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Maternity leave and children’s cognitive and behavioral development


We investigate the impact of maternity leave on the cognitive and behavioral development of children at ages 4 and 5, following up previous research on these children at younger ages. The impact is identified by legislated increases in the duration of maternity leave in Canada, which significantly increased the amount of first-year maternal care. Our results indicate no positive effect on indices of children’s cognitive and behavioral development. We uncover a small negative impact on PPVT scores.

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  1. 1.

    The “findings” preamble to the American Family and Medical leave Act states “…it is important for the development of children and the family unit that fathers and mothers be able to participate in early childrearing…” (Public Law No. 103-3 §2(a)(2). A recent Australian paid parental leave program was promoted with the claim that “the scheme will give more babies the best start in life. The payment will enable more parents to stay at home to care for their baby full-time during the vital early months of social, cognitive and physical development” (Commonwealth of Australia 2009). An extension of paid maternity leave in the United Kingdom seeks to “…give children the best start in life…” as the “…evidence confirms the value of consistent one-to-one care in the first year of a child’s life.” (Employment Relations Directorate 2006, p. 2).

  2. 2.

    See Baker and Milligan (2008b, 2010) and Hanratty and Trzcinski (2009).

  3. 3.

    For example, Ruhm (2004) reports reductions in PPVT scores of 7–8 % of a standard deviation from maternal employment in the first year while Bernal (2008) reports that a full year of full-time maternal employment in the first 5 years of life reduces test scores by 0.13 of standard deviation (PPVT and the Peabody Individual Achievement Tests). This research is reviewed in Lucas-Thompson et al. (2010).

  4. 4.

    For example, O’Brien Caughy et al. (1994) report that entrance into daycare before the first birthday was associated with higher test scores (Peabody Individual Achievement Tests) for lower income children and lower test scores for higher income children. For the UK, Gregg et al. (2005) find that children who receive informal care from friends and relatives in the first 18 months of life combined with full time maternal employment have lower cognitive outcomes. In the Canadian context, Lefebvre et al. (2008) report that Quebec’s universal, low fee childcare program, which serves children from birth, is related to reductions in PPVT scores of just under one-third of a standard deviation. Finally, Loeb et al. (2007) find that entry into non-parental center based care before the age of one can lead to problem behavior. Magnuson et al. (2007), Baker et al. (2008), and Datta Gupta and Simonsen (2010), and the research summarized in Belsky (2006) provide further evidence that non-parental care can have negative behavioral effects in some contexts.

  5. 5.

    Historically, mothers have taken the vast majority of the leave, although this is (slowly) changing in recent years. Marshall (2008) reports that in 2006, 23 % of eligible fathers took some parental leave.

  6. 6.

    The changes in provincial mandates were from 29–35 to 52–54 weeks with the exceptions of Alberta, where the change was from 18 to 52 weeks, and Quebec, where the entitlement did not change from a level of 70 weeks.

  7. 7.

    Two provinces, Saskatchewan and Alberta, did not change their job protected leave standards until 2001. Unfortunately there are not sufficient observations from these provinces over the 2–6 months of delay to take advantage of this feature of the reform.

  8. 8.

    The scale is: 0—the child has not reached the predimensional level, 1—the child has reached the predimensional level (4-year-old equivalent), 2—the child has reached the unidimensional level (6-year-old equivalent) and 3—the child has reached the bidimensional level (8-year-old equivalent).

  9. 9.

    Public childcare changes contemporaneous with the parental leave reform could also influence child outcomes. We are not aware of any other provinces with childcare changes over this time period.

  10. 10.

    The proportion of children age 0–5 living with two parents in 2000 in the Labour Force Survey is 92.2 %. Single-parenthood is much less prevalent in Canada than the USA.

  11. 11.

    Alberta and Saskatchewan did not change their maternity leave provisions to match the change in the federal EI rules until after December 2000. We therefore also exclude the very small number of children born in Alberta and Saskatchewan in the months between December 2000 and the point when the provincial maternity leave mandate changed a few months later.

  12. 12.

    Age-standardized PPVT scores are available in the NLSCY as provided by Statistics Canada. The age-standardization is based on the scores of respondents in the first five cycles of the Survey. Details are reported in Statistics Canada (n.d.). Age-standardized scores are not provided for the behavioral or other cognitive measures. For these measures we specify a full set of age (in months) and month of birth effects as additional control variables.

