (e.g., Reynolds 1951, Reder 1955, Okun 1973, Cutler and Katz 1991, Hines et 

al. 2002). In this case, the initial loss from graduating in a recession occurs due to a reduction in employer 

quality, and the catch-up process involves transitions to better employers.14 

 

13 Since the seminal characterization of the job mobility process over workers’ careers by Topel and Ward (1992), several 

authors have refined the notion and determinants of search. For example, McCall (1990) and Neal (1999) extend the 

definition of the search process to include occupations and industries and argue that some job switches have larger 

effects on future search than others. Yet, few papers estimate the determinants or differences in the pattern of job 

search, with the exception of the role of gender (e.g., Ureta and Light 1995, Keith and McWilliams 1999). Similarly, few 

papers analyze the role of firms. An exception are Topel and Ward (1992) themselves, who note that transitions to larger 

firms lead to more stable jobs and a related literature documenting correlation in turnover rates and firm size (e.g., 

Anderson and Meyer 1994) 

14 A process of cyclical upgrading can be motivated in a multi-sector model with queuing due to search frictions or some 

form of wage rigidity (McDonald and Solow 1985, Akerlof and Yellen 1985). Krause and Lubik (2005) formalize this 
8 

Such a process of cyclical upgrading is supported by several stylized facts. First, industries pay wage 

premiums that cannot be easily rationalized by worker characteristics (e.g., Krueger and Summers 1988, 

Dickens and Katz 1988). Second, sectors paying higher wages have more pro-cyclical job creation, partly 

because of more volatile demand for their products (Okun 1973, McLaughlin and Bils 2001, Aaronson and 

Christopher 2004). Third, less able workers tend to flow to larger firms and high wage sectors in booms 

(Vroman 1977, Albaek and Sorensen 1998, Devereux 2002). Yet, to our knowledge few papers analyze the 

effect of labor market shocks on job transitions between different types of firms and industries (von Wachter 

and Bender 2006). 

Standard career models do not predict differential effects of recessions on less advantaged workers; if 

wages are equal to workers’ marginal product, similar shocks should lead to comparable losses. Only in the 

presence of rents do firms have incentives to select the most able workers for employment, and to reduce the 

employment of less able workers. In this case, labor market entrants are more affected than workers with 

greater labor market experience. Similarly, among labor market entrants, the least advantaged workers are 

likely to be most affected by cyclical conditions in the labor market.15 

We know that low-skilled workers, non-whites, or women lose most in recessions in terms of wages 

(Hoynes 2000, Hines et al. 2002) and are most likely to drop out of the labor market (Bils 1985, Solon, 

Barsky, and Parker 1994). Yet, a high degree of heterogeneity in college quality or parental background 

suggests that even within the group of male college graduates those least advantaged may bear most of the 

burden of recessions. Large differences in the long-term cost of labor market shocks among prime earners 

would be a further indication that uniform measures may understate the true cost of business cycles (e.g., 

Lucas 1987, Barlevy 2005). 

3. Empirical Strategy and Data 

3.1. Basic Earnings Model 

To measure the long-term effects on earnings of starting to work in a recession, our main 

specification exploits cyclical variation in unemployment rates for young workers at the regional level. Since 

 process in a general equilibrium framework with good and bad employers and on the job search. Alternatively, it could 

arise due to a process of sorting by comparative advantage (Gibbons, Katz, Lemieux, and Parent 2002, McLaughlin and 

Bils 2001). A process of cyclical up- and down-grading due to the adjustment of hiring and promotion standards was 

first described by Reynolds (1951) and Reder (1955), and taken up by Okun (1973, 1981), and Hall (1974). Barsky and 

Solon (1989), Solon, Whatley, and Stevens (1997), and Devereux and Hart (2005), among others, examine the 

importance of such a process within firms. 

15 A cyclical process of adjustment in hiring and promotion standards has been often noted (e.g., Reder 1955). Barsky 

and Solon (1989) develop a numerical example. Generally, any model in which wages contain rents that vary inversely 

with worker ability predicts that high-wage firms should try to selectively hire more able workers, giving rise to the 

presence of queues. Such rents can arise due to rigid pay scales as in Hall (1974), or unions, as in Solow and MacDonald 

(1988). Lemieux (1998) describes the two-sided selection process for the case of unionized firms. While layoffs may be 

over-proportionately composed of less able workers in recessions if they accumulate less firm specific skills, as suggested 

by Neal (1998), this should affect low-skilled labor market entrants only indirectly via increased competition in the labor 

market. 
9 

our main independent variable – the rate of unemployment – varies across province and over time, we 

collapse the individual level data at the level of graduation cohort, initial region of residence, and calendar 

year and work only with the cell means crt y of the log of annual earnings and other variables (weighted by 

the corresponding cell sizes).16 The cell level model on which most of the estimates in the paper are based on 

is 

crt eURcr t r e c ucrt y = α + β 0 + φ +θ + γ + χ + (1) 

where θ r , χ c , e γ , and φt represent unrestricted fixed effects for first region of residence (r), year of 

graduation (c), year of potential labor market experience (e), and calendar year (t). The unemployment rate is 

measured at the time of graduation and the region of first residence (URcr 0 ). The coefficients β e on the 

interaction of potential experience with the unemployment rate in the year of college graduation are the main 

effects of interest. To account for group specific error-components, we cluster standard errors at the cohortregion level. 

Our data, further described below, allow us to observe almost the universe of male college graduates 

in Canada from the end of their first college degree every year for ten years into their careers. In our main 

specifications, we follow 20 cohorts entering from 1976 to 1995 in ten regions. Cohort effects control for 

permanent differences in earnings of different graduation years at the national level.17 Cohort dummies also 

eliminate earnings effects due to national time series variation in unemployment rates that is common across 

provinces. Province fixed effects ensure that we estimate effects of initial labor market conditions from 

changes in unemployment rates at the provincial level. Given the presence of unrestricted experience effects, 

the coefficients β e on the initial unemployment rates we estimate measure changes in experience profiles in 

earnings and other outcomes resulting from province-cohort-specific variation in unemployment rates. 

We interpret the variation in UR to arise from changes in aggregate labor demand that are 

uncorrelated with characteristics of different graduation cohorts. Year-to-year variations in cyclical labor 

market conditions that identify our estimates move at a higher frequency than age-specific population, 

 

16 The only individual characteristics we could include are the actual duration of college and age. Instead of including 

years of college in an individual level model, we split the sample between workers who graduated and all workers with 

some college. 

17 That earnings may permanently differ across cohorts has been noted in previous research (e.g., Welch 1979, Card and 

Lemieux 2001). As it is well-known, cohort, potential experience, and year effects cannot be identified separately without 

an additional restriction on cohort effects is needed. Since we are mainly interested in experience effects and in their 

change over the business cycle, we simply drop one additional cohort effect from the regression. We could have chosen 

to restrict cohort effects to sum to zero (as suggested by Deaton 1997). This alternative does not alter our estimates of 

the experience profile. Of course, the resulting cohort effects differ. We also have assessed the linearity assumption 

implicit in Equation (1) by plotting and regressing the residuals of earnings and unemployment rates (from first stage 

regressions on province, year, and cohort dummies) by experience years. The results (shown in Appendix Figure 2) 

suggest that the linearity assumption is highly plausible. 
10 

participation, or enrollment trends.18 To make sure we pick up mainly effects occurring due to demand 

conditions and avoid influences from cohort-specific labor supply changes of young workers, in our 

sensitivity analysis we have also used the unemployment rate for all workers as measure of initial labor market 

shock.19 Below and in the appendix we address other potential biases due to changes in cohort quality from 

selective graduation and selective employment. We also present a wide range of specification and robustness 

checks, such as different sample restrictions or different cohort restrictions. 

At experience year zero our regression estimates the effects on earnings from current unemployment 

rates in region of residence plus region and year fixed effects. This specification is familiar from the wagecurve literature.20 By extending this basic model to allow for dynamic effects of the initial unemployment rate, 

we introduce the history of labor market shocks a worker has been exposed to as possible determinant of 

earnings. Since the wage-curve literature has shown that current unemployment rates affect workers of all 

ages, in some specifications we also allow the complete unemployment rate history since the start of the 

career to affect current wages. 

On the one hand, this allows us to isolate the effect of labor market conditions at entry net of 

subsequent effects on earnings from exposure to a possibly prolonged recession or expansion. Among others, 

this distinguishes the role of labor market conditions at entry from the effect of labor market conditions 

when entering a new firm in mid-career (as stressed for example by Beaudry and DiNardo 1991). On the 

other hand, this allows us to directly compare the effect of aggregate unemployment rates at time of entry to 

the effect of unemployment rates for more mature workers. The result will help to assess the exceptional 

position of young labor market entrants and give us a benchmark to gauge the degree of persistence of early 

labor market shocks. 

While our basic model can be easily extended to include persistent effects of unemployment rates in 

the region of residence at higher experience years, identifying persistent effects of market conditions at each 

experience level faces difficulties because of a lack of variation in regional unemployment rates in the data. 

Thus, we first consider including the current unemployment rate interacted with experience as an additional 

control. This extension assumes that only the unemployment rate at the time of labor market entry has 

dynamic effects on earnings. 

To gauge the limitation this restriction places, we also experimented with a more complete dynamic 

model. Suppose we allow for dynamic effects of the aggregate unemployment rate a worker was exposed to at 

 

18 Despite increases in college enrollment rates in Canada since the 1980s, the correlation between unemployment rates 

for young and old workers is high and it has remained stable (Beaudry et al. 2000). 19 Similarly, to assess the role of participation changes, we also replicated our results using the employment-populationratio for 15 to 24 year olds. 

20 However, unlike the wage-curve literature, the main focus of our paper is to estimate the long-term effects on earnings 

of temporary unemployment rate shocks rather than uncovering an equilibrium relationship between wage levels and 

unemployment rates. 
11 

each in experience year (e) in the relevant region ( er ), denoted by r eeUR (note that this unemployment rate 

will differ for different graduation cohorts). Denote the effect on earnings in experience year e from the 

unemployment rate at experience year 0 (1,2,3,4,...) by β e,0 ( , , , ...) β e,1 β e,2 β e,3 . Dropping other 

regressors and the region subscripts on the unemployment rates for simplicity, the complete dynamic model 

can be written succinctly as 

crt t r c e e URcr e URr e URr e URr crt y = φ +θ + χ + γ + β + β + β + + β + ε 0 0 1 1 2 2 10 10 1 2 10 , , , , ... , (2) 

where we impose the restriction s e β e,s = 0 ∀ < . The full dynamic regression estimates the effect of the 

transitory component of each aggregate unemployment condition, net of its correlation with other 

unemployment rates affecting the worker in adjacent experience years. 

This more complex model allows us to interpret the coefficient of basic regression model that omits 

control variables for subsequent unemployment rates after labor market entry. Without dynamic UR controls, 

our basic estimate of the effect of the first unemployment rate exposure captures the average change in 

earnings from graduating in a recession given the regular evolution of the regional unemployment rate faced afterwards. 

In other words, it estimates the dynamic effect of the first unemployment rate plus the weighted sum of the 

effect of unemployment rates a worker faces in his career.21 

 Due to high inter-temporal correlation of aggregate unemployment rates, it is difficult to estimate the 

fully unrestricted model in Equation (2). However, it turns out that more parsimonious models can yield the 

desired results. For example, the time series process of unemployment is often characterized as an AR(2) with 

high persistence of the first lag.22 We estimate the full dynamic model including unemployment rates at all 

periods fully interacted with experience dummies, as well as a more restricted model with only two additional 

unemployment rates. As expected, these models give similar results. 

In contrast to the basic estimate from Equation (1), the estimates of these dynamic models capture 

the effect of the very first exposure to unemployment rates, net of the effects of persistent local labor market 

conditions. Interestingly, the effect of the early unemployment rate turns out to be remarkably stable in all 

specifications we try. We would also like to use the full model in Equation (2) to compare the dynamic effect 

of unemployment rates at labor market entry with dynamic effects of unemployment rates once a worker has 

 

21 The omitted variable bias of the coefficients on the first unemployment rates is 

= + ∑

= −

e

d rc

rc r d

e e e d d UR

UR UR

p d

1 0

0 ,0 ,0 , var( )

cov( , ) ˆ lim β β β . 

22 In this case, the first two leads of unemployment (i.e., unemployment at experience years one and two) should remove 

omitted variable bias from the dynamic effect of the unemployment rate at experience zero. Some prefer to model it as 

ARIMA(1,1,0), but often a prior of stationarity is invoked to distinguish a persistent AR(2) from the similar nonstationary process (e.g., Blanchard and Katz 1992). 
12 

entered the labor market. However, it is difficult to separately identify dynamic effects of unemployment rates 

at adjacent experience years given the high persistence of unemployment innovations. We therefore also 

consider a restricted model in which we constrain the effects of unemployment to be the same in groups of 

experience years. This model is further discussed in the Appendix.23 

 As suggested above, our identification strategy allows us to obtain an estimate of the effect of early 

labor market conditions if there are no unobserved characteristics correlated with the initial unemployment 

rate that vary with experience. This may fail if early labor market conditions affect workers’ labor supply 

decisions (e.g., see Bils 1985, Solon, Barsky, and Parker 1994). The labor force exit of less able workers would 

lead to an understatement of any negative effect of unemployment rates on earnings, and our estimates would 

represent a lower bound (however, if worse workers reenter the labor market with a time lag, it would lead us 

to over-estimate persistence). As we show below, initial unemployment rates have small effects on the 

propensity to temporarily exit our earnings sample due to unemployment or non-employment. To further 

address the problem of selective labor force participation, the large sample sizes allow us to re-estimate our 

main models using a sample of workers with positive earnings in each period.24 In addition, to focus on a 

sample of men with stable labor force attachment, we drop those who permanently stop filing taxes at any 

point in time. Since this group may contain workers who emigrate (mainly to the U.S.) for economic reasons, 

we compare our main results including those who permanently stop filing in the sensitivity analysis. 

Individual characteristics might also correlate with unemployment rates at college exit if more able 

students remain in college longer to avoid graduating in a recession. If better student delay, that would lead 

us to overestimate the initial negative effect and underestimate the degree of persistence. This is unlikely to 

explain an important part of our estimates of persistent effects because the process of reversion of initial 

losses takes too long to be explained by lagged entry of more able workers. Two additional results lead us to 

believe the bias from selective entry is small. First, a detailed analysis of college duration in response to 

unemployment rates at actual and predicted time of graduation indeed suggests there are only small increases 

in the duration of college in response to unemployment rates for our preferred sample. Second, to also 

address the problem directly, we use information on the date of entry into college and official degree duration 

to construct predicted graduation dates for all graduates in our sample. We then use the unemployment rate 

in the predicted year of graduation as an alternative source of variation to identify the long-term effect of initial 

labor market conditions. In the appendix, we discuss two sets of estimates based on this additional measure. 

 

23 Instead of imposing restriction across experience years, we have also experimented with more parametric models of 

the decay of the initial effect of unemployment. While an exponential rate of decay is rejected by the data, the dynamic 

behavior of unemployment rate effects could be capture reasonably well by a fourth order polynomial in the time of the 

shock. However, this approach doesn’t solve the problems inherent in the data – too persistent unemployment shocks 

and too few cohorts at older ages. 

24 This is generally not possible in the smaller survey data sets used to assess this kind of bias (e.g., Bils 1985). To further 

check for selective participation decisions, we also analyzed changes in predicted earnings with experience and with early 

unemployment shocks. While there is a very small but significantly negative gradient in predicted earnings with 

experience, perhaps suggesting a pattern of migration, there is basically no correlation with early unemployment shocks. 
13 

First, we show reduced form estimates using the unemployment at predicted graduation as the relevant labor 

market shock. The second set of estimates present instrumental variable estimates using the unemployment 

rate at the predicted date of graduation as an instrument of the unemployment rate at actual graduation. Since 

the case can be made for either approach, we present both.25

3.2. Analysis of the Channels of Catch-Up 

Since initial labor market conditions will affect earnings through multiple channels, the estimated 

coefficients in the regression of log earnings represent reduced form effects of the initial unemployment rate 

on earnings-experience profiles. We cannot identify the effect of the separate channels because each is in 

itself affected by choices made by the worker and by his possibly unobservable characteristics. Instead, we 

provide an assessment of their importance by directly analyzing the impact of the unemployment rate on the 

channels themselves. We examine three sets of variables: first, outcomes related to employment such as 

receipt of unemployment insurance, filing income taxes with zero earnings, and not filing taxes. Second, 

outcomes relating to job mobility, such as changes in workers’ main employers, changes in industry of main 

employer (at the one, two, and three digit level), mobility across provinces, and whether workers left their 

first employer or their first industry. Third, we analyze the effect of graduating in a recession on the 

characteristics of young workers’ employers, further described below. 

An important step in attributing job mobility to a productive search process is to analyze whether it 

is voluntary and beneficial. To obtain a first insight into this question, we first follow Topel and Ward (1992) 

and characterize the average association of job and industry mobility with wage growth. Topel and Ward 

(1992) established that about 30-50% of initial quarterly earnings growth occurs at job changes, and we 

confirm similar results for annual earnings in Canada.26 We then estimate simple descriptive models of wage 

growth as a function of initial unemployment rates where initial unemployment rate is interacted with mover 

status, i.e., 

( )icrt t r c e e cr icrt icrt cr icrt ∆ earnings = + + + + + UR + Move + Move UR + u β 0 λ δ 0 log α φ θ χ γ , (3) 

where δ is the difference in the effect of initial unemployment rates on earnings growth by different mover 

status (indexed by the Move-dummy). While these parameter estimates clearly have no causal interpretation, 

they capture the importance of job change in the dynamic pattern of unemployment rate effects by showing 

whether the correlation of job change and job mobility increases for workers experiencing adverse initial 

conditions. 

 

25 Note that if the graduation date is endogenous, so are potential labor market experience and initial cohort effects. We 

have estimated a series of reduced form models with predicted cohort effects and predicted experience with little change 

in the main results. 

26 This is a stable pattern across countries (Giuliano and von Wachter (2004) report very similar numbers for France and 

Germany). 
14 

To directly assess the benefits of job mobility and to gauge the potential importance of firm 

heterogeneity, we also characterize the effect of early unemployment rates on the quality of young workers’ 

employers. We construct measures of average characteristics of each employer controlling for year effects, 

and merge these to the individual level data using workers’ employer identification numbers reported on tax 

returns. In this way, we construct average log median earnings, average total payroll, and average number of 

employees at the firm level, averaged over the entire period of existence of the firm for all employees (not 

only workers with college education) controlling for aggregate year and region effects. If more able workers 

are sorted into better firms, the wage measures do not necessarily capture the fixed firm component in pay 

rates, but a more inclusive measure of ‘firm quality’. Results by Abowd et al. (2003) suggest that this 

correlation may be weak. We use multiple measures of firm quality in case some measures include workers 

more selectively than others. 

