Abstract
We examine the impact of the Allowance for Newborn Children, a universal baby bonus offered by the Canadian province of Quebec, on birth order, sibship sex composition, income, and education. We find a large response for third- and higher-order births for which the bonus was more generous. Interestingly, though, we find stronger response if there were two previous sons or a previous son and daughter rather than two previous daughters. We also find, in addition to a transitory effect, a permanent effect, with the greatest increase in one daughter-two son families among three-child households. Moreover, we find a hump shape response by income group, with the greatest response from middle-income families. Also, women with at least some post-secondary education respond more to the policy than those with less. These findings suggest that properly structured pro-natal policies can successfully increase fertility among different segments of the population while simultaneously diminishing the effect of gender preferences and fertility disparity related to women’s education.
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Notes
Unfortunately, the cancelation of the policy is announced well in advance and replaced by universal childcare; this creates a less credible experimental environment at the end of the policy period.
New theoretical models are accounting for observed heterogeneous effects. For example, to account for the effect of a child-care policy on fertility, Yakita (2018) allows for responses to differ by level of maternal education.
Milligan (2005) estimates a probit regression with the variable “family income.” He finds an overall positive coefficient, whereas we subgroup family income and estimate the same model to find the marginal effects of each income subgroup. Here, we are able to find a hump shape response for family income.
The baby bonus was paid to all births that were registered; we find no evidence of differences in ANC take-up rates by income.
We confirm this calculation.
In simulation, if applicable, we assume that the second child is 6 years old and the third child is 10 years old.
In 1997, after our sample period, Ontario introduced a means-tested child care supplement for working parents (Milligan and Stabile 2011).
Source: Statistics Canada. Table 106-9013.
With the confidential data, we are able to look at annual TFR for each year of age, whereas past papers using the public-use data have had to use 5-year age intervals.
From all the Canadian provinces, the province of Ontario is the most comparable to Quebec; they are neighbors, as well as the two most populated provinces in Canada. There are many cities and towns on the border of these two provinces, and in one instance, they even share the same metropolitan area (Ottawa-Gatineau).
See Hotz et al. (1997) for a detailed comparison on total fertility rates (TFR) and completed fertility rates (CFR).
In addition to graphical findings, we estimate a difference-in-differences (DID) model using the TFR as the outcome of interest for Quebec and Ontario with 1995 as the treatment year and 1988 as the comparison year. The DID model results in a 0.11 increase in the number of children born to Quebecois women in the treated year. As Manski and Pepper (2018) point out, such DID estimates require strong assumption on DID invariance. Following Manski and Pepper, we apply a class of the bounded-variation assumptions. We use the data prior to 1988 to calculate the bound parameter of bounded time variation, bounded inter-province variation, and bounded DID variation. The bounded DID estimates are between 0.104 and 0.199. These models are available upon request.
We also compared Quebec to the Rest of Canada and find that it closely mirrors that of Ontario illustrating that the gaps we are observing in Quebec are not just in comparison to Ontario.
Although the universal childcare policy is announced to start at the same time the baby bonus is canceled, no new subsidized childcare spaces were created before 2001 (Haeck et al. 2015). In Norway, Havnes and Mogstad (2011) find that formal childcare acts as a substitute for informal childcare (arrangements with relatives, friends, and so forth) instead of encouraging new female labor force participation. Baker et al. (2008) examine childcare use in Quebec and do find some crowding out of existing arrangements is evident.
Milligan (2002) writes that the rate for third and subsequent births in Quebec increased by 35%, from 0.217 per woman in 1987 to 0.294 in 1993, while falling elsewhere in Canada by 3%.
We also show a transitory effect took place both graphically and in regression analysis. Our graphical results can be seen in the Appendix, Fig. 6. The figure shows the birth cumulative distribution function for each of three cohorts by age of mother and parity, separately for Ontario and Quebec. The difference between Ontario and Quebec is most evident for the third child, where one can observe the “middle cohort” in Quebec having children much earlier than their Ontario counterparts.
The main shortcoming of Milligan’s (2005) study of the ANC is that the public-use census does not provide year of birth. This meant that the ANC policy period overlaps the 1991 census window, making it difficult to disentangle which births are part of the policy period. Also, the public-use census file has a very small sample size and does not allow for a thorough examination of heterogeneous effects like the confidential census file.
