Government programs designed to provide income safety nets often restrict eligibility to families with children, creating an unintended fertility incentive. This paper considers whether dramatically changing incentives in the earned income tax credit affect fertility rates in the USA. We use birth certificate data spanning the period 1990 to 1999 to test whether expansions in the credit influenced birthrate among targeted families. While economic theory would predict a positive fertility effect of the program for many eligible women, our results indicate that expanding the credit produced only extremely small reductions in higher order fertility among white women.
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A qualified child is a natural or adopted child or stepchild of taxpayers filing joint or single head of household returns. In 1990, a parent had to provide more than half the support for the child regardless of whether he or she lived with the child. Beginning in 1991, a parent could claim the EITC only if the child lived with him or her for more than half the year. For a thorough introduction to the EITC program, its historical development and provisions, see Hoffman and Seidman (2002).
In 1991, 1992, and 1993, the EITC was greater for filing units with children under 1 year of age. This so-called “wee-tots” credit increased the maximum credit by $388 in 1993.
Beginning in 1998, Minnesota had a two-tiered system where there is a second flat range above the initial one.
We note that there is no evidence that the EITC causes a behavioral response on the internal margin of hours worked (see, for example, Eissa and Liebman 1996).
According to the National Center for Health Statistics, a nurse attending at the birth collects this information from both the mother and father (if present). Contacts at the NCHS have told us that they assume the mother herself to be the source of all information in the certificates, except in a small number of states that impute marital status.
These results are consistent with other similar work. In the Current Population Survey, Eissa and Hoynes (2004) find that in tax year 1996, around 60% of married couples with children where the woman has less than a high school education are eligible for the EITC, and 19% of those where the woman has exactly a high school education are eligible. Using the cohort from the National Longitudinal Survey of Youth 1979 data, Dahl and Lochner (2005) find that 39% of mothers eligible for the EITC are high school dropouts, 48% have a high school degree, and 14% have some college.
The GAO (1996) estimated that 73% of married couples and only 53% of single parent receiving the EITC in 1994 were in the phase-out range.
Note that to be able to estimate populations for cell denominators using a 5% PUMS sample, we combine black and other non-whites into a single non-white category, and we combine 15- to 19- and 20- to 24-year-olds into a single category.
Ideally, women would be assigned to birth groups based upon total completed fertility. We use the number of children living in the household as a proxy for all children born to the mother because the 2000 Census did not ask a total fertility question.
Washington State did not record education on the birth certificate until 1992. In our 10-year sample, Washington and Connecticut average approximately 8 and 37% (respectively) of birth certificates missing information on mother’s education. In 1990, New York and New Jersey average approximately 41% of birth certificates missing data on mother’s education. In most cases, the missing data results because particular geographic regions within the state did not record mother’s education. We also exclude observations without a record of birth order. This is approximately 0.3% of all birth certificates in the nation. Connecticut accounts for a large share of these. We present specification tests that address the decision to drop these missing observations later in the paper.
The mean birthrate for all women in the dataset (including those with higher education) is 6.67 per 100 women, which compares well with national estimates ranging from 70.9 per 1,000 in 1990 to 65.9 per 1,000 in 1999 (US Census Bureau 2001).
As a falsification test, we separately estimate the regressions for samples of second births and third and higher order births. We use the EITC base and incremental values for second births as the primary independent variables for both samples. Because the EITC creates no incentive for third or higher order births, a significant effect would seemingly indicate a spurious relationship between the EITC incremental value variable and fertility. We do not find any convincing evidence of such a spurious relationship. While the coefficient on the second child incremental variable is still negative (−0.092) for third or higher order births, it is not statistically significant at standard levels and is smaller than the coefficient for second births only using the appropriate EITC values (−0.115). However, because neither coefficient is precisely estimated, we cannot reject equality between the two.
Dehejia and Lleras-Muney (2004) find that the health of babies is positively related to the unemployment rate and attribute at least some of this to the selection into motherhood during economic downturns.
In the cases where the EITC variables are zero, we put in 0.00001.
The following states inferred marital status, sometimes very badly, in the vital statistics data for some or all of the years in our data Michigan (all years), New York (all years), Connecticut (before 1998), California (before 1995), Nevada (before 1997), and Texas (before 1994) (US Department of Health and Human Services 2002; Ventura and Bachrach 2000).
It does not seem likely that there are differential labor supply responses between married and unmarried women in this case. An increase in the base EITC should provide incentive for married women to drop out of the labor force, given that most eligible married women are in the phase-out range of the EITC. This should be correlated with an increase in fertility rather than the increase that we see.
The welfare literature often restricts the sample to women with less than a high school diploma, but given the earnings requirements in the EITC, this restriction is not reasonable in our case.
For example, the estimated elasticity of −0.422 combined with a mean birthrate of 6.96 per 100 for higher order births to college-educated white women would imply more than a 2 percentage point reduction in the birthrate in response to the EITC.
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Thanks to Gabrielle Chapman, Cristian Meghea, and Karoline Mortenson for excellent research assistance. We are also grateful to Janet Holtzblatt, Saul Hoffman, Marianne Bitler, Elizabeth Peters, Irv Garfinkel, and anonymous referees for helpful comments and suggestions. Baughman gratefully acknowledges the support of the Robert Wood Johnson Scholars in Health Policy Research Program while the project was being completed. All remaining errors are our own.
Responsible editor: Junsen Zhang
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Baughman, R., Dickert-Conlin, S. The earned income tax credit and fertility. J Popul Econ 22, 537–563 (2009). https://doi.org/10.1007/s00148-007-0177-0
- Earned income tax credit