If workers simply search for better firms, then a large fraction of initial wage losses should fade when 

we compare workers starting at the same firm in different cyclical conditions. To gauge the role of 

heterogeneity in firms’ payment structures in initial wage losses and their reversion, we examine how our 

baseline results change after including fixed effects for initial firm and initial industry. Clearly, these estimates 

are affected by potential change in the degree of selectivity into firms across the business cycle. If, as 

suggested at the outset, more able workers are down-ranked to lower wage firms in recessions, we may 

underestimate the degree of within firm wage changes.27 

Last, to further analyze the role of the first employer, to see whether high-wage firms have more procyclical wages, and to relate our main results to papers analyzing careers within firms, we also separately 

estimated the dynamic effects of initial unemployment rates by average wage or size of the initial employer. If 

careers evolve within firms, then changes in contracting, job assignment, or training over the business cycle 

might lead workers starting at large firms to have smaller but possibly more persistent wage declines than 

workers starting in small firms.28 

Another potential channel of how workers regain their lost earnings is mobility across provinces 

(Waggoner 2004). However, the descriptive analysis shows that regional mobility is much lower in Canada 

than the U.S. To assess directly whether reversion occurs by moves to economically vibrant provinces, we 

included the history of aggregate unemployment rates and fixed effects for current province of residence 

interacted with year effects as explained above. To check whether provincial mobility is beneficial for other 

 

27 To allow for firm specific experience profiles, we also include an interaction of initial firm fixed effects and experience 

effects, effectively allowing for firm specific experience profiles. Since young workers’ earnings may not be entirely a 

function of their ability (due for example to the presence of employer learning), including worker fixed effects or 

working with changes in earnings is not an ideal strategy to deal with this problem. 28 Again, the probability of starting to work at a ‘high quality’ employer may be correlated with workers’ ability, and the 

degree of selectivity might be affected by early unemployment rates. To address this problem, we have included control 

functions in the fraction of workers starting to work at ‘high quality’ firms. Similarly, we have included average fathers’ 

income as control function. Neither affects our results, and is available upon request. 
15 

reasons (for example due to the wider scope for good job matches), we have also analyzed the incidence and 

returns to provincial mobility directly as an outcome and found them to be small. In addition, variation at the 

national level represents shocks affecting the entire labor market whose effect is unmitigated by inter-regional 

mobility. Thus, we also present estimates based on variation of unemployment rates at the national level.29

3.3. The Heterogeneous Cost of Recessions 

Even among college graduates there may be important heterogeneity in earnings capacity due to 

variation in college quality or choice of major. Yet, if wages are equal to marginal products, more or less 

advantaged labor market entrants should not suffer different consequences from early recessions unless they 

face different demand shocks. To assess the importance of heterogeneity we use the college of graduation, 

the major, and the years of schooling to predict future earnings for each individual and reanalyze our data by 

quintiles and deciles of predicted earnings. Since these estimates are based on differences in average realized 

earnings of workers in different colleges and majors, they reflect the effect of college quality as well as the fact 

that more able workers may attend certain universities. Thus, we can compare the initial losses and reversion 

process suffered by workers with different initial earnings capacities. In addition, these estimates deliver an 

additional dimension of variation in the channels of recovery and the decline in earnings losses that will be 

helpful in assessing the underlying economic mechanisms driving the catch-up pattern. 

To characterize the overall cost due to cyclical fluctuations sustained by different groups in the 

population, we can use our estimates to approximate the present discounted loss of annual earnings arising 

from actual early recession shocks. This complements existing estimates of the costs of recessions based on 

the average standard deviation of consumption or earnings process. Most of these estimates are based on 

Lucas’ (1987) original exercise of comparing the present discounted value of utility derived from two 

consumption streams, one uncertain and one certain. Lucas asked by what proportion consumption has to 

rise to make workers indifferent between the two paths.30 Lucas’ initial findings of extremely small valuations 

 

29 The national model we estimate is 

ct t e e c ct w = + + + UR + c + c + u 2

0 0 1 log α φ γ β χ χ , (1) 

where the dependent variable is real earnings, and φt and e γ represent calendar year and potential labor market 

experience effects, respectively. The national model includes either linear or quadratic cohort trends, since unrestricted 

cohort effects are not identified. 

30 Specifically, Lucas compares the present discounted value (PDV) of utility from two consumption streams; one 

certain, { , ,...} *

2 * C1 C , and one uncertain { , ,...} 1 2 C C , where ( ) * 1 Ct t Ct = + ε , and epsilon is a white noise shock with 

constant variance. He then asks by proportion μ the uncertain stream has to be higher in each period than the certain 

stream to be of equal PDV utility. Using a constant relative risk aversion (CRRA) utility function with coefficient of 

relative risk aversion equal to one and estimates of the standard deviation of aggregate consumption, he derives that for 

the average worker μ is extremely small. More generally, Lucas’ calculations suggest that costs of recessions are very 

small unless risk aversion is extremely high. Lucas’ original study has been extended to take into account different form 
16 

of uncertainty have been revised in the literature in favor of more nuanced estimates taking into account 

imperfect capital markets, lack of savings, or concentrated job losses (e.g., Barlevy 2005). We replicate the 

classic Lucas measure for different groups in the population using the actual changes in the streams of annual 

earnings we estimate. Since none of these estimates use actual changes in earnings or consumption in 

response to a recession shock to estimate the cost of recessions or explores the role of heterogeneity in the 

costs of recessions, our estimates provide a useful complement to the existing literature. 

3.4. Canadian Administrative Data and the Evolution of Careers 

Our results are based on a unique match between three large administrative data sets collected and 

compiled within Statistics Canada that is described in detail in the Data Appendix. The data combines 

administrative information on about 70% of Canadian college students and college graduates from 1976 to 

1995 with longitudinal individual income tax records and firms’ payroll information covering the years from 

1982 to 1999.31 The data contains detailed information about individual students’ course of study (such as 

type of degree, major, date of graduation), with detailed career information (e.g., annual earnings, province of 

residence, receipt of unemployment benefits) and information on employers. Exploiting the panel nature of 

our firm data, we calculate average firm size, average median wage, and total payroll taken at the 

establishment level, with year fixed effects taken out. All firm characteristics refer to permanent attributes so 

that they remain unchanged across the worker panel (i.e., an individual’s firm characteristics can change only 

if she moves employers). 

To generate a uniform sample with a common definition of labor market entry, we focus on the 

effect of recessions at the end of the first exit from college and exclude workers obtaining higher degrees from 

our sample.32 As shown in Appendix Table 1 even within this relatively homogeneous sample there is a high 

rate of drop out and high variance in college attendance. Despite the use of administrative data there is likely 

to still be some measurement error in actual graduation in our data. Thus, our main sample excludes college 

dropouts to focus on a more homogenous group of workers with better measured graduation date. To do so, 

we calculate the difference (D) between actual and predicted graduation year, and keep only workers with 

 of risk aversion, absence of savings, or unevenly distributed income shocks. To our knowledge, no one has used the 

effects of actual recessions shocks or considered heterogeneity in workers’ underlying earnings capacity. 31 The word ‘college’ is somewhat a misnomer in Canada because it is used usually to refer to one or two year 

community-level post-secondary institutions rather than degree-granting universities. In keeping with the terminology 

most often used, we shall refer to Canadian universities as colleges. 32 Since we find early recessions do not affect the probability of obtaining a graduate degree, this does not affect our 

results. We have experimented with other definitions of the relevant date of labor market entry (such as last degree or 

last degree of continuous education), with little effect on the results. In the sensitivity analysis, we also show results using 

a sample that includes workers obtaining a post-graduate degree. 
17 

non-negative difference. The right columns of Appendix Table 1 show characteristics for that sample. Among 

the sample of workers on or above grade 89% graduate, and average duration of college is about 4 years.33

 To assign the unemployment rate at the time of graduation, we have to choose a relevant province of 

residence. We settled for the province of first residence as the relevant labor market for young college 

graduates.34 We impose some additional basic sample restrictions and limit the degree of missing observations 

on earnings. In particular, we drop workers who permanently stop filing taxes with the purpose of removing 

individuals who stopped being recorded annually because they left the country, obtained a new personal 

identification number, entered the underground economy, or their file was simply miscoded along the way. 

Further sample restrictions are discussed in the Data Appendix. 

Figure 1 and Appendix Table 3 describe the general experience profiles and mobility patters in our 

data, and also document the many similarities between experience profiles in Canada to those in the US. To 

summarize briefly, the first years of the careers of young male Canadian college graduates are characterized by 

steep wage growth (also documented for the U.S. by Mincer 1974, and Murphy and Welch 1990), frequent 

job changes (Topel and Ward 1992, Giuliano and von Wachter 2004), initially unstable labor force attachment 

(Ryan 2001, Gardecki and Neumark 1998), some interregional mobility (Waggoner 2004), and frequent 

industry changes (McCall 1990, Neal 1995, Parent 2000). In addition, we document a strong experience 

gradient in average size and average wages paid by employers – from year one to ten, average firm size and 

average firm wage increase by 34% and 24%, respectively (Appendix Table 3). Male Canadian graduates tend 

to move to firms that on average pay more and are larger the longer they progress through the labor market.35 

The time series of unemployment rates at the provincial level and the provincial average are shown in 

Panel A of Figure 2. Canada experienced two major recessions in the early 1980s and 1990s that increased 

young workers’ unemployment rates for certain years by more than seven percentage points. This will work 

for our national specification.36 Figure 2A also displays a high degree of regional heterogeneity. During this 

 

33 By restricting our main discussion to graduates, we are also sure of picking up the effect of early unemployment rather 

than the drop out decision. Our data suggests undergraduates are unlikely to finish early or drop out because of labor 

market conditions. Not surprisingly, the most of the results hardly differ when replicated with the full sample. 34 The alternative, province of college, gives similar results. Appendix Figure 1 compares the effect of the two choices 

for our main estimates. With choice of province of college as the relevant labor market, the unemployment rate at 

experience year zero fails to pick up some effects of the unemployment rate at experience year one that are absorbed if 

we choose province of first residence as relevant market. This leads us to believe that province of first college has a 

stronger measurement error than province of first residence. If we group 0-1 together, the results of the two choices are 

indistinguishable. An examination of the incidence and gains of province mobility below leads us to believe that selective 

mobility from province of college to province of first residence is small. 35 Using data from the Current Population Survey Appendix Table 5 shows that a similar pattern is also present in the 

U.S. (see also Figure 1). 

36 The picture shows unemployment rates for 15 to 24 year olds. Using unemployment rates defined for workers age 20 

to 24 or for college graduates only does not substantially alter the pattern of unemployment over time or across regions, 

nor does it affect our results. 
18 

period, an increase of unemployment rates of 5 percentage points (or about two standard deviations, see 

Appendix Table 1) describes a typical large recession.37 

A potential concern is whether the variation in youth unemployment rates in Figure 2A represents 

changes in demand conditions facing male college graduates. Although education-specific unemployment 

rates are too noisy for most provinces, the unemployment rate for young college educated men for the larger 

states, such as Ontario or Quebec, are closely correlated with the youth unemployment rate and the average 

unemployment rate. As noted above unemployment rates between older and younger workers in Canada 

move in parallel at different levels. This appears to hold for college graduates as well. 

4. The Persistent Effect of Initial Labor Market Conditions on Earnings 

Panel B of Figure 2 shows average earnings-experience profiles of the graduation cohorts in our data 

at the national level together with their entry wage at experience one (their first full year of work) and the 

average wage for ‘mature’ workers (workers with 5 to 10 years of experience).38 One can clearly see 

differences in starting wages across graduation cohorts leading to differences in average cohort earnings. 

Thus, as found by others, if we were to add cohort effects in a simple earnings recession, they significantly 

improve the fit of the model. However, the figure also shows a clear pattern of convergence. That is, initial 

differences in starting conditions appear to fade over time. Cohort effects appear to have a time-varying 

component, or, as noted by Beaudry and Green (2001), experience profiles vary across cohorts. 

There exists a strong correlation between starting wages and initial unemployment rate conditions, 

which persists into higher experience years and slowly fades over time. This is shown in Panel A of Figure 3, 

which graphs national unemployment rates for young workers with wages at different years of experience by 

graduation cohort (both expressed as deviations from their means across cohorts). The correlations in Panel 

A strongly suggest that part of the initial but fading earnings differences in Figure 2B are driven by variation 

in initial labor market conditions. In addition, the Panel B in Figure 3 compares the effect of aggregate 

unemployment on wages for workers with different labor market experience – the figure clearly shows how 

college graduates just entering the labor market bear most of the burden of recession adjustment, something 

we will return to below. 

The correlation at the national level is exploited in Table 1, which shows the long-term effect of 

national unemployment rates on log real earnings, controlling for year and experience effects and linear or 

quadratic cohort trends. Column (1) and (4) show the shift in experience profiles due to an unemployment 

 

37 Panel B of Appendix Table 1 shows summary statistics. Appendix Table 1, Panel C also shows the distribution of our 

sample between provinces and the mean and standard deviation of unemployment across Canadian provinces. The 

sample of students differs across provinces because of population differences and college representation. We address 

this by including initial province fixed effects (and sometimes also current province fixed effects). 38 Graduation cohorts 1976 and 1994 have lower and higher average earnings then the rest of the sample, respectively, 

due to variation in college reporting rates (Appendix Table 2A shows the respective decline and increase in sample sizes). 

In the regression models, this is accounted for by cohort fixed effects; in the figures these two cohorts are dropped. 
19 

shock in experience year zero including a linear cohort trend for all workers with some college and those in 

the graduate sample, respectively. Standard errors are clustered at the level of graduation cohort to allow for 

group level error terms. The results suggest a strong initial effect that persists but fades after about five years 

in the labor market. 

4.1. Main Regional Models 

Our main results are drawn from regional models that include cohort effects as well as effects for 

initial province of residence as described in Section 2. The shifts in experience profiles due to an initial 

provincial unemployment shock are shown in Column 3 of Table 1 for all workers with at least some college 

and Column 6 for the graduate sample. The initial effects are similar in size to those from the national model, 

but starting at experience year four, the regional estimates remain more persistent, and converge to zero only 

after 10 experience years. Although estimates for the graduates are slightly more precise, there is little 

difference in the point estimates for graduates and all workers. This is apparent in Figure 3A, that plots the 

main coefficient estimates against potential experience. It does not appear that those with a college degree 

fare better than the full sample. 

Comparing national results with the regional results, we can exclude a strong correlation of initial 

unemployment rates at the national level with changing unobserved cohort characteristics. Similarly, as shown 

below, it does not appear that the regional results are driven by more persistent unemployment shocks. 

However, national estimates may be more strongly affected by measurement error problems due to misassignment of the relevant initial labor market shock. Inter-regional mobility is less common in Canada than 

in the U.S. Thus, the relevant labor market shock is at the regional level, an effect only partially absorbed by 

the national unemployment rate. Lower regional mobility (and, as we will see below, lower gains to mobility) 

also explains why results from the national model are not larger than the regional model. 

Due to the presence of continuing exposure to adverse labor market conditions, these estimates 

represent a summary of the earnings losses the average worker can expect due to entry in a depressed labor 

market. With an increase in unemployment of 5 percentage points -- roughly a shift from boom to recession 

– annual wages are about 9 percent lower in the first year after college, still 4 percent lower after 5 years out, 

and about 2 percent lower 9 years out. Overall, we view the regional and national results as telling a 

consistent story. Graduating during a recession leads to significantly lower earnings at the beginning of an 

individual’s labor market, but the gap converges to zero within six to eight years after graduation. These 

results are consistent with estimates from the wage curve literature.39 Similarly, they are consistent with 

 

39 The results are also consistent with previous evidence on the impact of unemployment on wages in Canada. 

Blanchflower and Oswald (1994) obtain a coefficient for log annual earnings on province unemployment controlling for 

region and year fixed effects for a representative sample of Canadian workers of -.013 (Table 8.3). This is very similar to 

what we find for male college graduates, despite the fact that their regression models include a range of individual 

characteristics. They also report an elasticity of -.065 for workers with university degrees and -.169 for workers less than 

age 25 (Table 8.4). If we divide our estimated coefficients by the average unemployment rate (14%), we obtain an 
20 

estimates by Bloom and Freeman (1988) who find that initial effects due to differences in cohort sizes fade 

after ten years. Similarly, Devereux (2003) finds among a sample of workers from all ages that half of a wageshock, instrumented by local unemployment conditions, is still present after about five years. Kahn’s (2005) 

finds somewhat more persistent losses than ours, perhaps due to her focus on graduates entering the strong 

recession of the early 1980s.40

The entry into a depressed labor market appears to leave persistent but temporary scars. These 

results (Figure 4A, Table 1) are robust to a variety of sensitivity checks. First, our results do not seem to be 

driven by any particular measure of labor market conditions. To counter the concern that the unemployment 

rate for young workers may be affected by cohort characteristics, we replicated our results with the 

unemployment rate for all workers (Appendix Figure 1, Panel A). We also find similar results from using the 

employment population rates for workers age 15 to 25 (or men only). Second, we compare the effect of 

average unemployment rates in experience year zero, 0 to 1, 0 to 2, and 0 to 3 (Appendix Figure 1, Panel B).41

While high average unemployment in the early years tends to make the effects more persistent, it does not 

appear that the effects captured in the main models are driven by periods of extended unemployment. As 

confirmed by the results in Section 4.3., the driving force behind our main results is the shock in the very first 

years after entry into the labor market. 

We have tried various other sample and specification choices, none of which substantially affected 

our results. Including college students who enter the labor market after a graduate degree has no effects on 

our results (Appendix Figure 3) suggesting workers do not selectively enter advanced degree programs due to 

unemployment. We also tried various ways of excluding workers with repeatedly missing wages, and find little 

effect on our results.42 We have re-estimated all of our results using the province of college as the region for 

 elasticity of -.11 to -.13 for our younger sample of graduates. The corresponding results for the U.S. are an elasticity of -

.102 for all workers, -.064 for workers with at least 13 years of schooling, and -.192 for workers age 25 or less. 40 The size of the initial loss and the degree of persistence depends on the strength of the initial unemployment rate, as 

well as on the life-time exposure of a cohort to unemployment. Since due to the cyclical nature of unemployment, lifetime unemployment exposure is partially determined by the timing of entry, the fortunes of cohorts may differ. Below 

we try to control for some of these differences directly. To assess whether the average profiles are driven by single 

‘unlucky’ cohorts, Panel D of Figure 5 shows the same estimate for several cohort groups. While some differences across 

cohorts are apparent in the size of the initial shock and the speed of reversion and a subset of cohorts graduating in the 

early 1980s appears to experience more persistent effects, overall the patterns of initial loss and reversion are very similar 

across cohorts. 

41 Since unemployment history is available only for cohorts entering after 1981, these show more persistent effects than 

the full set of cohorts used in Table 1. We also used the unemployment rate for young men instead of that for all young 

workers, again with no difference in results. Unfortunately, alternative measures based on young college graduates only 

proved to be too noisy in several of the smaller provinces and led to attenuated results. For the larger provinces, our 

results were confirmed even for this very narrow measure. 

42 Appendix Figure 3 shows the results with those who permanently stop filing included. We also used the sample of 

workers with a valid match to their father’s income to assess the degree of selective exclusion due to non-filing. 

