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Abortion Costs, Separation, and Non-marital Childbearing

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Abstract

How do abortion costs affect non-marital childbearing? While greater access to abortion has the first-order effect of reducing childbearing among pregnant women, it could nonetheless lead to unintended consequences through effects on marriage market norms. Single motherhood could rise if low-cost abortion makes it easier for men to avoid marriage. This study estimated the effect of abortion costs on separation, cohabitation and marriage following a birth by exploiting miscarriage and changes in state abortion laws. There is evidence that norms responded to abortion laws as women who gave birth under abortion restrictions experienced sizable decreases in single motherhood and increased cohabitation rates. The results underscore the importance of norms regulating relationship dynamics in explaining high levels of non-marital childbearing and single motherhood.

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Notes

  1. “Shot-gun” marriage refers to the marriage of a couple after pregnancy but before the birth of a child conceived outside of marriage. The shot-gun refers to coercion to ensure the man follows through with marriage.

  2. See Weiss and Willis (1997) who examined patterns in the US; see Weiss (1993) for a review. Brien et al. (1999) showed family childbearing and separation hazards share common unobservables, which is addressed below.

  3. Throughout the study we mainly followed the insights of Ashcraft et al. (2013) and Hotz et al. (2005), who showed that using OLS and IV estimators can deliver bounds on the effects of birth on labor market outcomes for conditionally random miscarriage. Akerlof et al. (1996) and Kane and Staiger (1996) provided models of this information flow, which empirically leads to a simultaneity bias when one conditions on birth.

  4. Horner (2014) showed women with children see lower levels of happiness when divorce barriers are reduced.

  5. Recent work by Beauchamp et al. (2015) showed men are more less likely to pay child support once they match with another partner.

  6. This is the assumption put forth in Hotz et al. (2005).

  7. Since the focus is on separation, the data only include miscarriages prior to twenty weeks of gestation in the empirical section. Results were largely insensitive to this cut-off. Gold et al. (2010) show a substantially higher risk of separation following a stillbirth (>20 weeks gestation) than a miscarriage.

  8. OLS and IV estimates are sufficient to sign the effect since the average treatment effect takes the following form:

    $$\begin{aligned} ATE = \left(\alpha \rho _{OLS} + (1-\alpha ) \beta \rho _{IV}\right)/\left(\alpha + (1-\alpha ) \beta \right). \end{aligned}$$
    (2)

    where (\(\alpha ,\beta\)) are both non-zero probabilities so the true ATE must lie between \(\rho _{OLS}\) and \(\rho _{IV}\). To calculate (\(\alpha ,\beta\)) requires more moments namely, (1) the fraction of women who would give birth if they did not miscarry, (2) the fraction of women who would have a miscarried had they not aborted, and (3) the fraction of all women not giving birth (who either miscarry or abort) who abort.

  9. Stillbirths have been documented to have larger influence on a couples’ likelihood of separation. See Gold et al. (2010).

  10. CDC data are drawn from 2000 and age-specific rates come only from 46 reporting areas in the US calculations come from Table 4 of the CDC report. The age of the Add Health Sample is roughly half 15–19 and half 20–24 in the pregnancy year. 75 % of pregnancies happened in 1997–2001.

  11. Separation results were nearly identical when using 9–24 months from conception as cutoffs. Respondents were asked to combine all periods of on-again off-again sexual intercourse with the partner so that separation measures the end of all sexual contact between the former partners.

  12. One cannot pin down the exact time of policy enactment because the data do not contain the state where the pregnancy occurred if a respondent moved states, a point returned to in the robustness exercise.

  13. Results below strengthen when the income threshold is reduced, and the median is admittedly arbitrary. See Medoff (2007) for a review of how these restrictions reduced abortion demand.

  14. Estimates of the policies’ association with birth, available upon request from the author, looked similar those from Kane and Staiger (1996), with increased abortion costs reducing the probability of births.

  15. Results, available upon request, showed very similar estimates from probit models; linear models were used throughout for consistency with the IV results below.

  16. AHPVT is an abbreviated Peabody Picture Vocabulary test, measuring vocabulary and verbal cognition.

  17. This last note is speculative since the clustering at the state level may be responsible for interpreting the coefficient as significant. Using the robust standard error calculation increased standard errors so that only the triple-interaction appeared significant.

  18. This strategy is sensible only for funding, because poor women are usually linked to their home-state address, even if they travel out-of-state for abortions, through Medicaid funding.

  19. The size of the coefficients on the triple interaction are larger than one, but the effect is given by adding this to other (positive) interaction terms.

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Authors and Affiliations

  1. Department of Economics, Wright State University, 3640 Colonel Glenn Hwy., Dayton, OH, 45435, USA

    Andrew Beauchamp

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Correspondence to Andrew Beauchamp.

Appendix

Appendix

The following sets of controls are discussed above.

Individual and Partner Characteristics

Female and partner age at pregnancy resolution; education level at pregnancy: less than a high school diploma, high school diploma, some college, bachelors degree or more and indicator of partner currently enrolled at time of pregnancy; male and female religious attendance in year of pregnancy (six values: 1 = never–6 = more than once per week) and its square; indicators for no religious attendance for men and women and unknown partner religious information; indicators of Black non-Hispanic, Hispanic, and Other; Welfare Recipient in year of pregnancy, and year prior to pregnancy; Work status (majority of time in pregnancy year) part time or full time; Total years work experience before pregnancy; cohabitation during pregnancy; indicators drinking alcohol daily during pregnancy; indicator for smoking one pack per day or more during pregnancy; exercise intensity at Wave I (none, moderate, intensive), age at first intercourse and its square, weight at Wave I and its square.

County Level Controls

Income: 1990 Census county per capita and median income; Population: 1990 Census population level, density, census designated percent urban; Religiosity: county percent adherents, percent adherents and percent population in conservative and liberal denominations, and proportion Catholic, from Churches and Church Membership 1990 data; Voting: county percent voting Republican and Democrat in 1992 presidential election; Marriage: census fraction of males never married, county level.

Fixed Effects

Fixed effects for thirty states and years from 1995 to 2002 are included.

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Beauchamp, A. Abortion Costs, Separation, and Non-marital Childbearing. J Fam Econ Iss 37, 182–196 (2016). https://doi.org/10.1007/s10834-015-9473-0

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