As a robustness check we also use a 3-year and 5-year window. See Section 6.2 for more detail.
We limit the sample to women aged 34 to ensure we can identify all children; there is a concern that if the woman is older than 34 years of age she may have children living outside the home.
As a robustness check, we examine the effect of the baby bonus on all single women. See Section 6.2 for more detail.
Another reason we only look at married women is because we do not want to model the relationship between the decision to be married and fertility as studied in Baudin et al. (2015).
We use the Canadian Consumer Price Index (CPI) for each province to convert nominal income into real income in 1992 constant Canadian dollars.
A household is located in an urban dwelling if it is located in a census metropolitan area (CMA), which is one or more municipalities with at least 100,000 people.
In some instances, we also utilized a triple-difference model; however, we prefer the ease of interpretation provided by subsampling the difference-in-differences model. The triple-difference results match well with our preferred model. Results of the triple-difference are available upon request.
Special care is taken into calculating average partial effects instead of partial effects evaluated at the mean. We observe individual’s characteristics to calculate an individual probability and then average all those probabilities, as opposed to mean marginal effects, where the mean for each variable is plugged in to calculate a probability. We calculate the marginal probability using the method described in Ai and Norton (2003).
The approach of using “probability of having a child” as the dependent variable is not new to this literature (see Cohen et al. 2013).
The definition of immigrant in this case comes from the Census definition, which represents all individuals not born as a Canadian citizen.
Nitsche et al. (2018) find evidence that it is important to also account for the male partner’s education level as it also significantly predicts fertility.
We also estimated n = 4 or more children and find that the results are similar to those for n = 3 or more.
We limit our sample to women aged 35 to 39 because they are near the end of childbearing, while still being young enough to have their children living at home. The census only accounts for the number of children present in the household; thus, if we include older women, we may be missing children that are no longer living at home.
This calculation is based on the average marginal effect for the interaction term divided by the proportion of women that had a child in Quebec in our pre-policy period (1987–88), which was 0.207.
The representative female used to calculate the probability of having a child is a married non-immigrant francophone woman who is 30–34 years old, with some post-secondary education, lives in an urban area, and has no previous children. These characteristics are chosen as they are the most common female we encounter in our sample and thus make the most general comparison.
Since younger women are likely to return to school, as a robustness check, we estimate our specification considering only women over 25 years of age and results do not change significantly.
Households’ response to having two children is negative since they are likely moving to a family with three children given the large cash incentive.
The sample size drops to 90,000 households. Also, Quebec has almost four times the number of observations than Ontario. Thus, this is not our preferred specification. The CMAs we selected are Temiskming Shores, North Bay, Petawawa, Pembroke, Hawkesbury, Cornwall, Rouyn-Noranda, Lachute, Salaberry-de-Valleyfield, Val-d’Or, and Amos.
We drop 12 strata because they contain less than 5 observations for each province and each period.
Results available upon request.
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Acknowledgements
We thank Byron Spencer for his guidance and support. We would also like to thank Philip DeCicca, Arthur Sweetman, Laura Turner, participants at the Canadian Population Society Annual Conference, the European Society for Population Economics Annual Conference, the Canadian Economic Association Annual Conference, the Annual Congress of the European Economic Association, and the University of New Brunswick for their helpful suggestions. We would also like to thank the anonymous referees and the editor, Alessandro Cigno, for their detailed and insightful comments. We would also like to thank Peter Kitchen from Statistics Canada for all his help. The analysis presented in this paper was conducted at the Research Data Centre at McMaster which is part of the Canadian Research Data Centre Network (CRDCN). The services and activities provided by the Research Data Centre at McMaster are made possible by the financial or in-kind support of the SSHRC, the CIHR, the CFI, Statistics Canada, and McMaster University. The views expressed in this paper do not necessarily represent the CRDCN’s or that of its partners’.
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Natalie Malak has received grants from Ontario Graduate Scholarships for her doctoral degree. Md Mahbubur Rahman and Terry A. Yip have received support from the Ontario Student Assistance Program for their doctoral degrees.
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Malak, N., Rahman, M.M. & Yip, T.A. Baby bonus, anyone? Examining heterogeneous responses to a pro-natalist policy. J Popul Econ 32, 1205–1246 (2019). https://doi.org/10.1007/s00148-019-00731-y
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DOI: https://doi.org/10.1007/s00148-019-00731-y