Although having a valid match to father’s earnings is affected by initial unemployment, perhaps through an effect on 

regional mobility, if we replicate Table 1 using only those with a valid match the results are very similar as for all 

workers. We conclude that workers with a valid father’s income match are similar to the entire sample. We then 

regressed average father’s lifetime income and “stopped filing”, with insignificant and small slopes. 
21 

the relevant initial shock with no basic change in our results. 43 Part of the reason why national results show 

more persistent effects of initial labor market conditions on wages might be that workers are ‘stuck’ in 

persistently slack regional labor markets. To address this possibility, we also included current province by 

current year fixed effects (shown in Appendix Figure 1, Panel C), which barely show any differences in the 

results. This is also a first indicator that mobility towards provinces with higher wages is not a strong source 

of catch-up in our sample. 

To isolate the effect of a short-term initial exposure to adverse labor market conditions, the large 

samples and ample cohort variation at our disposition allow us to control for the confounding effects of 

regional unemployment persistence. This is done in the first column of Table 2, which adds an interaction 

between unemployment rate in current state of residence and experience, as well as fixed effects for current 

state of residence.44 As predicted, the initial effect is reduced by persistence of labor market conditions, but 

the difference is small. Results by Beaudry and DiNardo (1991) suggest that ensuing labor market conditions 

may have persistent effects themselves for workers not changing employers. Similarly, job search predicts that 

unemployment conditions at the beginning of employment spells persist for job changers. As shown in the 

remaining columns of Table 2, and discussed in more detail in the sensitivity analysis in the appendix, the 

basic results are essentially unaffected if we allow for persistent effects of other labor market conditions as 

well. The results of the grouped model discussed in Section 3 and the Appendix are displayed graphically in 

Figure 5 (Panel A). A part of the effect of initial unemployment rates is due to persistent effects of continuing 

labor market conditions, but the bulk is driven by the very first shock alone. 

Figure 5 (Panel B) also summarizes another important step of the sensitivity analysis, the question of 

selective college graduation, which is discussed in detail in the Appendix. Summarizing, since most of our 

measures indicate insignificant effects of unemployment rates on college duration, selective timing of 

graduation does not appear to be an important phenomenon in our data. Not surprisingly, when we use the 

unemployment rate at predicted graduation as instrument our estimates confirm the main ordinary least 

squares results. Although all our results go through with the instrumental variable estimate, in what follows 

we report the more efficient ordinary least squares estimates.

4.2. Effects on Employment and Participation 

If unemployment rates affect participation, we might be concerned that our wage-estimates of the 

previous sections are estimated from a selected sample. Moreover, under a human capital model, individuals 

 

43 As shown in Appendix Figure 1, Panel C the results for earnings are marginally weaker initially but as persistent. This 

is likely due to measurement error, since in this case the shock in the province of residence at experience year one has 

very strong effects. If we group experience years zero and one together, the effects are very similar. While there may be a 

concern about selective mobility based on the unemployment shock in the province of college, we feel the effect of 

measurement error due to the mis-assignment of initial province is larger. This is supported by relatively low incidence 

and gains from regional mobility. 

44 Note that since we only observe full history of province of residence for cohorts graduating 1982 onward, Table 2 

uses only these cohorts. As further discussed in the sensitivity analysis, the effects for all cohorts are comparable. 
22 

not working lose time during which they acquire skills and knowledge that make them more productive. 

Thus, if entering the labor market in a recession reduces time worked, this might be a channel through which 

earnings are affected even in the medium run. Table 3 replicates the same results as in Table 1 using the 

fraction of workers claiming unemployment insurance benefits, the fraction of workers filing taxes with zero 

earnings, and the fraction of workers not filing taxes in a given year. The table shows an initially significant 

increase in fraction zero earnings and the fraction of unemployment insurance (UI) claimants that fades 

within three experience years.45 The effects are numerically small and can be completely explained by shortterm persistence of local unemployment rates (not shown). A temporary unemployment rate shock has no 

persistent effects on employment or participation of male college graduates.46

To gauge the magnitude of these numbers, add the fraction workers reporting zero earnings and the 

fraction UI claimants for the group on or above grade. In the first year in the labor market, a five percentage 

point increase in unemployment would induce an increase in this measure of ‘time not spent working’ of 

about 1 percentage point. If the typical length of a spell of non-employment were 6 months, then the 

expected amount of time lost would be about 2 days or 0.06 months (0.01 times 3). These back of the 

envelope calculations suggest that the loss of experience due to labor market entry in recessions is not very 

large for the average college graduate. These results are echoed by Kahn (2005), who finds small initial effects 

on hours, employment, and weeks worked for male college graduates in the U.S. after the 1982 recession. 

Since our sample does not contain information on time worked, we also replicated our results with 

the Canadian Census (available in a Web Appendix).47 Decomposing the effect of early unemployment rates 

on annual earnings into the effect on weeks worked and on weekly wages we find that the effect on weeks 

worked is short lived. The majority of the persistent effects we find is driven by a reduction in weekly 

earnings. Consistent with the small effects on employment we find our results change little if we restrict our 

sample to workers with positive earnings in each year (see Figure 5, Panel C). Thus, neither changes in labor 

market experience nor selective entry or exit from the earnings sample of workers of different abilities affect 

the main pattern of reversion we see.48

 

45 The fraction of missing earnings (those not filing) rises only in the year of college exit, during which most graduates 

only work part of the time. 

46 The effects are very similar for the sample of all workers. Table 3 also displays a pattern of ‘overshooting’ after 

experience year 7 some measures that would imply workers who had initially higher instability become more stable later 

relative to their more lucky counterparts. One could think of various hypothetical explanations of such a phenomenon. 

However, the estimates are numerically very small and never above 0.2 percentage points. 47 Similarly, although male college graduates are indeed a group of workers with very high labor force attachment, the 

cross-sectional experience gradient of weeks worked from Census data reveals that in the first three to four years of 

there is still some heterogeneity in work patterns that stabilizes afterwards. To further examine the importance of a loss 

in time worked, we replicated our results using four years from the Census (1981, 1986, 1991, and 1996) and 

decomposed the main effect on annual earnings into effects on weekly earnings and weeks worked. Although due to the 

different nature of the data we have to make some assumptions on the timing and province of college graduation, the 

main effects are very similar to our results. 48 This is corroborated by the fact that those who permanently stop filing do not appear to be any different from those 

who remain active (Panel A of Appendix Figure 3). The estimates based on the balanced panel in Figure 5 (Panel C) are 
23 

4.3. Mobility across Firms, Industries, and Regions 

Job search is a common explanation for both high wage growth and high job mobility in young 

workers’ careers (e.g., Topel and Ward 1992). Several studies aim at testing the basic elements of job search 

theory, such as the effect of past wages, tenure, and experience on the probability of job change (e.g., Topel 

and Ward 1992, Manning 2002, Farber 1994). While most of these studies try to control for unobserved 

heterogeneity, few exploit external sources of variation to identify the effects of interest. In this section we 

estimate the direct effect of early labor market conditions on the annual propensity of job change. 

Figure 4 (Panel B) graphs the effects of initial unemployment rates on the probability of changing 

employers. Initially, a 5 percentage point increase in regional unemployment rate raises the rate of job change 

by about 1.5 percentage points (depending on the sample); the effect gradually declines with experience until 

it is not significantly different from the baseline in experience year 8 to 9. Table 4 shows the coefficient 

estimates of the effect of unemployment rates at time of college exit for various measures of job mobility 

during the first ten experience years for the regional model estimated for the graduate sample (the results for 

all workers are very similar and available on our website). The increase in mobility is significantly larger but 

shorter lived at the national level.49 Overall, the results in Table 4 yield a persistent increase in mobility 

relative to the baseline of about 4-5% (see Appendix Table 3A). 

To gauge the magnitude of these effects, consider the reductions in job change with labor market 

experience apparent in Figure 1. Between experience years 2 and 4, the rate of job change for graduates 

declines by 3 percentage points annually. If this increasing stability reflects improving job matches due to 

search, a 2 percentage point increase in job mobility is comparable to holding workers back 3 to 4 months in 

their job search efforts. A similar pattern holds in experience years 5 and 6, where overall mobility declines 2 

percentage points, such that a 1 percentage point increase in mobility compares to a loss in job search of 

about 4-6 months. Thus, entering the labor market in a recession implies that workers lose about 4 months of 

search effort annually due to a bad initial start.50

Table 4 also reports effects on the cumulative fraction of workers who left the initial employer, as 

well as the propensity of change among 2-digit industry classes. While the effects on the rate of job change in 

columns (1) and (5) decline, the cumulated rate of departure from the 1st employer in columns (2) and (6) 

increases permanently by about .3 to .5 percentage points. Interestingly, the effect of an early unemployment 

 by 0.002 smaller in absolute value than our main estimates, a difference that is not statistically significant. Note that, if at 

all, the figure suggests negative initial selection, possibly consistent with a certain degree of out-migration to the U.S. of 

high earners. This is consistent with small decline in average predicted earnings with experience in our sample. 49 The national results are in the Web Appendix. Including current unemployment rates interacted with experience to 

control for persistent local labor market conditions leads to a slight increase in the effect on job mobility (since for all but 

labor market entrants higher unemployment depresses voluntary mobility). This is another sign that the effects on 

mobility we obtain are stemming for voluntary mobility – in case of involuntary mobility, would have expected inclusion 

of additional unemployment controls to lower the estimates. 50 The initial increase in job mobility we find is of comparable size as the effect of a 10% reduction in wages found by 

Topel and Ward (1992), consistent with the magnitude of wage losses we find. 
24 

rate on the frequency of 2-digit industry changes is almost as high as the frequency of changes in employers. 

Even mobility across 1-digit industries is relatively high (the coefficients are about three quarters the size of 

the coefficients for the 2-digit case in Table 4). This may either mean that the distinction among industries is 

meaningless in our sample, or that in addition to job shopping workers also actively search for a match with 

the ‘right’ industry. Several models of job search of younger workers would predict such a pattern (e.g., Neal 

1995 or McCall 1990). 

 To what extent does the increased job and industry mobility contribute to the reversion of earnings 

losses? Clearly, the initial increase and gradual fading of mobility-responses with experience follow similar 

patterns as the change in the experience-earnings gradient. Mobility is likely endogenous itself, and thus we 

cannot ‘condition out’ the contribution of mobility on earnings effects of early unemployment rates. To gauge 

the potential of job and industry mobility to explain the observed earnings pattern, the upper panel of Table 5 

shows the average earnings gain at job and industry changes by experience. Columns 1 through 5 show 

percentage annual earnings increases for movers and stayers, as well as for the full sample. 

The purpose of this descriptive table is to characterize the association of mobility and wage growth 

without any causal interpretation. Similar to Topel and Ward (1992)’s results, the table documents a strong 

correlation between job changes and wage growth. On average, wage changes at job changes account for 

about 40% of overall wage growth in the first five experience years, and thereafter steadily declines to reach 

about 20% in experience year 10. Despite the differences in samples (their sample included workers of all 

education levels), these fractions are remarkably similar from what results in Topel and Ward (1992) and 

Giuliano and von Wachter (2005). 

Earnings growth at job and industry mobility is 24% on average, and about double the growth for 

stayers from experience years 2 to 5, and then 1.5 times thereafter.51 If one took this as a typical gain 

associated with a job change, then the estimated 1.5 point increase in job changes due to a 5 point recession 

shock could explain about 20-25% of the reversion of initial losses. (Table 5B implies that an average increase 

in the rates of earnings growth for the first experience years due to 5 point initial UR shock is about 1.5-2 

points.) Thus, job and industry mobility have the potential to explain an important fraction of the decay of 

initial job losses. However, the actual effect is likely to be larger since in a search framework the gains for 

workers starting at lower wages are likely to exceed those of the average. 

To take this into account, the second panel of Table 5 presents models of the effect of initial 

unemployment rates on the rate of earnings growth by mover status. Due to selection into mover status, we 

cannot obtain a causal effect for wage growth of movers and stayers, neither is there a simple decomposition 

of the effect on total wage growth into the effects on its components. Instead, to complement the results in 

 

51 Experience year one includes transitions from jobs with half a year to jobs with a full year of earnings and thus is 

overstated. Note that these gains are higher than those found by Topel and Ward (1992) (Table VII), but they look at all 

workers and at quarterly earnings data. 
25 

Panel A, the goal of Panel B is to assess whether the correlation of earnings growth and job mobility 

strengthens for workers entering the labor market in a recession. Column (1) shows that the effect on 

changes in earnings for the full sample is of similar magnitude as the corresponding level estimates in Table 

1.52

 Columns 2 and 3 show that the correlation between earnings growth and job mobility rises in 

recessions. This implies that the average earnings gains shown in Panel A are likely to understate the true 

gains of those moving jobs in response to a recession. Job movers have persistently higher wage gains than 

stayers in response to an initial unemployment shock, that is, job movers catch up faster from the initial loss. 

Columns 4 and 5 suggest that earnings gains at moves across industries are less precisely estimated, but follow 

a similar pattern. Table 6 also shows estimated gains from regional mobility. As discussed in the appendix, 

while the regional mobility appears conducive to wage growth, most of reversion of the losses from initial 

labor market shocks appears to take place within regions. 

4.4. Careers Between Firms 

The experience profiles in firm size and firm wages shown in Figure 1 suggests workers search for 

better employers over time. This is consistent with a growing literature documenting large difference in firms’ 

wages not explained by worker and firm characteristics (e.g., Abowd, Creecy, Kramarz 2002, Idson and Oi 

1999). A similar gradient arises if high wage firms gradually screen for more able workers among labor market 

entrants, either because of comparative advantage (Gibbons, Katz, Lemieux, Parent 2005) or because they 

thereby minimize the rents they pay (Lemieux 1998).53 Adverse labor market conditions may impact these 

processes and reduce the quality of firms at which workers start their careers. Bils and McLaughlin (2001) 

find that better paying industries have pro-cyclical hiring patterns.54 Similar pattern are appear to hold for 

better paying or large firms; for example, this may arise due to changes in demand for products of different 

quality, differences in the costs of job creation, or because of changes in product market competition. In 

addition, it has long been speculated that firms raise their hiring standards in recessions (e.g., Hall 1974, Solon 

et al. 1997). This would lead to a temporary cyclical downgrading as workers tend to start at low wage firms. 

Figure 4 (Panel C) and Table 4 show some evidence of this process with the estimated effects of 

early labor market conditions on the average firm sizes and average firm wages by experience year. Table 4 

again shows results using regional unemployment shocks and for two samples. A reduction occurs in initial 

 

52 The effects based on changes are slightly more persistent, partially due to a slight difference in samples as well as due 

to the implicit control for worker fixed effects in the wage growth model. 

53 This process is reinforced if human capital increases with experience. Fox (2004) suggested that large firms will try to 

attract older, more experience workers because of span of control considerations. Or if, as in Neal (1998), high ability 

workers are better at acquiring specific human capital, and large high-wage firms value human capital more, over time 

more able workers will again transit to high wage firms. Alternatively, workers may start at low paying firms that allow 

for more general human capital investment on the job and then switch employers (Rosen 1972, Mincer 1974). 54 Typical high wage and pro-cyclical industries are durable goods manufacturing and construction. Typical low wage, 

less pro-cyclical sectors are retail trade or personal services. 
26 

firm size that fades within four years; for the graduate sample in Column 5, a 5 percentage point recession 

reduces firm size by 4-5% in the first years. As the remaining columns and Figure 4 (Panel C) shows, there 

are much larger effects on the average median (log) wage of a worker’s first employers. Column 7 suggests 

that the average median log wages of a workers’ employer falls 3-5% in the first years after entry into a 5 

point recession. This effect declines to a 2% reduction in years 5 to 9, and only fades by year 10. Since the 

effect of average log payroll combines the effects on average size and average median wages, the effects are 

initially larger (7-10%) than those on median earnings but decline more rapidly over time. 

These numbers suggest that about 40% to 50% of the effect of an initial 5 point unemployment 

shock on wages shown in Table 1 (ranging from -.092 initially, to -.059 after 3 years and -.0365 after 7 years 

for the graduate sample) could be explained by reductions in the average wage of an employer. To gain 

further insight about the economic significance of these results, compare the effects of early recessions on 

average median firm wages with the experience profiles in firm ‘quality’ in Figure 1. The increase in average 

median firm wages due to experience is 8%, 6%, 4%, 4%, and 2% from year zero to year five (in the graduate 

sample).55 If workers search continuously throughout the year, and job search entails a continuous increase in 

firm size, then the effects of recessions set people back by about half a year in their job search process 

consistently in each of the first five years in the job market.56

It appears that a considerable part of earnings losses from graduating in a recession can be explained 

by the start of working life in lower paying industries and firms. Over time, affected workers improve their 

relative position vis-à-vis other more lucky workers by switching to better paying establishments. These 

moves entail switches across industries and across regions as well, but little losses in the time spent working. 

Thus, firms appear to play an important role in the determination of early wage growth and in the persistence 

of early labor market shocks on wages. This is consistent with a pattern of cyclical down- and upgrading of 

workers between industries and firm-types (e.g., Okun 1973). However, workers do not appear to be 

confined to their initial employer and can remedy an initial bad draw due to temporary changes in hiring 

standard in a recession by switching employers as the economy turns back to normal. 

4.5. The Role of the First Employer 

To further describe the career moves that lead to the catch-up up pattern observed in Figure 4 (Panel 

A), we assess the role of the initial employer in ensuing career outcomes. The results in the previous section 

in particular suggest that workers might lose initially due to access to firms that pay less and perhaps offers a 

less favorable career environment. To do so, we first include fixed effects for the first employer in the basic 

 

55 Relative to the increase in average firm size (-4%, 7%, 4.4%, 1.3%, and 4% in years 1 to 5 for the graduate sample), the 

effect of initial firm size sets workers back by about a little more than half a year. 

56 Similar results are also obtained for average one, two, and three-digit industry wage premiums, consistent with the fact 

that high wage industries have more pro-cyclical employment creation. However, changes in average industries wage 

premiums for labor market entrants can only partially explain decline in average firm wages (see the Web Appendix for 

direct estimates of changes of average industry wages). 
27 

regression model, and also interact these fixed effects with unrestricted experience profiles. An interaction 

with experience also allows for shifts in experience profiles related to the choice of the first employer to 

account for heterogeneity in training or learning opportunities across firms. If all that happens is that workers 

search for rents, controlling for initial firm fixed effects should eliminate losses. Clearly, these results can only 

be indicative, since recessions are likely to change the ability of the marginal worker hired. 

The results, shown in Figure 6A, suggest that wage losses for workers starting in a recession at a 

given firm relative to workers starting at better times are about 60-70% of original losses initially, and then 

decline to about 60% and fade completely after year seven. Thus, 30-40% of losses are driven by the choice 

of the very first employer alone (see also Appendix Table 6). Including an interaction with initial firm and 

experience effects does further reduce the initial earnings loss, but not by much. Thus, it does not appear that 

on average choice of initial employer is associated with different growth paths. Effects of first industry at the 

2-digit level also explain a smaller but important part of the initial wage differences. 

These results confirm that average wage differences between firms and industries play an important 

role in explaining persistent recession effects. Yet, there are important persistent effects left for workers 

starting at the same firm. This is not surprising since previous research have found firms do not fully shelter 

workers from external labor market conditions, specially for job changers (Solon et al. 1994, Devereux 2003). 

