Abstract
This paper adapts the population balancing equation to develop a framework for studying the proximate determinants of educational homogamy. Using data from the National Longitudinal Survey of Youth on a cohort of women born between 1957 and 1964, we decompose the odds of homogamy in prevailing marriages into four proximate determinants: (1) first marriages, (2) first and later marital dissolutions, (3) remarriages, and (4) educational attainment after marriage. The odds of homogamy among new first marriages are lower than among prevailing marriages, but not because of selective marital dissolution, remarriage, and educational attainment after marriage, as has been speculated. Prevailing marriages are more likely to be educationally homogamous than new first marriages because of the accumulation of homogamous first marriages in the stock of marriages. First marriages overwhelmingly account for the odds of homogamy in prevailing marriages in this cohort. Marital dissolutions, remarriages, and educational upgrades after marriage have relatively small and offsetting effects. Our results suggest that, despite the high prevalence of divorce, remarriage, and continued schooling after marriage in the United States, the key to understanding trends in educational homogamy lies primarily in variation in assortative mating into first marriage.
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Notes
Our education classification captures the main education credential groupings in the United States and avoids categories with very small sample sizes. The odds of homogamy in the stocks and flows are quite similar using three-category (<12, 12, >12) and five-category (<10, 10–11, 12, 13–15, ≥16) education groupings.
Figure 1 summarizes the states that women may be in at any given age, but because we are interested in couples’ joint marital and educational characteristics, women may actually be in 1 of 40 states at any given time. When respondents are single (or divorced, separated, or widowed), they may be in 1 of 4 education categories; when they are first married (or remarried), they may be in 1 of 16 joint education categories. This means that a woman may be in one of 40 ([4 × 2] + [16 × 2]) states at any given age. Thus, our multistate life tables consists of a series of transition matrices (one for each year of wife’s age between 16 and 41) that are each 40 × 40, although not all transitions are possible (e.g., educational downgrades).
There is weak evidence for the statistical significance of age patterns of homogamy among prevailing marriages, but stronger evidence for the significance of these patterns among new first marriages (see Online Resource 1 for details). That there is weak evidence for the significance of age patterns among prevailing marriages is not problematic for our analysis because we are interested in how the flows shift the odds of homogamy upward or downward among a wide cross section of marriages, not how these flows contribute to changes in resemblance by wife’s age.
Because the percentage contribution of each flow is estimated from counterfactual marriage distributions using multistate life table data rather than sample data, traditional confidence intervals are invalid. Thus, we bootstrap 95% confidence intervals by taking 1,000 samples of female-respondent person-years with replacement of size n = 81,589 (the total sample size upon which the rates for the multistate life table are constructed). All analyses are run using these 1,000 samples; 95% confidence intervals are calculated as , where is the original sample estimate and \( s{{\widehat{e}}_{{1,000}}} \) is the bootstrapped estimate of the standard error of from 1,000 samples (Efron and Tibshirani 1993:168–173).
The odds of homogamy among couples about to dissolve their first marriages are 84% those among prevailing marriages (2.29 / 2.74 = 0.84, Fig. 4), an estimate consistent with previous findings (Schwartz 2010a). The low volume of marital dissolutions relative to first marriages dampens the effects of these transitions on homogamy in the stock of marriages.
Based on the results shown in Fig. 4, we expected the impacts of later marital dissolutions and educational upgrades to be positive. If we omit the controls for changes in the distribution of educational attainment by wife’s age from Eq. 2, the impacts of both of these flows are positive and small (<0.5%), indicating that these results are sensitive to model specifications.
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Acknowledgments
The first author gratefully acknowledges financial support from the Jacob K. Javits Fellowship Program. The second author received support from the John D. and Catherine T. MacArthur Foundation and the Council on Research of the UCLA Academic Senate. This research was carried out using the facilities of the Center for Demography and Ecology at the University of Wisconsin–Madison (R24 HD047873) and the California Center for Population Research at the University of California, Los Angeles (R24 HD041022). The authors are grateful to Thomas DiPrete, Shoshana Grossbard, V. Joseph Hotz, Jenna Nobles, Judith Seltzer, Megan Sweeney, Donald Treiman, and anonymous reviewers for valuable comments. Earlier versions were presented at the 2001 annual meeting of the Population Association of America in Washington, DC, and the 2003 meeting of the International Sociological Association Research Committee on Social Stratification and Mobility (RC28) in New York.
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Schwartz, C.R., Mare, R.D. The Proximate Determinants of Educational Homogamy: The Effects of First Marriage, Marital Dissolution, Remarriage, and Educational Upgrading. Demography 49, 629–650 (2012). https://doi.org/10.1007/s13524-012-0093-0
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DOI: https://doi.org/10.1007/s13524-012-0093-0