Abstract
This paper analyzes the family structure experiences of children in the U.S. Childbearing and transitions among single, cohabiting, and married states are analyzed jointly. A novel contribution is to distinguish men by their relationship to children: biological father or stepfather. The analysis uses data from the NLSY79. A key finding is that children of black mothers spend on average only 33% of their childhood living with the biological father and mother, compared to 74% for children of white mothers. The two most important proximate demographic determinants of the large racial gap are the much higher propensity of black women to conceive children outside of a union, and the lower rate of “shotgun” unions for blacks compared to whites. Another notable finding is that cohabitation plays a negligible role in the family structure experiences of children of white mothers, and even for children of black mothers accounts for less than one fifth of time spent living with both biological parents.
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Well-known examples include McLanahan and Sandefur (1994), Chase-Lansdale et al. (1995), and Hetherington and Stanley-Hagan (1999). Recent analyses include Aughinbaugh et al. (2005), Gennetian (2005), Ginther and Pollak (2004), Hofferth (2006), Lang and Zagorsky (2001), and Sigle-Rushton et al. (2005).
See Andersson (2002) for an interesting descriptive comparison of the family structure experiences of children in 15 countries, including the U.S. His findings show that “The USA stands out as one extreme case with its very high proportion of children born to a lone mother, with a higher probability of children who experience a union disruption of their parents than anywhere else, and with many children having the experience of living in a stepfamily.” (p. 343). Heuveline et al. (2003) also report that the U.S. is an outlier in the sense that over half of all time spent by American children in single parent families is accounted for by children born to lone mothers. This is true of only three countries in their 17-country sample.
Here and throughout the paper, whites and blacks include only non-Hispanics. Hispanics can be either white or black (or another race). To save space, we focus mostly on children of black and white mothers. Results for Hispanics are almost always in between those for whites and blacks, and closer to whites. See the working paper version (Blau and van der Klaauw 2007) for a more detailed discussion of results for Hispanics.
Lillard’s (1993) continuous time model of simultaneous hazards is in principle an attractive framework for the analysis. However, it is very difficult to deal with missing data in Lillard’s model. We discuss below how missing data are handled.
The assumption that at most one event occurs per month implies that a woman cannot end a co-residential relationship with one man and begin a co-residential relationship with another man in the same month. A new co-residential relationship can be formed only if the woman is single at the beginning of the month. And a newly formed co-residential relationship cannot be dissolved in the period in which it was formed. Hence, we assume that all partnerships last at least 1 month, and all single spells last at least 1 month.
Ignoring pregnancies that do not lead to a live birth results in some women being treated as if they were at risk of conception in some months in which they actually are pregnant. And the occurrence of a miscarriage or abortion could affect the demographic processes of interest. Accounting for pregnancies that do not result in a live birth would require modeling at least two additional events—miscarriage and abortion—in an analysis that is already quite rich.
Note that this assumption does not rule out a woman having multiple children with a given man with whom she does not live. Such behavior is ruled out only if the woman bears a child with a different man between pregnancies with the man in question.
Note that the new man this period could be different from the new man in previous periods. There is no way to determine this empirically, because we know nothing about non-co-residential romantic relationships unless and until they become co-residential. For example, we do not know how long a woman may have been dating a given man. This could be relevant in determining the risk that he fathers a child or enters a union with the woman. See Carlson et al. (2004) for an analysis of this issue using data from the Fragile Families study.
Because we are interested in the implications of marital behavior for the family structure experiences of children, the end of a marriage is defined to occur at the time of separation rather than divorce. However, some separations are temporary, and modeling the process that determines whether a separated couple reunites would make the analysis overly complex. Therefore, we ignore separations that result in reuniting rather than divorce, if the temporary separation lasts less than 2 years. If a temporary separation lasts more than 2 years, we censor the observation at the date of separation. The only exception is if the woman had not conceived any children before the end of the separation. We also allow for transitory separations involving cohabitation, again if the separation lasts no more than 2 years. The date of separation was not ascertained for marriages that ended in divorce before the 1979 interview. We use the date of divorce as the ending date of the marriage in these cases (2.4% of first marriages). In the sample of 4,926 women, there were 1,676 separations, of which 19% were temporary. The median duration of a temporary separation is 17 months, and 60% were shorter than 2 years and therefore ignored. The other 40% were right-censored.