In addition, the empirical and theoretical literature suggests firms may have pro-cyclical internal job 

assignment (e.g., Okun 1973, Devereux 2005, Gibbons and Waldman 2004) or write contracts based on 

external market conditions (e.g., Beaudry and DiNardo 1991), potentially leading to persistent earnings effects 

within firms. These mechanisms should be particularly relevant for larger and high paying firms since despite 

reduced earnings workers may not have an incentive to change employers. 

To gauge whether workers starting at large or high wage firms have more persistent wage losses due 

to unemployment rate shocks, we interacted a dummy for whether the first employer is large (firm size 

greater than 1000) or pays high wages (log median earnings greater than 75th percentile of distribution) with 

unemployment at time of graduation. We find that there’s a substantial gain from starting at larger or high 

paying firms that is unaffected by early labor market conditions (Figure 6B). That is, relative to workers in the 

same firm type, the initial wage losses from starting in a recession are similar. However, the losses of workers 

starting at large, high wage firms are more persistent (i.e., the difference in losses between initial firm types turns 

significantly negative with experience). 

Workers starting at large, high paying firms always benefit from a wage premium (about 15-20% for 

firms with 1000 or more workers, as in Oi and Idson (1999), and about 30-40% for firms above the upper 

75th percentile of average log median earnings), but may be stuck with a worse contract, consistent with 

findings in Baker, Gibbs, and Holmstrom (1994). Workers starting at smaller, low wage firms are more likely 

to move and improve their position in terms of contract or firm type, consistent with the pattern documented 
28 

in Section 4. Although none of the controls for selection bias we tried affected our results, it should be kept 

in mind that these results may partially be affected by differential hiring patterns over the business cycle.57

5. Heterogeneity of Early Recession Effects 

5.1. Larger Effects for Labor Market Entrants 

To put the magnitude of the shock in the initial period into further perspective, Figure 5 (Panel A) 

shows the dynamic effect of a shock occurring at experience years two to three from the grouped model with 

full history controls (Table 2, Column 7). To make the dynamic pattern comparable with that of the first 

group, the table shows coefficients relative to the time of the shock (i.e., experience zero now relates to the 

moment of the shock). The effect of a shock experienced at experience years 2-3 is much smaller than the 

effect of a shock at entry (0-1) for all experience years. Our period is too short to observe complete reversion 

but the point estimates are insignificant after 4-6 years. Inspection of the data leads us to believe that the 

dynamic effects for shocks at later experience years are small.58

To explore the difference between labor market entrants and more mature workers further, Table 6 

analyzes the profile in the effect of unemployment rates on wages and other outcomes by five experience 

groups. To make our estimates comparable with the previous literature, we show effects of the natural 

logarithm of unemployment rates controlling for current province fixed effects. The upper panel uses the 

unemployment rate for workers age 15 to 24 and the lower panel considers the effect of unemployment rates 

for all workers. The first rows of Panel A and B show the effect of unemployment without experience 

interaction. The elasticities in the first row of Column 1 of the two panels essentially replicate the results 

typically found in wage-curve estimates. The remaining columns show the effects of unemployment on other 

outcomes; the remaining rows of the table show separate estimates by experience groups. 

The table makes strongly confirms the exceptional role of labor market entrants vis-à-vis mature 

workers. First, in all estimates there is an important experience gradient in the effect of current 

unemployment rates. Thus, the pooled estimates in the first row potentially obscure important effects present 

in the data. Second, the initial effects in early experience years are the strongest across all groups. 

Unemployment conditions in the local labor market matter three to four times as much for labor market 

entrants than for young workers who already progressed into their career by a few years. Third, the estimated 

gradient is as expected from results of the previous literature. For example, job to job mobility of mature 

workers declines in recessions (Shimer 2005b), effects on non-employment are small, and average firm size 

rises for mature workers since smaller plants are more likely to close (Krashinsky 2002). Note that since later 

experience years pick up some of the persistent effect of the initial shock, the difference between the effect of 

 

57 We included several control functions based on the quadratic average father’s income at the cell level and cell size, 

none of which turned out to be significant or to alter our results. 58 Our sample of cohorts is small at later experience years, such that the cohort variation shown in Appendix Figure 1 

does not allow us to estimate the average dynamic effects of shocks at later experience years. 
29 

unemployment at experience years 0-1 and 2-3 or later years is understated. A replication of the table with full 

dynamic controls yields qualitatively similar results but larger initial differences. 

5.2. Larger Effects for Entrants at the Bottom of the Skill and Earnings Distribution 

The baseline results reveal that, on average, individuals that graduate in high unemployment 

conditions experience lower relative wages and that this gap dissipates over a 10 year period. Consequences 

of graduating in poor economic conditions, however, might differ by innate ability or initial background 

characteristics. The most productive workers may possess superior skills or network connections that assist 

them to find higher paying jobs. Similarly, firms may have an incentive to hire more able workers if they pay 

wage premiums. While there are some estimates of recession effects for the very bottom of the skill 

distribution, less evidence exists for the importance of heterogeneity among higher-skilled workers. 

Moreover, very little is known about the degree of heterogeneity in the persistence of initial effects, or how 

different groups of workers catch-up after an initial shock. 

These differences are of special interest for our purposes, because they provide an additional source 

of variation to assess the economic mechanisms underlying the adjustment process. Yet, they have been 

understudied because few data sources contain information on both employer quality and workers’ earnings 

capacities. We use our data to examine whether college graduates with lower predicted wages, based on 

college background, are more adversely affected by higher unemployment rate conditions in terms of lower 

earnings and ending up at firms that tend to pay less. We first use an OLS model to predict log earnings 

based on college attended, program of graduation, and years of study, conditional on province of study and 

cohort year.59 Since individuals are likely to be sorted into colleges, these estimates will capture both 

differences in innate ability as well as differences in college quality.60 We then group individuals into quintiles 

based on these predicted wages. 

Figure 7 shows the same coefficients for the effects of the initial unemployment rate on log earnings, 

job mobility, individual’s firm’s log median earnings, and employment as in the baseline model, but for the 

first, third, and fifth predicted wage quintiles (this figure corresponds to Figure 4 for the full sample). Table 7 

summarizes the key structure of losses by quintile and compares them to results for the full sample. For the 

sake of exposition, the table displaces three parameter estimates for the initial dip, first recovery, and final 

 

59 A similar approach to assess college quality is followed by Betts, Ferrall, and Finnie (2000), who use the same college 

data and information on wages after graduation as we do. After analyzing majors and colleges separately, in our final 

specification we interact major and college dummies. The results by major are show in Appendix Figure 4 for earnings, 

the analysis for colleges and other outcomes is in the Web Appendix. Differences by major or college in itself are as 

expected (e.g., humanities graduates do worst, then come social sciences, economists and engineers are in the middle 

range), but too broad to yield a prediction of individual earnings capacity. This exercise is done for the graduate sample 

only, since it is conceptually harder to assign college quality for drop-outs. 60 This is discussed extensively in Black and Smith (2003), Black, Kermit, and Smith (2005), or Dale and Krueger (2002). 

An advantage of our data relative to the literature on college quality in the U.S. is that we have access to earnings 

histories. Using the same data as ours, Betts et al. (2000) find that the effects of observable measures of college quality 

on earnings are small. 
30 

fade of earnings losses in an approach mirroring that of Jacobson, Lalonde, and Sullivan (1993). As apparent 

from the table and figure, those with the lowest predicted annual earnings are most affected by higher initial 

unemployment conditions and experience permanent earnings losses. Earnings one year into the labor 

market are about 15 percent lower from a 5 percentage point increase in the initial unemployment rate, and, 

in this case, remain about 7.5 percent lower even after 10 years. The top quintile’s earnings are about 7.5 

percent lower, on average, in the first year out, but the gap falls to less than 2 percent after only 4 years. 

The pattern is altered if we consider the effects on firm quality, as measured by the median wage paid 

at the firm. While those in the highest quintile end up in lower paying firms only for the first 4 to 5 years, 

firm quality is permanently worse among the lowest quintile. Different from before, the middle group sees 

quick initial improvements but lasting declines in firm quality (Panel C); nevertheless, earnings continue to 

rise and job mobility remains high (Panel B). Thus, while initial catch up is driven by improvements in firm 

size, continuing catch-up appears to be driven by beneficial job mobility within firm-classes. 

These patterns are summarized in Figure 8 (Panel A). By deciles of predicted earnings, the figure 

shows the fraction of earnings losses that have faded after five years in the labor market, as well as the 

improvements in firm quality and the fraction of workers that left their first employer. Those deciles with 

highest rate of job mobility and larges changes in firm quality appear to have faster reversion of earnings. The 

correlations in the figure lend additional support to the result based on the average in our sample that 

increasing job mobility and improvements in firm quality are important channels of recovery from an initial 

recession shock.61

The longitudinal data also allows us to obtain a direct measure of the cost of recessions that is a 

useful complement to measures in the literature based on the standard deviations of earnings. Figure 8 (Panel 

B) graphs two summary measures of the present discounted loss due to entry into the labor market in a 

recession by deciles of the predicted earnings distribution. First, it plots the percentage decline in the present 

discounted value of annual earnings; second, it shows the fraction increase in annual earnings a worker would 

require to be indifferent between the noisy earnings path and an alternative, stable path. The latter 

corresponds conceptually to the original Lucas measure where we have replaced consumption by annual 

earnings and is comparable to several estimates of costs of recessions in the literature.62 

Figure 8B has two key messages. First, there is an important gradient in the cost of recessions in 

predicted earnings – those individuals with lower earnings capacity have four to five times costs of recessions 

than the most advantaged workers. The least advantaged appear to bear most of the costs of recessions. 

 

61 As Figure 7 (Panel D) shows, the lowest ability workers are an exception and tend to converge by improving labor 

force attachment relative to similar workers graduating in booms. 62 This approximation has clearly important flaws, since social insurance programs smooth temporary earnings shocks 

and may lead consumption to be less volatile than earnings. On the other hand, this might be less of a concern for 

highly educated workers whose take up of social programs is low. Here we follow the literature on the costs of 

recessions by approximating the risk faced by individuals with earnings risk. Cutler and Katz (1991) discuss the 

usefulness of earnings as a measure of inequality and the effects cyclical shocks. 
31 

Second, the losses from starting to work in a recession as measured by actual changes in the present 

discounted values of earnings or utility losses are high even for the more able workers. In particular, they are 

much higher for the median worker in our sample than what is typically found in the literature.63

6. Discussion of Career Models 

The preceding sections document several facts that a model of career determination should be able 

to explain. Although no single model is likely to explain all patterns or is completely rejected by the data, for 

most workers we find little evidence for either permanent scarring or short-term frictionless adjustment. 

Instead, a combination of models based on search and contracting frictions hold could explain the main 

pattern we find. 

First, the persistence of a short-term initial labor market shock appears too high to be explained by 

the standard neo-classical model, especially once we control for persistence of unemployment shocks. 

Similarly, the standard model cannot explain the systematic responses of job mobility and firm quality, the 

fact that less advantaged college graduates are hurt more, and that labor market entrants fare significantly 

worse with young workers with two or three years of labor market experience. While each of these 

phenomena could be explained by an extension of the standard model, the balance of the evidence suggests 

important short- to medium run departures from a frictionless view of wage setting. 

This conclusion is corroborated by the fact that the initial losses in employment we find appear too 

small to cause significant losses in general human capital accumulation. Analyzing survey information on 

weeks and hours worked, Kahn (2005) comes to a similar conclusion. Labor force attachment of male college 

graduates is very high even in recessions. Moreover, since our results persist even controlling for subsequent 

labor market shocks, it appears unlikely that continuing low economic activity within firms reduces the 

amount of learning-by-doing.64

A second important insight from our results is that the importance of job changes in early wage 

growth and the increase in job mobility for workers graduating in recessions suggests that initial losses occur 

due to a temporary worsening of the wage-offer distribution and a time-intensive search for a better job after 

the distribution has shifted back to normal. Topel and Ward (1992) and von Wachter and Bender (2006) 

suggest that the job shopping process can last 5 to 10 years. The heightened mobility across industries also 

suggests that workers actively modify their careers in response to an initial shock to avoid persistent losses. 

 

63 We find that an uncertain stream of earnings had to be increased by about 7% for the median worker in our sample to 

be of equal utility as a comparable certain path. The typical estimate in the literature is below 1%. Some studies, such as 

Storesletten, Telmer, and Yaron (2001) or Krusel and Smith (1999) find effects comparable to ours for households with 

no wealth. 

64 We also find that a substantial fraction of improvements in wages occur at job changes, and that this correlation 

strengthens for workers graduating in a recession. This result and the persistence of losses from just a single initial shock 

also makes it appear unlikely that initial losses arise from lack of accumulation of firm or industry specific capital due to 

excess initial mobility. While Kahn (2005) finds no effect of recessions on job tenure, Devereux (2002) finds significant 

yet small effects. Due to measurement error in self-reported job mobility, to small sample sizes, and to the presence of 

cohort effects, the authors may obtain estimates that are likely to be imprecise and downward biased. 
32 

This is consistent with Neal’s (1999) concept of ‘complex’ job mobility according to which workers search 

both over careers and jobs. Since we observe that most job and career switches occur early, losses from 

career-specific capital are likely to be small. 

Third, the positive experience gradient we find in firm wages and firm size suggests workers search 

for better firms over time. Building on a large literature documenting important differences in average firm 

wages (e.g., Dickens and Katz 1988, Krueger and Summers 1988, Abowd et al. 2002), our paper is the first to 

characterize the experience gradient in firm quality in detail, extending and complementing earlier analyses by 

Fox (2004).65 Consistent with a large literature documenting that high-wage sectors have more pro-cyclical 

job creation, we find that an important part of initial losses is due to a down-ranking of workers into smaller, 

less-paying firms. Graduates lose initially because they face a worsening distribution of firm quality in the 

market during a recession, a channel often invoked in the literature but not documented before in this detail 

(e.g., Cutler and Katz 1991, Hines, Hoynes, and Krueger 2002). 

Consistent with these observations, we also find that catch-up from an early recession shock involves 

a systematic recovery in firm quality, a pattern not due to persistent labor market conditions. These results 

point towards an environment where young workers search for jobs in high paying firms and industries, and 

where wage losses and an important part of the ensuing convergence occurs through improvement in firm 

and industry characteristics. Such a process of temporary cyclical down-grading could either result from an 

environment in which certain sectors pay rents (Okun 1973, Krause and Lubik 2005), or from a situation in 

which workers of heterogeneous ability face firms with different payment structures (Mclaughlin and Bils 

2001, Gibbons et al. 2005).66

Fourth, our results also suggest an important role for the presence of long-term contracts and careers 

within firms. This can explain the fact that labor market entrants suffer stronger earnings losses than workers 

who are already employed. It also suggests that contracts can explain an important part of the remaining 

persistence we find within firms. However, since the effects we find arise just from a single recession shock 

and are robust for detailed controls for unemployment history, in our sample initial shocks keep influencing 

contracts despite the advent of better labor market conditions (Beaudry and DiNardo 1991). The ‘contract’ 

view could also explain our finding that workers starting in large high wage firms have more persistent 

earnings losses from graduating in recessions. Since on average these workers benefit from significant wage 

premiums, this may explain the lack of mobility that must underlie persistent effects within firms. This 

 

65 Using three size categories in the Survey of Income and Program Participation Fox (2004) analyzes whether larger 

firms search for better (older) workers to fill higher-ranked positions. Similarly, high-wage firms may try to hire older 

workers to minimize rents paid. While these explanations are complementary to our story, they are unlikely to explain 

the whole experience gradient, since workers are often promoted internally into higher positions, something not 

captured by the experience profiles. 

66 While we cannot rule out an explanation based on sorting and employer learning, it is not clear why information losses 

in recessions should down-rank the average worker; similarly, without additional assumptions a model of sorting cannot 

explain positive gradients in firm characteristics. 
33 

suggests that the pattern of permanent firm-entry cohort effects found by Baker, Gibbs, and Holmstrom 

(1994) are concentrated at very large, high paying firms. 

A fifth insight from our results is that there is important heterogeneity in the long-term costs of 

recessions, even among college graduates.67 These differences suggest that firms raise their hiring standards in 

recessions and that wages deviate from marginal products.68 Our results also imply that the pattern of catchup differs among high and low wage workers. While the majority of the population catch-up by switching to 

better employers, in particularly in the first years after labor market entry, the bottom 20% is permanently 

down ranked to lower quality firms (as predicted by Okun 1973). These workers still see some reversion of 

initial losses, partially due to an increase in labor supply. Since low-wage occupations and industries are 

typically less pro-cyclical, these differences are unlikely to be driven by skill-specific demand shocks. 

7. Conclusion 

We have estimated the long term effects of entering the labor market in a recession for a large sample 

of Canadian men leaving college whose earnings, employers, and career outcomes are tracked for ten years. 

Our main results suggest that the average worker graduating college in a recession faces earnings losses that 

are very persistent but not permanent. On average, a two standard deviation increase in the unemployment 

rate (roughly comparing the difference between those exiting college in a bust versus boom) leads to an initial 

wage gap of about 10 percent. This gap declines relatively slowly, and fades to zero after about the eighth 

year. Controlling for unemployment rate conditions after the first year of labor market entry, we also 

conclude that virtually all of the wage deficit can be attributed to the unemployment rate variation in the very 

first year after leaving school. The results are robust to selective graduation or selective labor force 

participation, and to the many alternative specifications we tried. 

We have also analyzed the possible sources of these losses, and how workers that graduate in a 

recession eventually catch-up to their counterparts that graduate in a boom. Job and industry mobility rise 

initially and decline gradually in response to an initial adverse shock, implying that exiting college in a 

recession reduces the quality of initial job matches. However, job moves are highly productive for the young 

 

67 Changes in the decomposition of employers over the business cycle are also often observed in the academic labor 

market for economists. In particular, while the type of jobs available across the cycle varies little for the top graduate 

students, it is matters a lot for the median student or the lower tail. The analogy to the wider labor market is imperfect, 

however, since job mobility and the access to alternative employers for the average economist in the first years of the 

career is much lower. Not surprisingly, as found by Oyer (2005), cyclical swings in the market for graduates from top 

Ph.D. programs lead to persistent declines in productivity. 68 Any model in which wages contain rents that decline with worker ability implies high-wage firms selectively hire more 

able workers and reduce employment of less able workers first in a downturn (e.g., Lemieux 1998). A cyclical adjustment 

of hiring standards has been often noted (e.g., Reder 1955, Hall 1974), and it has often been suggested that a loss in job 

quality is an important feature of recessions for less skilled workers (e.g., Okun 1973, Hines et al. 2002). It is more 

difficult to explain this phenomenon in equilibrium models of wage determination in which wages always equal marginal 

product. While for employed workers complementarities between specific skill accumulation and general ability may 

reduce the job stability of less able workers (Neal 1998), this is less relevant for labor market entrants. Counter-cyclical 

demand for skilled labor could explain the patterns observed in the data. 
34 

graduates in our sample, and this association strengthens for workers affected by adverse early market 

conditions. Calibrating the effects of job search using estimates of average wage gains of moving by 

experience, we find that increased job search can explain about 30% of the reversion in initial wage losses. 