  13. 13.

    The substitution of TFY i for T i provides a neat solution for observations with censored values of T i by limiting the time frame to 1 year. For example, for children surveyed at older ages, the observation of T i may be censored because the mother is still at home at the date of the survey.

  14. 14.

    The regressions for age-standardized PPVT scores omit the month of birth and single month of child’s age. The results including these controls are reported in footnotes below. We control for place of birth to account for any differences in immigration patterns. Over this time period, however, there were no strong shifts in the class or source of immigration to Canada. Robust standard errors reported for all estimates. These standard errors are “conservative,” in the sense that the standard errors clustering on province, year of birth or a pre-/post-policy reform indicator are generally smaller.

  15. 15.

    The polynomials in quarter are not included here given the specification of year of birth effects.

  16. 16.

    Adding controls for the child’s age and month of birth the estimate is −0.415 (0.486). As noted below (i.e., footnote 20) this change in the estimate is due to the sensitivity of the estimate for girls to the change in the specification of the age and month of birth effects.

  17. 17.

    The second stage collapses when we specify a quintic in time. The model is not identified when we use quarter of birth dummy variables since they perfectly predict the instrument.

  18. 18.

    The number of observations varies across the different outcomes. Restricting the sample to children with valid observations on all outcomes leads to similar estimates.

  19. 19.

    The exception is hyperactivity—the estimates are negative, quite large and in some specifications statistically significant.

  20. 20.

    Controlling for child’s age and month of birth the estimate for boys is −1.443 (0.743) and for girls, 0.953 (0.672).

  21. 21.

    The estimates for the 8 year of birth effects are reported in Baker and Milligan (2011).

  22. 22.

    Cycles 6–8 are the source of information on the non parental care of post reform children at these older ages. These sampling issues make the comparison of these responses to the responses of pre reform cohorts from earlier waves problematic. The issues include a computer glitch that led to missing values for one-quarter of children in cycle 6.

  23. 23.

    We investigated the potential impact of differential fertility on the composition of our sample by looking at the fraction of childless couples, the average number of children among those with children, and the fraction of those having a child and who are married. No differences in the trend after 2000/2001 are observed.

  24. 24.

    As noted earlier, the year of birth is not directly reported in the LFS. Instead, we identify the year of birth for children by selecting a sample of children in December of each year. The regressions also include controls for province, urban/rural residence, mother’s and spouse’s age and education.

  25. 25.

    We create real earnings by converting the earnings reports to 2002 dollars using the Consumer Price Index. In the presence of a general upward trend in wages across years, our estimates here based on just the time series variation will attribute to the policy what is really just a trend in real wage growth. We have also re-estimated these regressions deflating earnings by the growth in the Industrial Aggregate Wage from Statistics Canada’s Survey of Employment Payrolls and Hours (catalogue 72-002-XIB). Using these wage-growth-adjusted earnings tells a somewhat different story. They indicate a smaller (by half) increase in family earnings at ages 13–35 months, and little increase at older ages. These results are reported in Baker and Milligan (2011).

  26. 26.

    The LFS does not provide any direct information on whether the child is in non-parental care.

  27. 27.

    Our brief discussion of these milestones follows Scher and Harel (2009).


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We have benefited from comments from Josh Angrist and seminar participants at the 2011 AEA meetings, UC Davis, Laval, Manitoba, Stavanger, Texas, W.E. Upjohn and Waterloo. We also thank the referees for helpful suggestions. We gratefully acknowledge the research support of SSHRC (Baker Grant, #410-2008-0346, #410-2011-0724 Milligan Grant #410-2006-0928). We thank the staffs of the Toronto and B.C. Research Data Centres for their technical support. This paper represents the views of the authors and does not necessarily reflect the views of Statistics Canada.

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Correspondence to Michael Baker.

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Baker, M., Milligan, K. Maternity leave and children’s cognitive and behavioral development. J Popul Econ 28, 373–391 (2015).

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  • Maternity leave
  • Child development
  • Children
  • Cognitive development

JEL Classification

  • J13
  • I18
  • J38