Sixty percent of freestanding cohabitations had a beginning date that was not known to the nearest month, and 95% had an uncertain end date. Forty percent of cohabitations that turned into marriages had an uncertain begin date (the end date in this case is the date of marriage, which is always known to the nearest month). Another consequence of the fact that some cohabitation dates are not known to the exact month is that there are cases in which the sequence of events cannot be determined. For example, if a cohabitation began sometime between the 1986 and 1987 interview dates and a birth also occurred in this interval, then we do not know whether the woman was single or cohabiting at the time of the birth. Ambiguity about the sequence of events occurs at least once for 9% of women in the sample. Rather than discarding such cases, we modify the likelihood function to account for all the possible sequences in which events could have occurred, weighted by the probability of each sequence.
Sixty percent of live births had an observed conception date. For these cases in which the date of conception is known, the birth occurred 9 months after the conception in 55% of cases, 8 months after in 30% of cases, and 10 months after in 7% of cases. Child deaths are treated as exogenous censoring events, and the spell in progress at the time a child died is right-censored. A new spell then begins with one less child present.
There are 415 cases in which there were important inconsistencies in the union history that could not be resolved. Cases that violate the assumptions of the model include women who (1) end a marriage or cohabitation with a man, and subsequently form a new co-residential union with the same man (except cases of temporary separation, as discussed above); (2) women who bear a child with one man, then with a second man, and then with the first man; and (3) cases in which two or more demographic events occur in the same month (e.g. marriage and birth; conception and exit from cohabitation). There were 114 cases in which a woman apparently violated assumption (1) and otherwise would have been kept in the sample; 68 cases in which a woman apparently violated assumption (2); and 65 cases in which assumption (3) was apparently violated. We say “apparently” because it is likely that some of these cases are a result of errors in identifying men, but there is no way to determine this. A total of 446 cases with inconsistencies in the union history or violations of model assumptions were dropped.
As discussed above, cases in which the sequence of events is uncertain are included in the analysis. They contribute terms in the likelihood function for all possible sequences, weighted by the estimated probability from the model of each alternative sequence. The descriptive statistics presented in Tables 2 and 3 weight each alternative possible sequence equally.
Data from the early 1980s show that 4 and 13% of births were to cohabiting women, for whites and blacks, respectively (Bumpass and Lu 2000). The comparable figures for the early 1990s were 9 and 16, respectively.
The sample includes women up to age 45, but the average age at the last observation is about 39. The simulations are truncated at age 39 in order to allow a comparison to the data. See the working paper version (Blau and van der Klaauw 2007) for a comparison of the simulations to the data. As in the data, some children are not observed for their entire childhood in the simulations. The simulated data for children are truncated at age 18, as were the real data. In the simulations, there are no deaths or children who move in or out of the mother’s household. The only exogenous explanatory variable other than race/ethnicity is the woman’s date of birth, which is set to the sample mean for all women in the simulations.
It is implicit here and throughout the remaining discussion that the child lives with the biological mother. Hence, “time spent with the biological father” means “time spent with the biological father and mother together.”
Marital status at the time of conception plays a dominant role in family structure experiences for children of black mothers, but much less so for children of white mothers. If the mother was single when the child was conceived, the mean percent of childhood spent with the biological father is 46% for whites and 14% for blacks (not shown).
See Blau and van der Klaauw (2007) for details on these calculations.
These numbers are derived from calculations for all simulated children, regardless of the last age at which they are observed.
The figures in Table 5 may underestimate the true number of family structure changes, as our analysis ignores temporary separations and misses cohabitations that begin and end between interviews before 2000, and cohabitations that ended before the first interview.
The categories are lived with (1) biological mother and biological father; (2) biological mother and another man; (3) biological mother and no man; (4) another woman and biological father; (5) no woman and biological father; and (6) any other living arrangement.
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Acknowledgements
Financial support from NICHD grant HD45587 is gratefully acknowledged. Thanks to Karin Gleiter for expert programming. A previous version of this paper was presented at the 2005 Annual Meeting of the Population Association of America in Philadelphia, and in seminars at the Carolina Population Center, Cornell, Syracuse, NYU, and the 2005 NIH Workshop on Intergenerational Family Resource Allocation. We are grateful for comments by the editor, referees, and seminar and conference participants. The authors alone are responsible for the contents. The views expressed are those of the authors and do not necessarily reflect those of the Federal Reserve Bank of New York.
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Blau, D.M., van der Klaauw, W. A demographic analysis of the family structure experiences of children in the United States. Rev Econ Household 6, 193–221 (2008). https://doi.org/10.1007/s11150-008-9030-9
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DOI: https://doi.org/10.1007/s11150-008-9030-9