Consistent with a pattern of increased productive job mobility, we find that recessions lead workers 

to start at employers that are on average smaller and pay less. Possibly due to a lack of information on firms, 

few papers document that workers seem to face positive experience profiles in firm-size and average firmwages. We document this pattern, compare it to similar data for the US, and show that declines in the size 

and average wages of first employers of young college graduates could explain about 30 to 40% of initial wage 

losses from starting a career in a recession. Labor market entry in bad times leads to worse job placement or 

mis-matches of workers into firms, and workers catch-up by searching for or sorting themselves into better 

establishments. Such a process of cyclical up- or down-grading could either result from an environment in 

which certain sectors pay rents (e.g., Okun 1973), or from a situation in which workers of heterogeneous 

ability face firms with different payment structures (e.g., Gibbons, Katz, Lemieux, and Parent 2005). As 

suggested by Freeman (1975) and Katz et al. (1995), these patterns are strongest for young workers, while 

older workers are less affected by recession shocks. 

We also find that college graduates at the bottom of the wage and ability distribution have larger and 

more persistent losses, while the effects at the top are small and short lived. Our estimates of the path of 

earnings declines suggests that the present discounted value of losses in annual earning or utility could be 

three to four times larger for the least relative to the most advantaged workers, indicating that even within the 

group of college graduates there is a large degree of heterogeneity in the costs of recessions. Patterns across 

deciles of the distribution of predicted wages support the importance of job mobility and changes in firm 

quality, with exception of those least advantaged, who suffer permanent earnings losses and are permanently 

down-ranked to lower wage firms. 
35 

Data Appendix

Our data combines three administrative datasets from Statistics Canada. The first is the University 

Student Information System (USIS), which includes enrollment and graduate information of post-secondary 

students in Canada from 1974 to 1997. We augment the USIS data by linking it to income data from the T1 

Family File (T1FF) between 1982 and 1999, and to an employer-employee matched dataset called the 

Longitudinal Employment Analysis Program database (LEAP). Each is described below, followed by how we 

defined the variables used in our analysis. 

USIS is a national database containing pertinent up-to-date information on student participation in 

and graduation from Canadian degree granting institutions obtained from administrative records provided at 

the individual level. USIS has two main components. The enrolment survey collects information on student 

counts, and requests information on a broad array of student and program characteristics including 

institution, province, gender, age, mother tongue, immigration status, country of citizenship and country of 

origin, full- or part-time status, type of qualification sought (e.g., bachelor, masters, etc., or none), field of 

study, year of study in program and an individual identifier. The degrees survey collects information on all 

students who have received a degree, diploma or certificate during the calendar year. The degrees survey has a 

more limited number of data elements than the enrolment survey. These datasets have been merged by the 

Education, Culture and Tourism Division of Statistics Canada, creating a third file commonly referred to as 

the linkage file. We use the linkage file in this analysis. 

The information is obtained from the administrative records of Canadian degree-granting 

institutions, generally in an individual record format. Approximately 70 percent of post-secondary 

institutions provided regular annual individual information, including student identifiers that allow matching 

to the other two administrative datasets. We therefore focus on students from these institutions.69 All 

information in the USIS are checked for validity edited by the universities and, in some cases, by the province 

and by Statistics Canada. 

The enrolment survey collects information on student counts as of December 1st in all provinces 

except Ontario, where the reference date is November 1st. This means that each student who attends 

university in the fall session is counted only once annually, even though the student may be enrolled in more 

than one program. This student count is used as a proxy for the total number of students enrolled during a 

complete academic year. 

The degrees survey collects information on all students who have received a degree, diploma or 

certificate during the calendar year ending in December. It is a count of the number of degrees, diplomas 

and certificates awarded, not the number of individual students who receive them. 

 

69 For more on the USIS and the match to the T1FF, see Heisz (2001) and Heisz (2003). 
36 

From the enrolment data, we keep all males that began a full-time undergraduate program at a postsecondary school institution between the ages of 17 and 20. We note students’ graduation date, or last year 

enrolled full time (plus one since enrolment was recorded as of December 1). Experience is defined as 

number of years since graduation or number of years since ending full-time post-secondary education. We 

examine earnings starting when experience equals zero, since students are likely to have worked for 7 months 

since graduation. We remove any student taking longer than 8 years to complete an undergraduate degree 

(dropping less than 1 percent of the sample). We also calculate predicted graduation year based on entry year 

plus four. 

The enrolment data includes information on home province. If missing, home province was 

assumed to be the province of the institution the student began their program. After finding that national and 

regional unemployment rates at time of graduation were not correlated with obtaining a subsequent degree, 

we focus on students that obtain no more than one degree. 

 The post-secondary students we examine from the USIS are matched to the T1FF using the student 

identifier. The T1FF is a data set of individual tax records from 1982 to 1999. The T1FF includes 

information on earnings, defined as the sum of taxable earnings from employment and self-employment. 

The dataset also contains information on transfers, as well as age, gender, residential address and an 

identification number for the firm at which the individual is employed. Some students (fewer than 15 percent 

of the sample) were not matched, mostly due to missing identifiers. Missing ID may be because (1) the 

student did not have an IDcode (perhaps because he or she was a foreign student), (2) the student had an ID 

code, but either did not give it to the institution or the institution did not request it, or (3) the institution 

collected the ID code but did not report it on the USIS survey. To remove individuals that have left the 

country, we drop any student that does not file in the last two years of the T1FF data. 

The cross-section outcome variables we examine include whether a student receives a degree, and 

years in post-secondary school. The annual outcome variables we focus on are log earnings, dummy variables 

for not filing taxes, zero earnings, and living in different province than initial province. 

Individuals working in the USIS-T1FF are also matched annually to information about their firms 

from Statistics Canada’s Longitudinal Employment Analysis Program database (LEAP), beginning in 1983. 

The match rate was 96 percent.70 LEAP is a company-level database that includes all employers in Canada, 

both corporate and unincorporated. The database tracks the employment and payroll characteristics of 

 

70 In the case of multiple employers, the main employer is the one from which a worker has the most earnings. In 

defining our mobility measures, we have taken particular care with missing values for firm identifiers and industry codes. 

To address the problem of missing values, we first fill in single missing values with the adjacent past firm identifier or 

industry code. We then estimate a conservative and a more inclusive measure of mobility. The first only considers 

changes between two valid firm identifiers or industry codes. The second treats remaining missing values as a job or 

industry change. The two measures approximate upper and lower bounds of job mobility. 
37 

individual firms from their year of entry to their year of exit.71 Employers in Canada are required to register a 

payroll deduction account and issue a T4 slip to each employee that summarizes earnings received in a given 

fiscal year. The LEAP database includes every business that issues a T4 taxation slip. 

The LEAP includes a 3-digit industry code and information on annual firm size and total payroll 

amounts. We recorded average firm size, and total firm size between 1982 and 1999, and also subtracted the 

mean amounts for each year before averaging. Both methods produced similar results.72 We also recorded 

when individuals switched firms and industries. 

The data are collapsed into cell means by home province, year left post-secondary education, 

predicted year left post-secondary education, and experience. The cell means are matched to national and 

provincial unemployment rates both at time of school exit and predicted school exit. We use Statistics 

Canada’s youth unemployment rate (ages 16 to 25). Results with the full unemployment rate were similar. 

We work with two samples – the two-way student-earnings match, and the three-way match that also 

includes firm variables. The main results are obtained on the former, but estimates differ little between the 

two samples. To maximize the range of cohorts with as much as possible experience history we focus on the 

full range of graduation cohorts that we can match to unemployment rates at time of labor market entry 

(1976-1995). In the empirical analysis, we also report alternative results with subsets of cohorts. Appendix 

Tables 2A and 2B show sample sizes of the two-way match by graduation and experience year for graduation 

cohorts from 1977 to 1995 (including and excluding observations with missing earnings). 

Sensitivity Appendix A: Accounting for Selective College Graduation 

The decision to leave college may be a function of the business cycle.73 If workers postpone college 

exit in recessions, we would expect that the unemployment rate in the year of predicted graduation is positively 

related to college duration. Similarly, since workers with shorter durations are more likely to be able to further 

postpone graduation labor market entrants in a recession are more likely to have longer durations. Appendix 

Table 6 shows the effects on various basic measures of college duration of the national and regional 

unemployment rates, as well as of predicted regional rates, separately for all workers and for those at least on 

grade. We see no significant correlations at the national level or for regional unemployment at the time when 

workers should have graduated were they on grade. However, we see some significant effect of early 

 

71 The self-employed that do not draw a salary are not included on the LEAP database. In addition, businesses 

comprised solely of individuals or partnerships who do not draw a salary are also excluded from the LEAP. 72 The USIS industry code is documented in Statistics Canada’s USIS user guide, 1995. 73 College enrollment decisions also depend on the state of the local labor market. However, the effects appear to be 

small in the U.S. since the 1960s (e.g., the fraction of men age 19 to 21 in college is not affected by the unemployment 

rate for mature workers, see Card and Lemieux (2000) Table 4, nor is the proportion of workers who finish 12th grade 

and start college (Table 5). The unemployment rate at age 17 does not affect the probability of having a college degree, 

but raises the fraction of workers with some college (Table6)). Note that if unemployment triggers entry into college of 

workers with particular unobserved characteristics, this could affect our instrumental variable strategy even if workers 

are not forward looking due to correlation of the unemployment rate at entry and at exit. However, as shown in the next 

section, most of the correlation of unemployment rates fades after three years. 
38 

unemployment rates at actual graduation with duration. For a five percent change in unemployment rates, this 

would imply an increase of 2.5 percentage points (10% relative to the 0.26 average shown in Appendix Table 

1). 

Panels D to F of Appendix Table 7 show the same specifications for those workers on or above 

grade (see also the appendix available on our website for more detail). The effects are somewhat smaller. A 

five point shock to unemployment implies a 0.05 increase in average years of college (corresponding to three 

weeks or 1.4% relative to a mean of 4.11 years). These results suggest that a very small fraction of workers 

who are barely on or above grade tend to extend their stay in college by one or two years.74 The fact that 

unemployment at predicted graduation matters less suggests this is driven primarily by workers who are 

already beyond grade. Consistently, the fact that the results are even weaker for the full sample and the fact 

that being on or above grade is not affected indicates that students overall do not make significant attempts 

to avoid leaving school in a recession by delaying graduation or enrolling in a new program.75

To directly address endogenous college exit we instrument unemployment in the actual year of exit 

with unemployment in the predicted year of exit based on official degree duration. Predicted year of exit is a 

valid instrument for actual year if college entry is uncorrelated with unemployment rates in the year of 

predicted exit, if it has no direct effect beyond the actual unemployment rate, and if it correlates with 

unemployment at actual exit. We believe the exclusion restrictions are valid, since even if students wanted, 

given the covariance structure of unemployment rates it would be hard for them to forecast future 

unemployment rates. The case could be made that the unemployment rate at predicted graduation could in 

itself be viewed as the relevant ‘shock’ to workers’ careers. Thus, we present and discuss both reduced form 

and instrumental variable (IV) estimates. 

The first two columns of Appendix Table 8 present the reduced form estimates of the interactions of 

potential labor market experience for the same specifications as in Table 1 (OLS). Columns 3 and 4 shows the 

IV results and the coefficients on the instrument from the corresponding first stage. The reduced form 

estimates are either equal (all workers) or slightly smaller (graduates) than the corresponding OLS estimates. 

The numbers in Appendix Table 7 imply that delayed entry is unlikely to affect the estimates of the catch-up 

 

74 Additional results in an appendix available on our website suggests that for this sample the probability of being above 

grade 1-3 years is raised marginally. Taking the results from Panel F, if 0.85% of workers stay longer and raise average 

college duration by 0.0056 years, the average additional time spent in college must be more than one year. 75 Note that as pointed out in Section 2, the propensity of obtaining a graduate degree is also not affected by the 

unemployment rate in the year of the first exit from college (a 5 point unemployment shock leads to an increase in the 

probability to obtaining a post graduate degree of 0.008, relative to a mean of 0.2, with the lowest p-value of 0.157 in the 

regional sample for all workers). Post-graduate degrees are specially concentrated in the health professions, social 

sciences, and other majors (25-30% of all graduates obtain a graduate degree) and less concentrated in business, 

engineering, and teaching (8-12% obtain a graduate degree). Our sample restriction tends to more heavily exclude health 

profession and the social sciences than economics and engineering. To assess whether for some of these subjects the 

propensity to obtain a higher degree responds more strongly to unemployment at time of graduation, we ran the 

regressions by major. Social sciences is the only major experiencing consistent increases in the fraction of post-grad 

degrees during recessions, while health professions experiences consistent declines. All other majors show no clear 

patterns. 
39 

pattern in the reduced form. The first stage coefficient is highly significantly different from zero and different 

from one. The ensuing IV results are either the same as OLS (for those on/above grade), or slightly more 

negative and more persistent (for all workers). All IV coefficient estimates are well within the confidence 

intervals for OLS results.76 Since the general effects of unemployment rates on labor market entry are quite 

small, it would have been surprising to find much of a difference. We conclude that OLS is appropriate to 

analyze the effects of early labor market conditions on the long-term career outcomes of Canadian college 

graduates. 

Sensitivity Appendix B: Accounting for Labor Market History 

All estimates presented so far represent summary effects of the dynamic impact of the initial 

unemployment rate plus the dynamic effects of ensuing unemployment rates that correlate with the first. 

They characterize the expected earnings loss of a worker graduating in a recession and help to assess the 

implications of different models of career determination. Another estimate of interest is the long-term impact 

of an isolated temporary shock of labor market conditions for individuals entering the full-time labor market 

for the first time, holding all else constant. This effect can also be compared to similar shocks at later 

experience years to benchmark whether initial shocks, when virtually all labor market entrants must search for 

employment, generate different permanent and transitory effects than subsequent shocks. 

Since the current province of residence is available from income tax records, we can use our data to 

construct unemployment rate histories for each individual starting in 1982. We interact these histories with 

unrestricted experience dummies and include them into the basic model as additional control variables to 

isolate the effect of the unemployment rate at time of college exit. Since we only have complete data for 

‘market history’ of individuals graduated starting in 1982, we focus on this restricted group of cohorts.77

Although shocks are highly persistent initially, the auto-covariance structure dips to zero after three to four 

years.78 Thus, the inclusion of two to three lags should suffice to absorb most of omitted variable bias. 

Table 2 shows a series of models with augmented controls for unemployment history, each 

interacted with experience. The table shows the basic regional model with the graduate sample for two 

models with outcomes recorded between 1982 and 1995. To compare similarly defined unemployment 

 

76 Note that Hausman tests cannot be read off the tables since standard are clustered at either graduation cohort or 

graduation cohort-initial province level. Although we could implement a test based on Davidson and McKinnon’s (1989) 

approach, we believe that the differences so small that it would not reverse our conclusions. 77 As shown in Appendix Figure 1, this group of cohorts has slightly more persistent effects of initial labor market 

conditions. We have also experimented with including cohorts with incomplete unemployment histories. The results, 

available in an Appendix on our website, are qualitatively the same. We also included unemployment histories based on 

unemployment rates for all workers, with no differences in the results. 

78 If as commonly done we specify the time series process of the unemployment rate as an AR(2), the coefficients are 

0.87 and -.158 for the first and second lag, respectively, in a sample pooling all states and including year and state fixed 

effects (a procedure followed by Blanchard and Katz 1992). Figures of the auto-covariance structure and further 

discussion are available in an Appendix accessible on our website. 
40 

shocks, all models include current province fixed effects.79 The first model includes the unemployment rate at 

the current experience year interacted with experience dummies, without additional labor market history. As 

expected, this has some small initial effects for experience years one to three, but little thereafter. Given that 

each of these unemployment rates has itself a potentially dynamic effect, the next models include interactions 

of these unemployment rates with experience dummies. 

The first model, shown in Column 3 of Table 2 only includes dynamic effects of unemployment rates 

occurring in experience years one to three. The result shows an increasing spread in the two estimates that 

flattens out after experience year 5, exactly as predicted by the omitted variable bias calculations in Section 2. 

At each experience year the worker is exposed to additional shocks correlated with the initial shock that in 

itself have dynamic effects, leading to an increasing bias; as the effects of shocks decline for mature workers 

(as shown in Table 7) and the correlation with unemployment fades or becomes slightly negative, the size of 

the gap stabilizes. Towards experience year eight the estimates become imprecise as the number of cohorts 

decline. The next model in Column 4 includes the entire interacted history for each experience year from one 

to ten. As predicted, the model is extremely similar to the one in Column 3 (however, the joint hypothesis 

that all additional coefficients or that all dynamic effects at higher experience years are jointly equal to zero is 

rejected by an F-test). Overall, the effect of the unemployment rate a worker faces in the year of college entry 

has a long term effect even when controlling for unrestricted dynamic effects of each single unemployment 

shock experience afterwards. 

Since the estimates at later experience become imprecise, we now turn to a grouped model. We 

restrict the dynamic effects to be equal in two-year intervals (i.e., the effects of the unemployment rate at 

experience years 0-1, 2-3, 4-5, etc., is constrained to be equal). To keep the size of the coefficients comparable 

to that of the main model, we take the averages of unemployment rates within groups (the results are the 

same if we were to compare coefficients at two standard deviations of the respective regressors). The fully 

interacted model with grouped unemployment rates then is 

crt t r c e e cr r e r r crt log w = + + + + (UR +UR )/2 + (UR +UR )/2 + ... + u ,0 0 1 ,1 2 3 1 2 3

φ θ χ γ β β . (3) 

Our data does not allow us to estimate the dynamic effects of unemployment shocks at experience years 

greater than three with a sufficient degree of precision due to a declining number of cohorts.80 Thus, we 

present dynamic estimates for groups 0-1 and 2-3, and include additional dynamic interactions as controls for 

omitted variable bias. The dynamic effect at experience year 2-3 will help us to give a benchmark for the size 

of the impact of initial labor market conditions. 

 

79 As shown in Appendix Figure 1 and discussed in Section 3, this has little bearing on our original results. 80 Thus, dynamic estimates for unemployment shocks at higher experience years pick up the behavior of a limited 

number of cohorts. While interesting in its own right, the analysis of single cohorts is left to a separate study. 
41 

The effect of a single shock at experience zero and the effect of the average unemployment in 

experience years zero and one are very similar. The last columns of Table 2 then show the model with fully 

interacted controls for grouped unemployment history. The coefficient estimates are graphed in Figure 4 

(Panel A). The effect of omitted variable bias is again as predicted. Moreover, now the estimated effects are 

smooth and show a similar convergence pattern as before.81

Sensitivity Appendix C: The Role of Regional Mobility 

In columns 6 and 7 of Table 6, we compare the effect of initial unemployment rates on the gains 

from regional mobility by experience. Interestingly, while regional movers gain more if affected by an early 

recession initially, these gains fade after experience year three. It is those who stay in the region or residence 

who have consistently higher earnings growth. Thus, while regional mobility may still be as beneficial in 

booms as in recessions, it appears regional movers do not have permanently higher rates of catch up than 

regional stayers. That gains at regional mobility are not as exceptional as gains at job or industry moves also 

results from the fact that average earnings growth for region movers and stayers is quite similar, as shown in 

the last columns of Panel A. 

It appears that regional mobility is not as important in Canada as in the U.S. (Waggoner 2004). To 

further explore whether the higher job mobility for workers entering the job market in recessions is 

associated with higher mobility across provinces, the last columns of Table 3 shows the effects of the 

unemployment rate at college exit on subsequent provincial mobility. The national unemployment rate is 

uncorrelated with moving to other provinces for both the full sample and graduate sample in Columns 5 to 6 

respectively. The results here suggest no inter-provincial mobility response from worsening in overall 

economic conditions. For the regression models identifying regional economic shocks, however, we do 

observe initially increased provincial mobility for cohorts exposed to higher unemployment conditions at time 

of college exit. For the graduate sample, a 5 percentage point difference in the unemployment rate at entry is 

associated with about a .75 percentage point difference in the provincial mobility rate in the first two years. 

This rate is about half that for firm mobility, and drops quickly after the third year.82 The small effect of 

unemployment at college exit on provincial mobility suggests that most of the pattern of catch-up in wages 

over time for individuals that began the labor market in a recession occurs within provinces.

 

81 If we repeat the exercise with the full set of cohorts (for which we do not have complete history controls) the results 

are very similar for the grouped model, with complete convergence occurring after six years in the labor market (shown 

in an Appendix Figure available on our website). 82 After the fifth year out of college, the unemployment rate at time of exit is negatively correlated with provincial 

mobility. Those induced to move to another province from entering the local labor market during high unemployment 

appear to be less likely to move thereafter. We also replicated our estimates separately for workers who never switch 

region and for movers. Those never moving, about three quarters of our sample, behave very similar as the full sample 

(see Web Appendix). 
42 

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Figure 1: Experience-Profiles in Earnings, Mobility, and Firm Charachteristics Canada and U.S.

Panel A: Change in Annual Earnings Panel B: Fraction Job C

Panel C: Change in Firm Size Panel D: Labor Force Att

-0.2

-0.1

0

0.1

0.20.30.40.50.60.7

0 1 2 3 4 5 6 7 8 9 10

Experience

% Change in Firm Size

CanadaUSAPoly. (USA)

0

0.05

0.1

0.15

0.2

0.25

0.3

0.35

0.4

0.45

1 2 3 4 5 6 

Experience

Fraction Changed Job

0

0.20.40.60.8

1

1.2

1 2 3 4 5 6 7 8 9 10

Experience

% Change Annual Earnings Canada USA 

0

0.1

0.20.30.40.50.6

0 1 2 3 4 5 6 

Average Log Firm Earnings Avg. Log 

Avg. Log 
Figure 2B: Mature and Entry Level Earnings and Experience Profiles by Graduation Year

Figure 2A: Unemployment Rates Ages 14-24 for Canada and Provinces 1976-2000

0

5

10

15

20

25

30

35

1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000

Year

Percent

Canada New Foundland Prince-Edward Island

Nova Scotia New Brunswick Quebec

Ontario Manitoba Saskatchewan

Alberta British Columbia

4.5

4.7

4.9

5.1

5.3

5.5

5.7

5.9

6.1

6.3

6.5

82 83 84 85 86 87 88 89 90 91 92 93 94 95 96 97 98

Calendar Year

Log Real Annual Earnings

Entry Level Annual Earnings 

(Experience Year One)

Mature Workers' Annual Earnings 

(Experience Years 5 to 10)

Annual Earnings by Potential Labor 

Market Experience
Figure 3A: Earnings By Experience Year For Cohorts Entering Labor Market 1978 to 1993

Figure 3B: Aggregate Unemployment and Wage Fluctuations by Experience-Level

-0.08

-0.06

-0.04

-0.02

0

0.02

0.04

0.06

0.08

0.1

78 80 82 84 86 88 90 92

Year of Graduation

Earnings Residuals

-8

-6

-4

-2

0

2

4

6

8

10

Exp =1 Exp = 2

Exp = 3 Exp = 4

Exp = 5 Exp = 6-10

 UR 15-24

-0.2

-0.15

-0.1

-0.05

0

0.05

0.1

0.15

0.2

0.25

82 84 86 88 90 92 94 96 98

Year

Log Real Annual Earnings

-8

-6

-4

-2

0

2

4

6

8

10

Unemployment Rate

Exp = 0 Exp =1 

Exp = 2 Exp = 3

Exp = 4 Exp = 5-10

 UR 15-24
National/Regional National National Regional National National Regional

Trend Linear Quadratic NA Linear Quadratic NA

D>=0? No No No Yes Yes Yes

(1) (2) (3) (4) (5) (6)

Experience Year

0 -0.021 -0.0224 -0.0168 -0.0231 -0.0232 -0.0187

[0.0047]*** [0.0039]*** [0.0026]*** [0.0037]*** [0.0036]*** [0.0024]***

1 -0.0177 -0.0187 -0.0194 -0.0168 -0.0169 -0.0181

[0.0052]*** [0.0028]*** [0.0024]*** [0.0049]*** [0.0026]*** [0.0021]***

2 -0.0128 -0.0137 -0.0166 -0.0116 -0.012 -0.0154

[0.0033]*** [0.0026]*** [0.0022]*** [0.0030]*** [0.0021]*** [0.0019]***

3 -0.0084 -0.0089 -0.012 -0.006 -0.0066 -0.0117

[0.0022]*** [0.0022]*** [0.0021]*** [0.0022]** [0.0015]*** [0.0017]***

4 -0.0061 -0.006 -0.0093 -0.0036 -0.004 -0.0096

[0.0025]** [0.0027]** [0.0020]*** [0.0028] [0.0021]* [0.0016]***

5 -0.0065 -0.0055 -0.0072 -0.0035 -0.0032 -0.0081

[0.0029]** [0.0020]** [0.0019]*** [0.0024] [0.0015]** [0.0016]***

6 -0.0027 -0.0023 -0.0062 -0.0018 -0.0012 -0.0071

[0.0032] [0.0020] [0.0020]*** [0.0027] [0.0018] [0.0017]***

7 -0.003 -0.0027 -0.0061 -0.0019 -0.001 -0.0071

[0.0043] [0.0023] [0.0020]*** [0.0034] [0.0018] [0.0017]***

8 -0.0001 0.0002 -0.0043 -0.0008 0.0006 -0.0061

[0.0049] [0.0028] [0.0019]** [0.0034] [0.0016] [0.0017]***

9 0.0035 0.0038 -0.0035 0.0021 0.0038 -0.0051

[0.0047] [0.0027] [0.0019]* [0.0033] [0.0017]** [0.0017]***

10 0.0066 0.0051 -0.0015 0.0047 0.0049 -0.0032

[0.0048] [0.0028]* [0.0020] [0.0034] [0.0022]** [0.0017]*

Constant 7.3951 -3.6341 8.8017 7.673 -2.0294 9.0456

[0.2571]*** [2.3916] [0.1012]*** [0.2095]*** [0.8040]** [0.0668]***

N 14407 14407 14407 8679 8679 8679

R-squared 0.76 0.77 0.8 0.93 0.93 0.95

Specification

Table 1: Effect of Unemployment Rate at time of Graduation on Log Real Earnings by 

Potential Experience

Note: The sample includes males in Canada leaving university between 1976 and 1995. 'D' indicates the difference

between the actual year left and the predicted year of graduation based on year of entry and program. Sample sizes

reflect cell sample sizes after collapsing the micro data by graduation cohort, province of residence in each year of

graduation, and experience year. The national model regresses log annual earnings on the youth unemployment rate

in the country at the year of college exit, interacted with experience years 0 to 10, plus province of residence fixed

effects, experience fixed effects, and a linear or quadratic graduation cohort trend. The regional model regresses log

annual earnings on the youth unemployment rate in the province of first residence, interacted with experience years 0

to 10, plus province of residence fixed effects, experience fixed effects, and year of graduation fixed effects. The

coefficients shown are the unemployment rate at college exit and experience interactions. One, two, and three asterix

indicates statistical significance at the 10 percent, 5 percent, and 1 percent levels respectively. See text for more

details.
Panel A: Log Real Annual Earnings Panel B: Probability of Annual Change in Emplo

Panel C: Average Firm 'Quality', Graduates Only Panel D: Fraction not Working, Differen

Figure 4: The Persistent Effects of Unemployment in the Year of Graduation on Earnings, Job Mobility, and F(Graduation Cohorts 1976-1995)

0

0.0005

0.001

0.0015

0.002

0.0025

0.003

0.0035

0 1 2 3 4 5 6 

Experience

Log Real Earnings

Graduates OnlyAll Workers with Some Coll

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

All Workers with Some CollegeGraduates Only

-0.025

-0.02

-0.015

-0.01

-0.005

0

0.005

0.01

0 1 2 3 4 5 6 7 8 9 10

Experience

Percent Difference

-0.012-0.01-0.008-0.006-0.004-0.0020

Average Log PayrollAverage Log Firm SizeAverage Median Wage

-0.001

-0.0005

0

0.0005

0.001

0.0015

0.002

0 1 2 3 4 5 6 

Experience

On Zero Mis
Model

Baseline 

(No UR 

History)

With 

Current 

UR*Exp

With 

History in 

Exp=1,2,3

With Full 

UR 

History

Baseline 

Group 0-1 

(No Hist.)

Group 01 

With Full 

History

Group 23 

With Full 

History

(1) (2) (3) (4) (5) (6) (7)

Experience Year

0 -0.0174 -0.0184 -0.0173 -0.0159 -0.0165 -0.0147 ---

[0.0028]*** [0.0028]*** [0.0027]*** [0.0028]*** [0.0030]*** [0.0031]*** ---

1 -0.0176 -0.0178 -0.017 -0.0172 -0.0177 -0.016 ---

[0.0024]*** [0.0047]*** [0.0044]*** [0.0048]*** [0.0025]*** [0.0026]*** ---

2 -0.016 -0.0142 -0.014 -0.0121 -0.0167 -0.0141 -0.0026

[0.0021]*** [0.0025]*** [0.0041]*** [0.0041]*** [0.0020]*** [0.0025]*** [0.0025]

3 -0.0128 -0.0117 -0.0087 -0.0094 -0.0134 -0.0107 -0.004

[0.0019]*** [0.0020]*** [0.0037]** [0.0035]*** [0.0018]*** [0.0024]*** [0.0024]*

4 -0.0117 -0.0113 -0.0063 -0.008 -0.0122 -0.0102 -0.0036

[0.0018]*** [0.0019]*** [0.0039] [0.0038]** [0.0018]*** [0.0024]*** [0.0030]

5 -0.011 -0.0108 -0.0076 -0.0082 -0.0113 -0.0095 -0.0026

[0.0018]*** [0.0018]*** [0.0046] [0.0043]* [0.0016]*** [0.0026]*** [0.0041]

6 -0.0104 -0.0102 -0.008 -0.0076 -0.0109 -0.0065 -0.0071

[0.0019]*** [0.0019]*** [0.0055] [0.0048] [0.0018]*** [0.0030]** [0.0043]

7 -0.0105 -0.0105 -0.0104 -0.0099 -0.0108 -0.0072 -0.0044

[0.0019]*** [0.0019]*** [0.0049]** [0.0046]** [0.0019]*** [0.0032]** [0.0041]

8 -0.0095 -0.0095 -0.0067 -0.0049 -0.0099 -0.0066 -0.0024

[0.0019]*** [0.0019]*** [0.0050] [0.0043] [0.0020]*** [0.0029]** [0.0040]

9 -0.0086 -0.0085 -0.0103 -0.0091 -0.009 -0.0054 -0.0029

[0.0019]*** [0.0019]*** [0.0051]** [0.0038]** [0.0020]*** [0.0032]* [0.0045]

10 -0.006 -0.0054 -0.0125 -0.0115 -0.0062 -0.0021 -0.0032

[0.0021]*** [0.0021]*** [0.0055]** [0.0050]** [0.0023]*** [0.0039] [0.0051]

Constant 9.2257 9.2636 9.2633 9.2379 9.2195 9.2031 ---

[0.0982]*** [0.1023]*** [0.0969]*** [0.1034]*** [0.1040]*** [0.1102]*** ---

N 7536 7536 7536 6994 7536 7299 ---

R2 0.96 0.96 0.96 0.97 0.96 0.96 ---

Specification

Table 2: Effect of Unemployment Rate at time of Graduation With Controls for UR 

History, Basic and Grouped Model - Graduate Sample, Regional Model, Cohorts 1982-

1995

Note: The sample includes males in Canada graduating university (with D>=0) between 1982 and 1995 . Sample

sizes reflect cell sample sizes after collapsing the micro data by graduation cohort, province of residence in each

year of graduation, and experience year. The national model regresses log annual earnings on the youth

unemployment rate in the country, instrumented with the youth unemployment rate in the province of first

residence (the columns indicate whether this rate is averaged over the first 1 to 3 years), interacted with experience

years 0 to 10, plus province of residence fixed effects, experience fixed effects, and year of graduation fixed

effects. The columns indicate additional controls for experience interacted with later unemployment rates. One,

two, and three asterix indicates statistical significance at the 10 percent, 5 percent, and 1 percent levels

respectively. See text for more details.
Panel A: Effect of Unemployment Rate at Time of Graduation on Earnings Controlling 

for Dynamic Effects of Further Unemployment Shocks (by Experience Groups)

Panel B: Effect of Unemployment Rate at Time of Graduation Unemployment at Time of Predicted Graduation

Figure 5: Selected Results from Sensitivity Analysis (Effect of Further Unemployment Shocks, Selective Graduation, Selective ParDifferences)

Panel C: Effect of Unemployment Rate at Time of Graduation for Workers with Positive Earnings Each Period

Panel D: Effect of Unemployment Rate at Time of GraduatioGroups of Graduation Cohorts

-0.02

-0.018-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings Baseline (No History)

Group 01 (No History)Group 01 (Full History)Group 23 (Full History)

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 

Experience

Log Real Earnings

1976-1991982-1991976-1981976-1991984-199

Graduati

-0.02

-0.018-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0 1 2 3 4 5 6 

Experience

Log Real Earnings

Instrumental Variable EstimateOLS Estimates (Main Model)

-0.02

-0.018-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

Panel SampleMain Model
Nat./Reg.

Outcome Fraction 

Zero 

Earnings

Fraction 

Not in 

Sample

Fraction on 

UI

Fraction 

Changed 

Province

Fraction 

Left First 

Province

Fraction 

Changed 

Province

Fraction 

Left First 

Province

(1) (2) (3) (4) (5) (6) (7)

Experience Year

0 0.0008 0.0014 0.0017 - -

[0.0001]*** [0.0005]*** [0.0003]*** - -

1 0.0003 0.0002 0.0011 0 0.0007 0.0023 0.001

[0.0001]*** [0.0003] [0.0002]*** [0.0003] [0.0005] [0.0004]*** [0.0010]

2 0.0002 -0.0001 0.0009 0.0001 0.0003 0.0014 0.0029

[0.0001]** [0.0003] [0.0002]*** [0.0002] [0.0006] [0.0002]*** [0.0008]***

3 -0.0001 0 0 0 -0.0001 0.0008 0.0036

[0.0001] [0.0003] [0.0002] [0.0002] [0.0005] [0.0002]*** [0.0008]***

4 0 0.0001 0.0002 0 -0.0005 0.0001 0.0039

[0.0001] [0.0002] [0.0001] [0.0002] [0.0005] [0.0002] [0.0008]***

5 0 -0.0001 -0.0001 -0.0001 -0.0007 -0.0003 0.0038

[0.0001] [0.0003] [0.0002] [0.0002] [0.0006] [0.0002] [0.0008]***

6 0 -0.0005 -0.0002 0.0002 -0.0005 -0.0004 0.0036

[0.0001] [0.0003]* [0.0002] [0.0002] [0.0006] [0.0002]* [0.0008]***

7 0 -0.0002 -0.0002 -0.0003 -0.0008 -0.0008 0.0035

[0.0001] [0.0002] [0.0002] [0.0002]* [0.0005] [0.0002]*** [0.0008]***

8 0 -0.0002 -0.0005 -0.0001 -0.0003 -0.0008 0.0034

[0.0001] [0.0003] [0.0002]*** [0.0003] [0.0008] [0.0003]*** [0.0008]***

9 -0.0002 -0.0004 -0.0007 0.0001 -0.0003 -0.0008 0.0031

[0.0001]*** [0.0002]* [0.0002]*** [0.0002] [0.0009] [0.0003]*** [0.0008]***

10 0 -0.0008 -0.0005 0 -0.0005 -0.0006 0.0031

[0.0001] [0.0003]*** [0.0002]*** [0.0002] [0.0008] [0.0003]** [0.0009]***

Constant -0.0032 0.0227 0.0162 0.006 -0.0399 0.0227 0.0305

[0.0025] [0.0118]* [0.0072]** [0.0097] [0.0315] [0.0068]*** [0.0307]

N 8679 8679 8679 5909 5942 5909 5942

R2 0.2 0.39 0.34 0.08 0.14 0.4 0.71

Table 3: Effect of Unemployment Rate at time of Graduation on Labor Force Participation - and 

Province Mobility - Graduate Sample

Specification

Regional National Regional

Note: Columns indicate outcome variable used as the dependent variable. Each model regresses these outcomes on the youth

unemployment rate in the province of first residence, interacted with experience years 0 to 10, plus province of residence fixed

effects, experience fixed effects, and year of graduation fixed effects. The coefficients shown are the unemployment rate at

college exit and experience interactions. One, two, and three asterix indicates statistical significance at the 10 percent, 5 percent,

and 1 percent levels respectively. See text for more details.
Outcome Fraction 

Changed 

Firm

Fraction 

Changed 

Industry

Fraction 

Left First 

Firm

Fraction 

Left First 

Industry

Log Firm 

Size

Fraction 

Firm Size 

> 1000

Average 

Median 

Firm Wage

Average 

Log Firm 

Payroll

(1) (2) (3) (4) (5) (6) (7) (8)

Experience Year

0 ---- -0.008 -0.0016 -0.0097 -0.0169

---- [0.0050] [0.0008]* [0.0014]*** [0.0058]***

1 0.0029 0.0021 0.0038 0.0025 -0.0115 -0.002 -0.0096 -0.0224

[0.0008]*** [0.0007]*** [0.0010]*** [0.0011]** [0.0049]** [0.0009]** [0.0011]*** [0.0055]***

2 0.0031 0.0034 0.0046 0.0041 -0.0088 -0.002 -0.0073 -0.0173

[0.0007]*** [0.0006]*** [0.0011]*** [0.0011]*** [0.0050]* [0.0008]** [0.0011]*** [0.0056]***

3 0.0021 0.0023 0.0049 0.0045 -0.0034 -0.0012 -0.0057 -0.0107

[0.0007]*** [0.0006]*** [0.0009]*** [0.0009]*** [0.0047] [0.0008] [0.0010]*** [0.0052]**

4 0.0018 0.0015 0.0052 0.0046 -0.0022 -0.0009 -0.0044 -0.008

[0.0006]*** [0.0006]** [0.0009]*** [0.0009]*** [0.0048] [0.0008] [0.0011]*** [0.0054]

5 0.0022 0.0019 0.0043 0.0039 0.0025 -0.0003 -0.0039 -0.0023

[0.0005]*** [0.0005]*** [0.0010]*** [0.0010]*** [0.0051] [0.0009] [0.0012]*** [0.0057]

6 0.0015 0.0011 0.0043 0.004 0.0014 -0.0005 -0.0049 -0.0046

[0.0005]*** [0.0005]** [0.0010]*** [0.0010]*** [0.0050] [0.0009] [0.0012]*** [0.0056]

7 0.0018 0.002 0.0041 0.0039 0.0013 -0.0007 -0.005 -0.0047

[0.0006]*** [0.0006]*** [0.0011]*** [0.0010]*** [0.0054] [0.0009] [0.0012]*** [0.0060]

8 0.0018 0.002 0.0044 0.0042 0.0029 -0.0003 -0.0044 -0.0022

[0.0008]** [0.0007]*** [0.0011]*** [0.0010]*** [0.0054] [0.0009] [0.0011]*** [0.0060]

9 0.0016 0.002 0.0047 0.0052 0.0044 0.0001 -0.0035 0.0002

[0.0010] [0.0009]** [0.0010]*** [0.0010]*** [0.0055] [0.0009] [0.0011]*** [0.0063]

10 0.0013 0.0015 0.005 0.0055 0.0048 0.0002 -0.002 0.0021

[0.0011] [0.0011] [0.0010]*** [0.0010]*** [0.0068] [0.0010] [0.0015] [0.0077]

Constant 0.3407 0.3151 0.1391 0.523 8.1745 0.719 0.8069 7.2971

[0.0184]*** [0.0187]*** [0.0428]*** [0.0403]*** [0.1953]*** [0.0283]*** [0.0368]*** [0.2203]***

N 5871 5871 5863 5861 8435 8435 8435 8435

R2 0.8 0.79 0.86 0.77 0.53 0.47 0.75 0.6

Specification

Table 4: Effect of Unemployment Rate at time of Graduation on Job and Industry Mobility and 

Average Employer Characteristics - Graduates Only

Note: Columns indicate the firm or industry mobility variable used as the dependent variable. Each model regresses these

outcomes on the youth unemployment rate in the province of first residence, interacted with experience years 0 to 10, plus

province of residence fixed effects, experience fixed effects, and year of graduation fixed effects. The coefficients shown are the

unemployment rate at college exit and experience interactions. One, two, and three asterix indicates statistical significance at the

10 percent, 5 percent, and 1 percent levels respectively. See text for more details.
(1) (2) (3) (4) (5) (6) (7)

Experience Year

1 0.47 0.49 0.46 0.46 0.46 0.56 0.47

2 0.21 0.27 0.18 0.29 0.18 0.27 0.21

3 0.18 0.25 0.15 0.28 0.15 0.24 0.18

4 0.15 0.23 0.12 0.23 0.12 0.20 0.15

5 0.12 0.19 0.10 0.22 0.10 0.15 0.12

6 0.09 0.13 0.08 0.11 0.08 0.11 0.09

7 0.08 0.12 0.07 0.10 0.07 0.10 0.08

8 0.07 0.09 0.06 0.05 0.06 0.06 0.07

9 0.06 0.08 0.06 0.05 0.06 0.08 0.06

10 0.06 0.08 0.05 0.06 0.05 0.06 0.06

Experience Year

1 -0.0004 0.001 -0.002 0.006 -0.002 0.011 -0.001

2 0.006 0.006 0.005 0.007 0.005 0.001 0.006

3 0.007 0.008 0.006 0.015 0.006 0.005 0.007

4 0.005 0.006 0.004 -0.001 0.004 -0.010 0.005

5 0.004 0.007 0.003 0.007 0.003 -0.005 0.004

6 0.003 0.005 0.002 0.006 0.002 -0.017 0.003

7 0.002 0.002 0.002 0.006 0.002 -0.002 0.002

8 0.003 0.004 0.003 0.004 0.003 -0.004 0.003

9 0.002 -0.0002 0.002 -0.007 0.002 -0.008 0.002

10 0.002 0.003 0.002 -0.002 0.002 -0.013 0.003

Table 5: Wage Growth for Stayers and Movers Between Firms, Industries, and Provinces -- Averages 

and Effects of Initial Unemployment Rate, Graduates Only

 Gains of 

Province 

Stayers

 Gains of 

Job 

Movers

 Gains of 

Job 

Stayers

Panel B: Effect of Unemployment Rate on Wage Growth by Mover Status

 Gains of 

Industry 

Movers

 Gains of 

Industry 

Stayers

 Gains of 

Province 

Movers

Panel A: Average Wage Growth by Mover Status 

 Overall 

Earnings 

Growth

Note: Panel A contains simple averages of annual wage growth among males in Canada graduating university (with D>=0)

between 1976 and 1995. For Panel B, columns indicate selected sample for regression models of annual increases in log

earnings on the youth unemployment rate in the province of first residence, interacted with experience years 0 to 10, plus

province of residence fixed effects, experience fixed effects, and year of graduation fixed effects. The coefficients shown are

the unemployment rate at college exit and experience interactions. Bold (italicized) coefficients signify significance at the 1

(10) percent levels respectively. See text for more details.
Figure 6A: Effect of Initial Unemployment Rate Controlling for First Firm and Industry 

Fixed Effects Interacted With Experience - Graduates Only

Figure 6B: Effect of Initial Unemployment Rate by Firm Type - Graduates Only

-0.02

-0.015

-0.01

-0.005

0

0.005

0 1 2 3 4 5 6 7 8 9 10

Experience

Percent Difference

Original (No FE)

With Initial Firm*Exp FE

With Initial Firm FE

With Initial Ind FE

-0.018

-0.016

-0.014

-0.012

-0.01

-0.008

-0.006

-0.004

-0.002

0

0 1 2 3 4 5 6 7 8 9 10

Experience

lpayroll<75th

lpayroll>=75th

lmedearn<75th

lmedearn>=75th
Outcome Annual Earnings On UI Zero Earnings Changed Firm Changed 2-Digit Industry Firm Avg. Median Wage Fir

>

Experience Group

0-10 -0.088 0.005 0.013 -0.046 -0.031 -0.019 0

(0.027) (0.002) (0.003) (0.012) (0.010) (0.020) (0

0-1 -0.277 0.012 0.028 0.021 0.016 -0.142 -

(0.035) (0.005) (0.004) (0.014) (0.013) (0.020) (0

2-3 -0.186 0.007 0.019 -0.024 -0.008 -0.074 0

(0.030) (0.003) (0.004) (0.016) (0.013) (0.018) (0

4-5 -0.065 0.001 0.005 -0.043 -0.025 0.020 0

(0.028) (0.003) (0.003) (0.016) (0.013) (0.016) (0

6+ -0.037 0.000 -0.001 -0.061 -0.043 0.050 0

(0.023) (0.002) (0.003) (0.017) (0.013) (0.019) (0

Experience Group

0-10 -0.115 0.005 0.012 -0.041 -0.026 -0.043 0

(0.023) (0.002) (0.003) (0.012) (0.010) (0.015) (0

0-1 -0.276 0.009 0.024 0.026 0.020 -0.150 -

(0.036) (0.005) (0.004) (0.014) (0.013) (0.022) (0

2-3 -0.188 0.007 0.018 -0.022 -0.005 -0.098 0

(0.034) (0.004) (0.004) (0.017) (0.015) (0.018) (0

4-5 -0.086 0.003 0.005 -0.048 -0.028 -0.005 0

(0.031) (0.003) (0.004) (0.017) (0.014) (0.018) (0

6+ -0.102 0.002 0.002 -0.057 -0.039 0.035 0

(0.028) (0.002) (0.003) (0.018) (0.013) (0.022) (0

Panel A: Log(Unemployment Rate Ages 15-24)

Panel B: Log(Unemployment Rate All Ages)

Table 6: Experience Gradient in Annual Earnings, Job Mobility, and Other Outcomes (Elasticities) - All Workers

Notes: Standard errors clustered at the province-cohort level are in parentheses. The unemployment rate refers to the unemployment rcurrent residence. Coefficients are from models ran separately by indicated group of potential experience. Controls include fixed effectprovince, and current province. The unemployment rate for job-mobility is lagged by one period. In the case of linear probability modreplaced by quartic trends. 
Table 7: Heterogeneity in Initial Loss and Reversion by Worker Groups

x

Bottom 

Quintile

Middle 

Quintile

Top 

Quintile

Drop -0.0183 -0.0282 -0.0240 -0.0134

(0.0000) (0.0000) (0.0000) (0.0000)

Slope 0.0020 0.0021 0.0025 0.0019

(0.0000) (0.0021) (0.0025) (0.0019)

Fade 0.0015 0.0016 0.0021 0.0012

(0.0000) (0.0000) (0.0000) (0.0000)

Drop -0.0093 -0.0098 -0.0125 -0.0074

(0.0000) (0.0000) (0.0000) (0.0000)

Slope 0.0010 -0.0001 0.0014 0.0014

(0.0000) -(0.0001) (0.0014) (0.0014)

Fade 0.0007 -0.0001 0.0009 0.0009

(0.0000) (0.0000) (0.0000) (0.0000)

Drop -0.0102 -0.0078 -0.0162 -0.0073

(0.0000) (0.0003) (0.0001) (0.0000)

Slope 0.0021 0.0004 0.0006 0.0036

(0.0000) (0.0004) (0.0006) (0.0036)

Fade 0.0016 -0.0009 0.0008 0.0033

(0.0000) (0.0000) (0.0000) (0.0000)

Jump 0.0032 0.0012 0.0025 0.0043

(0.0000) (0.0000) (0.0000) (0.0000)

Slope 0.0002 -0.0002 0.0004 0.0003

(0.0000) -(0.0002) (0.0004) (0.0003)

Fade 0.0001 0.0002 0.0005 -0.0001

(0.0000) (0.0000) (0.0000) (0.0000)

Drop 0.0030 -0.0010 0.0011 0.0066

(0.0000) (0.0000) (0.0000) (0.0000)

Slope -0.0002 0.0005 0.0004 -0.0013

(0.0000) (0.0005) (0.0004) -(0.0013)

Fade -0.0001 0.0003 0.0003 -0.0008

(0.0000) (0.0000) (0.0000) (0.0000)

Drop 0.0012 -0.0002 0.0009 0.0014

(0.0000) (0.0000) (0.0000) (0.0000)

Slope -0.0003 -0.0003 -0.0001 -0.0004

(0.0000) -(0.0003) -(0.0001) -(0.0004)

Fade -0.0002 -0.0002 0.0000 -0.0002

(0.0000) (0.0000) (0.0000) (0.0000)

Position in Distribution of Predicted 

Annual Earnings

All 

Outcome Variable Graduates

Notes: Coefficients from separate regression models. The initial loss (DROP) is the 

effect of unemployment at graduation (UR) at experience zero and one, the first 

phase of the catch up (SLOPE) is the coefficient on the interaction of UR with linear 

experience for experience years two to six, and the second phase (FADE) of the catch 

up is same interaction for experience years seven to ten.

Average Firm 

Median Log 

Earnings

Annual 

Earnings

Average Firm 

Employment

Fraction 

Changed 

Employer

Fraction Left 1st 

Employer

Fraction Zero 

Earnings
Panel A: Log Real Annual Earnings Panel B: Probability of Annual Change in E

Panel C: Average Firm 'Quality', Graduates Only Panel D: Fraction Filing With Zero Annual Figure 7: Changes of Earnings, Job Mobility, Firm Quality, and Employment due to Entering the Labor Mfor Workers with Different Predicted Earnings

-0.035

-0.03

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 7 8 9 10

Years of Potential Experience

Log Real Earnings

TopMiddleBottom

-0.002

-0.001

0

0.001

0.0020.0030.0040.0050.0060.0070.008

0 1 2 3 4 5 6 

Years of Potential Experience

Fraction Job Change

-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0.0020.004

0 1 2 3 4 5 6 7 8 9 10

Years of Potential Experience

Log Real Earnings

Top Middle Bottom

-0.0025

-0.002

-0.0015

-0.001

-0.0005

0

0.0005

0.001

0.0015

0.002

0.0025

0.003

0 1 2 3 4 5 6 

Years of Potential Experience

Fraction Zero Earnings

z
Panel B: Heterogeneity of Losses as Measured by Present Disctounted Value of 

Earnings or of Utility 

Panel A: Reversion of Losses After 5 Years in the Labor Market by Decile of Predicted 

Earning Distribution, College Graduates

Figure 8: Heterogeneity of Losses from Graduating in a Recession and Channels of 

Catch-Up by Deciles of Distribution of Predicted Earnings

-0.4

-0.2

0

0.2

0.4

0.6

1 2 3 4 5 6 7 8 9 10

Decile of Predicted Earnings Distribution

-0.007

-0.005

-0.003

-0.001

0.001

0.003

0.005

Fraction Earnings Loss Remaining

Closed Gap in Firm Average Wage

Fraction Left First Firm (Right Axis)

-0.05

-0.03

-0.01

0.01

0.03

0.05

0.07

0.09

0.11

0.13

1 2 3 4 5 6 7 8 9 10

Decile of Predicted Earnings Distribution

-0.1

-0.08

-0.06

-0.04

-0.02

0

0.02

% Gap in PDV of Utility (Lucas Measure): Left Axis

% Loss in PDV of Annual Earnings: Right Axis
Panel A: Duration of College

Years Until 

BA

In 

Graduate 

Sample

Fraction 

Above 

Grade

PredictedActual BA 

Years

Years Until 

BA

In 

Graduate 

Sample

Fraction 

Above 

Grade

PredictedActual BA 

Years

3.31 0.63 0.26 -0.10 4.11 0.89 0.40 0.86

(1.29) (0.38) (0.37) (1.69) (0.59) (0.11) (0.39) (1.08)

Fraction D 

>1

Fraction D 

>2

Fraction 

D<-1

Fraction 

D<-2

Fraction D 

>0

Fraction D 

>1

Fraction D 

>2 --

At Exp. 

Zero 0.13 0.06 0.23 0.10 0.52 0.20 0.09 --

Panel B: Unemployment Rates Ages 15-24

Average

Standard 

Deviation Maximum Minimum

National 14.76 2.42 19.2 11.0

Province 14.13 3.98 32.7 6.3

National Detrended 0 2.41 4.53 -3.83

Province Demeaned 0 3.01 6.53 -7.12

Panel C: Provinces

N Fraction Average Std. Dev.

Nova Scotia 1,143 0.84 18.99 2.50

PEI 109 0.08 18.91 2.08

Newfoundland 2,535 1.86 27.11 3.51

New Brunswick 7,281 5.33 20.07 2.13

Quebec 10,472 7.66 17.20 2.60

Ontario 71,995 52.69 13.03 3.14

Manitoba 10,308 7.54 12.59 1.81

Saskatchewan 4,557 3.34 11.84 2.26

Alberta 11,742 8.59 11.68 3.08

British Columbia 16,493 12.07 15.93 3.86

Sample Size Unemployment Rate

Appendix Table 1: Descriptive Statistics from Administrative College Data 1976-1995

Notes: See text and Data Appendix. D=Actual Graduation Year - Graduation Year Based on Program Duration.

At Exp. 

Zero

Entire Sample (Some College) Graduates (Actual ≥ Predicted Year)
Appendix Table 2A. Sample Size by Graduation Cohort and Experience

0 1 2 3 4 5 6 7 8 9 10 Total

1976 3732 3732 3732 3732 3732 18660

1977 6875 6875 6875 6875 6875 6875 41250

1978 7863 7863 7863 7863 7863 7863 7863 55041

1979 7780 7780 7780 7780 7780 7780 7780 7780 62240

1980 7869 7869 7869 7869 7869 7869 7869 7869 7869 70821

1981 7899 7899 7899 7899 7899 7899 7899 7899 7899 7899 78990

1982 8033 8033 8033 8033 8033 8033 8033 8033 8033 8033 8033 88363

1983 9146 9146 9146 9146 9146 9146 9146 9146 9146 9146 9146 100606

1984 8746 8746 8746 8746 8746 8746 8746 8746 8746 8746 8746 96206

1985 9584 9584 9584 9584 9584 9584 9584 9584 9584 9584 9584 105424

1986 9379 9379 9379 9379 9379 9379 9379 9379 9379 9379 9379 103169

1987 9307 9307 9307 9307 9307 9307 9307 9307 9307 9307 9307 102377

1988 9621 9621 9621 9621 9621 9621 9621 9621 9621 9621 9621 105831

1989 9391 9391 9391 9391 9391 9391 9391 9391 9391 9391 9391 103301

1990 9408 9408 9408 9408 9408 9408 9408 9408 9408 9408 94080

1991 9288 9288 9288 9288 9288 9288 9288 9288 9288 83592

1992 9770 9770 9770 9770 9770 9770 9770 9770 78160

1993 10429 10429 10429 10429 10429 10429 10429 73003

1994 14416 14416 14416 14416 14416 14416 86496

1995 10117 10117 10117 10117 10117 50585

Total 136,635 144,534 152,403 160,183 168,046 164,804 154,120 143,691 133,921 124,633 115,225

Appendix Table 2B. Sample with Non-Missing Earnings by Graduation Cohort and Experience

0 1 2 3 4 5 6 7 8 9 10 Total

1976 3416 3364 3387 3367 3429 16963

1977 6320 6263 6322 6227 6303 6233 37668

1978 7284 7199 7199 7073 7173 7050 7168 50146

1979 7119 7058 7088 6934 7026 6937 7032 7097 56291

1980 7226 7134 7208 7073 7139 7041 7135 7194 7138 64288

1981 7166 7115 7160 7069 7097 7004 7102 7139 7131 7096 71079

1982 7083 7204 7274 7170 7214 7131 7209 7287 7218 7201 7288 79279

1983 7863 8144 8130 8249 8201 8221 8284 8221 8214 8221 8351 90099

1984 7723 7796 7896 7763 7858 7906 7876 7774 7835 7895 7899 86221

1985 8422 8637 8561 8637 8689 8672 8599 8616 8689 8693 8742 94957

1986 8443 8456 8512 8557 8524 8440 8457 8475 8476 8560 8571 93471

1987 8308 8428 8453 8375 8318 8311 8364 8383 8453 8473 8672 92538

1988 8790 8776 8717 8661 8670 8658 8668 8746 8773 8854 9029 96342

1989 8621 8530 8451 8433 8460 8411 8440 8557 8666 8785 9391 94745

1990 8532 8454 8427 8421 8445 8452 8532 8658 8742 9408 86071

1991 8325 8300 8294 8302 8392 8410 8510 8632 9288 76453

1992 8650 8707 8737 8806 8814 8895 9044 9770 71423

1993 9284 9389 9410 9371 9462 9650 10429 66995

1994 12756 12863 12941 13160 13376 14416 79512

1995 9149 9152 9291 9403 10117 47112

Total 121949 130002 137435 144721 153159 150350 140367 131047 122352 114167 106104

Graduation 

Year

Years Since Graduation

Graduation 

Year

Years Since Graduation
Year 

of 

Exp.

Average 

Log 

Earnings

Fraction 

on UI

Frac. Not 

in Labor 

Force

Fraction 

Changed 

Firm

Average 

Log 

Earnings

Fraction 

on UI

Frac. Not 

in Labor 

Force

Fraction 

Changed 

Firm

0 8.83 0.016 0.111 - 8.93 0.020 0.102 -

1 9.30 0.023 0.103 0.42 9.49 0.020 0.094 0.40

2 9.51 0.023 0.100 0.35 9.71 0.020 0.093 0.31

3 9.69 0.021 0.099 0.31 9.87 0.016 0.093 0.28

4 9.84 0.017 0.091 0.28 9.99 0.013 0.085 0.25

5 9.96 0.016 0.090 0.25 10.10 0.012 0.085 0.22

6 10.05 0.015 0.092 0.22 10.18 0.011 0.086 0.20

7 10.13 0.013 0.090 0.20 10.25 0.009 0.084 0.18

8 10.20 0.012 0.089 0.18 10.30 0.008 0.082 0.17

9 10.25 0.011 0.086 0.17 10.36 0.007 0.082 0.16

10 10.30 0.010 0.081 0.17 10.40 0.007 0.077 0.16

Year 

of 

Exp.

Average 

Log 

Earnings

Fraction 

Unemployed

Frac. Not 

in Labor 

Force

Fraction 

Changed 

Firma

Average 

Log 

Earnings

Fraction 

Unemployed

Frac. Not 

in Labor 

Force

Fraction 

Changed 

Firma

1 8.94 0.047 0.150 0.349 8.91 0.044 0.144 0.386

2 9.21 0.068 0.132 0.310 9.30 0.064 0.128 0.326

3 9.49 0.045 0.120 0.267 9.57 0.041 0.119 0.258

4 9.59 0.038 0.054 0.216 9.62 0.036 0.054 0.208

5 9.79 0.028 0.055 0.202 9.84 0.025 0.059 0.198

6 9.87 0.040 0.052 0.190 9.91 0.032 0.055 0.180

7 9.81 0.030 0.048 0.171 9.89 0.024 0.048 0.183

8 9.92 0.028 0.039 0.170 9.98 0.019 0.036 0.169

9 9.98 0.015 0.037 0.155 10.05 0.012 0.037 0.146

10 10.03 0.023 0.034 0.142 10.12 0.021 0.035 0.133

a

These figures are calculated as the fraction of workers with one year of tenure from the CPS’ tenure, 

mobility, and pension supplements from 1979 to 2000.

Entire Sample (Some College) Graduates (Actual ≥ Predicted Year)

Appendix Table 3: Experience Profiles in Wages, Participation, and Job Change, Canada and USA 

Panel B: Average Experience Profile USA (Current Population Survey 1994-1996)

Panel A: Average Experience Profile Canada (Income Tax Records, 1982-1999)

Notes: See text and Data Appendix.
Panel A. Mobility Outcomes by Potential Labor Market Experience

Year 

of 

Exp.

Fraction 

Changed 

Industry 1

Fraction 

Changed 

Industry 2

Fraction 

Changed 

Industry 3

Fraction 

Changed 

Province

Fraction 

Left 1st 

Firm

Fraction 

Left 1st 

Industry 1

Fraction 

Left 1st 

Industry 2

Fraction 

Left 1st 

Province

0 -- -- -- -- -- -- -- --

1 0.308 0.352 0.365 0.040 0.399 0.31 0.35 0.052

2 0.220 0.257 0.270 0.029 0.558 0.42 0.48 0.086

3 0.186 0.220 0.233 0.027 0.65 0.48 0.56 0.104

4 0.163 0.194 0.207 0.024 0.709 0.52 0.61 0.115

5 0.141 0.169 0.181 0.021 0.745 0.55 0.64 0.124

6 0.126 0.151 0.163 0.020 0.769 0.56 0.66 0.133

7 0.113 0.135 0.146 0.015 0.784 0.57 0.67 0.138

8 0.104 0.124 0.134 0.012 0.799 0.58 0.68 0.143

9 0.098 0.118 0.128 0.011 0.813 0.59 0.69 0.147

10 0.098 0.116 0.126 0.009 0.827 0.61 0.71 0.150

Panel B. Firm Outcomes by Potential Labor Market Experience

Year 

of 

Exp.

Mean Log 

Firm Size 

Actual 

Mean Firm 

Size 

Fraction 

Firm > 100

Fraction 

Firm > 500

Fraction 

Firm > 

1000

Fraction 

Firm > 

5000

Avg. Log 

Med. Firm 

Earnings

Avg. Log 

Firm 

Payroll

0 6.94 27705 0.73 0.59 0.53 0.34 0.62 5.94

1 6.95 26563 0.74 0.59 0.53 0.33 0.70 6.00

2 7.03 28549 0.75 0.60 0.54 0.33 0.76 6.14

3 7.07 29701 0.75 0.61 0.55 0.34 0.81 6.22

4 7.08 30210 0.75 0.61 0.55 0.34 0.84 6.26

5 7.13 31429 0.76 0.62 0.55 0.35 0.87 6.34

6 7.17 33207 0.76 0.62 0.56 0.36 0.89 6.41

7 7.20 34164 0.76 0.63 0.56 0.36 0.91 6.45

8 7.21 34981 0.76 0.63 0.56 0.37 0.92 6.48

9 7.21 35286 0.76 0.63 0.57 0.37 0.93 6.50

10 7.20 35810 0.76 0.63 0.57 0.37 0.94 6.50

Difference >=0 (Graduates)

Difference >=0 (Graduates)

Appendix Table 4: Experience Profile in Mobility and Firm Characteristics, Canada 1982-1999, 

Graduates Only

Notes: See text and Data Appendix.
Year of 

Experience

Log Firm 

Size

Fraction 

Firm Size 

> 100

Fraction 

Firm Size 

> 500

Fraction 

Firm Size 

> 1000

Log Firm 

Size

Fraction 

Firm Size 

> 100

Fraction 

Firm Size 

> 500

Fra

Firm 

> 

0 5.30 0.58 0.42 0.33 5.70 0.62 0.49 0

1 5.16 0.52 0.40 0.33 5.65 0.61 0.47 0

2 5.58 0.62 0.46 0.37 5.86 0.66 0.51 03 5.43 0.59 0.42 0.34 5.52 0.59 0.44 04 5.32 0.58 0.39 0.33 5.52 0.60 0.42 05 5.65 0.61 0.47 0.36 5.89 0.64 0.50 06 5.79 0.64 0.48 0.39 5.89 0.64 0.50 07 5.70 0.63 0.48 0.38 5.80 0.65 0.50 08 5.56 0.59 0.45 0.37 5.68 0.63 0.47 09 5.96 0.67 0.51 0.44 6.18 0.71 0.54 0

10 5.73 0.63 0.48 0.40 5.88 0.67 0.50 0

All Workers (Some College) At Least 16 Years of Schooling

Notes: Pension and Benefit Supplements to The Current Population Survey, 1979, 1983, 1988. Sample size is 

for all workers with 13 to 18 years of schooling and 2987 for workers with at least 16 years of schooling.Appendix Table 5: Firm Size and Average Firm Wages Experience -- USA
Panel A: Different Early Labor Market Conditions (2 Std.Dev. Shock) Panel B: Different Early Labor Market Horizons (Averag

Panel C: Shock in Region of College vs. Region of First Residence Panel D: Current Province and Current Province-Year C

Appendix Figure 1: Effect of Unemployment Rate at Time of Graduation on Log Real Earnings: Regional GradCohort 1982-1995 (Unless Otherwise Noted) 

-0.18-0.16-0.14-0.12

-0.1

-0.08-0.06-0.04-0.02

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

UR Age 15-24 (Main Model)UR All AgesUR Age 20-24E-Pop Age 15-24

-0.02

-0.018-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0 1 2 3 4 5 6 7 

Experience

Log Real Earnings

UR at Exp=0 (Main Model)Average UR Exp=0-1Average UR Exp=0-2Average UR Exp=0-3

-0.02

-0.018-0.016-0.014-0.012

-0.01

-0.008-0.006-0.004-0.002

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

Prov. 1st Residence (1982-1995)Prov. Coll. (1982-1995)Prov. 1st Residence (1976-1995)Prov. Coll. (1976-1995)

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 7 

Experience

Log Real Earnings

With Current Province-Year FEWith Current-Province FEOnly 1st Province of Residence FE
Notes: Circles correspond to cell sizes.

Appendix Figure 2: Residuals of Separate Regressions of Earnings and Unemployment Rates including YeaGraduation Cohort Dummies, Plotted by Experience Year with Corresponding Regression Line

Experience=0

Unemployment Rate Residuals

 Log-Earnings Residuals Fitted values

-5.44531 3.78972

-.3-.2-.1

0

.1.2.3.4

Experience=3

Unemployment Rate Residuals

 Log-Earnings Residuals Fitted 

-5.44222 

-.3-.2-.1

0

.1.2.3.4

Experience=5

Unemployment Rate Residuals

 Log-Earnings Residuals Fitted values

-5.44312 3.76533

-.3-.2-.1

0

.1.2.3.4

Experience=8

Unemployment Rate Residuals

 Log-Earnings Residuals Fitted 

-4.19008 

-.3-.2-.1

0

.1.2.3.4
Appendix Figure 3: Effect of Unemployment Rate at Time of Graduation on Log Real Earnings for 

Different Samples: Regional Models, Some College (All) and Graduate Sample, All Cohorts

Panel A: Including Workers Who Stop Filing

Panel B: Estimates for Sample Including Graduate Degrees

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

Main Model: Regional, All

Regional, All

Main Model: Regional, Graduates

Regional, Graduates

-0.025

-0.02

-0.015

-0.01

-0.005

0

0 1 2 3 4 5 6 7 8 9 10

Experience

Log Real Earnings

Without Graduate Degrees (Main Model: All Workers)

With Graduate Degrees (All Workers)

Without Graduate Degrees (Main Model: BA Sample)

With Graduate Degrees (BA Sample)
Main Effect Difference Main Effect Difference

Firm Industry

<75th 

Percentile

>=75th 

Percentile

<75th 

Percentile

>=75th 

Percentile

(1) (2) (3) (4) (5) (6)

Experience Year

0 -0.0091 -0.0091 -0.0146 -0.0021 -0.0157 -0.0025

[0.0033]*** [0.0033]*** [0.0028]*** [0.0022] [0.0028]*** [0.0020]

1 -0.0111 -0.0111 -0.0159 -0.0036 -0.0159 -0.0038

[0.0028]*** [0.0028]*** [0.0026]*** [0.0023] [0.0024]*** [0.0021]*

2 -0.009 -0.009 -0.0134 -0.0053 -0.0136 -0.0058

[0.0027]*** [0.0027]*** [0.0024]*** [0.0022]** [0.0024]*** [0.0021]***

3 -0.0069 -0.0069 -0.0109 -0.0051 -0.0101 -0.0068

[0.0025]*** [0.0025]*** [0.0025]*** [0.0023]** [0.0023]*** [0.0023]***

4 -0.0051 -0.0051 -0.0096 -0.0039 -0.0096 -0.0044

[0.0027]* [0.0027]* [0.0020]*** [0.0019]** [0.0021]*** [0.0018]**

5 -0.0032 -0.0032 -0.0082 -0.0026 -0.0088 -0.0026

[0.0024] [0.0024] [0.0018]*** [0.0016]* [0.0020]*** [0.0017]

6 -0.0031 -0.0031 -0.0067 -0.0043 -0.007 -0.0041

[0.0025] [0.0025] [0.0021]*** [0.0017]*** [0.0020]*** [0.0017]**

7 -0.0027 -0.0027 -0.0052 -0.0058 -0.0069 -0.0039

[0.0027] [0.0027] [0.0019]*** [0.0016]*** [0.0019]*** [0.0014]***

8 -0.0005 -0.0005 -0.005 -0.0033 -0.0073 -0.0006

[0.0023] [0.0023] [0.0020]** [0.0018]* [0.0021]*** [0.0016]

9 -0.0013 -0.0013 -0.0036 -0.006 -0.0061 -0.0013

[0.0027] [0.0027] [0.0023] [0.0017]*** [0.0022]*** [0.0020]

10 0.0028 0.0028 -0.0011 -0.0049 -0.0022 -0.0034

[0.0030] [0.0030] [0.0026] [0.0016]*** [0.0025] [0.0021]

Constant 15.3696 15.3696 8.9546 -- 8.8768 --

[.] [.] [0.0908]*** -- [0.0664]*** --

N 418600 418600 12700 -- 14614 --

R-squared 0.8 0.8 0.93 -- 0.93 --

By Average Median Firm 

Wage

By Average Log Firm Payroll

Appendix Table 6: Effect of Unemployment Rate at time of Graduation on Log Real Earnings 

Controlling for Fixed Effects for First Industry or First Firm

Specification

Fixed Effects for First 

Firm/ Industry

Note: First two columns indicate models with firm or industry fixed effects. The remainign columns display

coefficients from two interacted regression models, respectively. Each columns shows the unemployment rate and

experience interactions from regressing log annual earnings on the youth unemployment rate in the youth

unemployment rate in the province of first residence, interacted with experience years 0 to 10, plus province of

residence fixed effects, experience fixed effects, and year of graduation fixed effects. One, two, and three asterix

indicates statistical significance at the 10 percent, 5 percent, and 1 percent levels respectively. See text for more

details.
Appendix Figure 4: Effect of Graduating in Recession on Annual Earnings by Major of 

Study (Graduates Only)

-0.03

-0.02

-0.01

0

0.01

0.02

0 1 2 3 4 5 6 7 8 9 10

Potential Experience

Log Annual Earnings

All Majors

Arts and Humanities

Social Sciences

Business and Economics

Engineering and Computer Science

Physical Sciences
Years Until 

BA

Fraction 

Above 

Grade

Fraction < 4 

Years

Fraction > 4 

Years

In Graduate 

Sample

Difference 

(D)

Panel A: National, All Workers

0.007 -0.0019 -0.0018 0.001 0.0012 -0.0006

[0.0138] [0.0039] [0.0039] [0.0025] [0.0043] [0.0157]

N 1591 1591 1591 1591 1591 1591

R2 0.01 0.02 0.01 0.02 0.01 0.02

Panel B: Regional, All Workers

0.0072 0.0046 0.0003 0.0041 -0.0032 0.0034

[0.0074] [0.0028] [0.0024] [0.0020]** [0.0028] [0.0108]

N 1591 1591 1591 1591 1591 1591

R2 0.08 0.05 0.06 0.09 0.09 0.06

Panel C: Regional, Predicted UR, All Workers

0.0001 -0.0003 0.0019 0.0025 -0.0048 -0.0042

[0.0410] [0.0101] [0.0115] [0.0081] [0.0112] [0.0523]

N 1566 1566 1566 1566 1566 1566

R2 0.14 0.07 0.15 0.1 0.16 0.11

Panel D: National, D>=0

0.0062 -0.0025 -0.0012 0.0017 0.0001 -0.0052

[0.0063] [0.0043] [0.0010] [0.0025] [0.0014] [0.0082]

N 955 955 955 955 955 955

R2 0.01 0.02 0.04 0.01 0 0.01

Panel E: Regional, D>=0

0.011 0.0083 -0.0002 0.0061 -0.0007 0.0157

[0.0052]** [0.0035]** [0.0009] [0.0027]** [0.0015] [0.0065]**

N 955 955 955 955 955 955

R2 0.22 0.04 0.27 0.15 0.38 0.04

Panel F: Regional, Predicted UR, D>=0

0.006 0.0024 -0.0001 0.0042 -0.0002 0

[0.0042] [0.0038] [0.0008] [0.0026] [0.0016] [0.0000]***

N 930 930 930 930 930 930

R2 0.83 0.64 0.46 0.71 0.54 1

Unemployment Rate

Appendix Table 7: Effect of Unemployment Rate on Duration of College -- National, Regional, and 

Predicted

Unemployment Rate

Unemployment Rate

Unemployment Rate

Unemployment Rate

Unemployment Rate

Note: The sample includes males in Canada leaving university between 1976 and 1995. 'D' indicates the difference between the

actual year left and the predicted year of graduation based on year of entry and program. The dependent variable is indicated

in the column heading. The national model regresses the dependent variable on the youth unemployment rate in the country

at the year of college exit, plus province of residence fixed effects, and a linear or quadratic graduation cohort trend. The

regional model regresses log annual earnings on the youth unemployment rate in the province of first residence, plus province

of residence fixed effects, and year of graduation fixed effects. One, two, and three asterix indicates statistical significance at

the 10 percent, 5 percent, and 1 percent levels respectively. See text for more details.
Model

D>=0? No Yes No Yes

(1) (2) (3) (4)

First Stage Coefficient --- --- 0.8841 0.8984

--- --- [0.0502]*** [0.0391]***

Experience Year

0 -0.0119 -0.0134 -0.0162 -0.0186

[0.0023]*** [0.0023]*** [0.0030]*** [0.0034]***

1 -0.0154 -0.0134 -0.0215 -0.0179

[0.0030]*** [0.0024]*** [0.0041]*** [0.0033]***

2 -0.0145 -0.0114 -0.0204 -0.0147

[0.0030]*** [0.0021]*** [0.0042]*** [0.0028]***

3 -0.0117 -0.0086 -0.0165 -0.0106

[0.0027]*** [0.0019]*** [0.0038]*** [0.0024]***

4 -0.0093 -0.0072 -0.013 -0.0086

[0.0025]*** [0.0019]*** [0.0035]*** [0.0023]***

5 -0.0068 -0.0059 -0.0093 -0.0069

[0.0024]*** [0.0017]*** [0.0034]*** [0.0021]***

6 -0.0054 -0.0045 -0.0072 -0.0053

[0.0027]** [0.0019]** [0.0038]* [0.0024]**

7 -0.0059 -0.0046 -0.0079 -0.0058

[0.0026]** [0.0018]** [0.0036]** [0.0023]**

8 -0.0053 -0.0045 -0.0073 -0.0061

[0.0024]** [0.0018]** [0.0034]** [0.0023]***

9 -0.0046 -0.0041 -0.0065 -0.0056

[0.0024]* [0.0020]** [0.0034]* [0.0024]**

10 -0.0027 -0.003 -0.0043 -0.0044

[0.0025] [0.0020] [0.0034] [0.0024]*

Constant 6.9933 8.7117 7.0555 8.7857

[0.1012]*** [0.0668]*** [0.0981]*** [0.1075]***

N 14223 8495 14223 8495

R-squared 0.92 0.95 0.92 0.95

Reduced Form Instrumental Variables

Specification

Appendix Table 8: Effect of Unemployment Rate at Time of Predicted Graduation on Log Real Earnings 

by Potential Experience (Reduced Form) and Instrumental Variable Estimates, Regional Model 

Note: The sample includes males in Canada leaving university between 1976 and 1995. 'D' indicates the difference between

the actual year left and the predicted year of graduation based on year of entry and program. The reduced form model

regresses log annual earnings on the predicted youth unemployment rate in the province of first residence when D=0,

interacted with experience years 0 to 10, plus province of residence fixed effects, experience fixed effects, and year of

graduation fixed effects. The instrumental variable model regresses log annual earnings on the instrumented youth

unemployment rate in the province of first residence, interacted with experience years 0 to 10, plus province of residence

fixed effects, experience fixed effects, and year of graduation fixed effects. One, two, and three asterix indicates statistical

significance at the 10 percent, 5 percent, and 1 percent levels respectively. See text for more details.