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Is there a ‘marriage premium’ for gay men?

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Abstract

It is well-known that married men earn more than comparable single men, with typical estimates of the male marriage premium in the range of 10–20%. Some research also finds that cohabiting men earn more than men not living with a female partner. This study uses data from the General Social Survey and the National Health and Social Life Survey to examine whether a similar premium accrues to gay men who live with a male partner and whether cohabiting gay men have different observable characteristics than non-cohabiting gay men. Controlling for observable characteristics, cohabiting gay men do not earn significantly more than other gay men or more than unmarried heterosexual men. Cohabiting heterosexual men also do not earn more than non-cohabiting heterosexual men.

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Notes

  1. Alm et al. (2000) estimate that legalizing gay marriages would increase federal tax revenues because most homosexual couples consist of two earners and many would be subject to the “marriage tax penalty.” In an examination of whether gay-partnership laws in Europe lowered the prevalence of sexually transmitted diseases, Dee (2007) finds negative effects on syphilis rates (and smaller, less precise effects on gonorrhea and HIV).

  2. Mueller (2007) suggests that men living in same-sex relationships may experience more negative discrimination than other gay men because they are more likely to be openly gay. Also, Frank (2006) argues that marriage serves as a signal of heterosexuality, and discrimination against gay men may explain at least part of the male marriage premium. Blandford (2003) makes a similar point, noting that estimates of the male marriage premium may be confounded by mixing unmarried heterosexual and homosexual men in the comparison group. If gay men experience an earnings penalty due to discrimination based on their sexual orientation, including these men in the unmarried comparison group inflates estimates of the male marriage premium.

  3. Research focusing on females indicates that cohabiting lesbians earn more than women in opposite-sex couples (including married women) but that this earnings premium does not appear to be due to specialization (Jepsen 2007).

  4. Similar to Jepsen and Jepsen’s (2002) results for the US, Andersson et al. (2006) report larger average age differences as well as more differences in nativity among registered same-sex couples than among registered opposite-sex couples in Norway and Sweden. Schoen and Weinick (1993) report that cohabiting opposite-sex couples in the US are more alike in educational attainment than married couples, which they interpret as evidence of less specialization.

  5. However, results in Oppenheimer (2003) and Xie et al. (2003) suggest that selection plays a larger role in marriage than in cohabitation.

  6. There is a smaller literature on heterosexual-gay earnings differences in other countries, including Plug and Berkhout’s (2004) study of the Netherlands and Carpenter’s (2007b) study of Canada.

  7. Arabsheibani et al. (2005) find that cohabiting gay men are paid less than cohabiting heterosexual men with the same characteristics in the UK and interpret this result as due at least in part to discrimination. Another potential contributor to the gay earnings penalty is compensating differentials. Carpenter (2007b) finds that gay men in Canada are more likely to have “good” jobs than heterosexual men; gay men’s jobs may pay less but have better non-pecuniary characteristics.

  8. The 1990 and 2000 Censuses have much larger samples, but because the surveys did not ask about sexual orientation or sexual partners, gays and lesbians can be identified solely on the basis of living with a same-sex partner. This makes it impossible to compare the earnings of cohabiting and non-cohabiting gays. Black et al. (2000) provide a critical review of the quality of the GSS and NHSLS (as well as the 1990 Census) data for examining gays and lesbians.

  9. The sample does not include 11 men who are married to a woman but who appear to be gay in terms of having had exclusively male sex partners during the last year (i.e., closeted gays). Badgett (1995) reports that over 40% of her sample of gay men are married, with gay defined as having had more same-sex partners than opposite-sex partners since age 18. Individuals whose reports of the sex of their partners within the last year are inconsistent with reports of the sex of their partners since age 18 are also dropped from the sample, as are individuals whose own sex is reported once as male and once as female in the GSS.

  10. This method would mistakenly classify heterosexual men with female roommates and gay men with male roommates as cohabiting and would misclassify cohabiters who do not report having a regular sex partner as non-cohabiting. Any such measurement error should bias the coefficients on the cohabitation variables toward zero. The robustness section examines the sensitivity of the results. In Carpenter’s (2005) sample from the 2001 California Health Interview Survey, about one-third of gay men report living with a partner. Black et al. (2000) estimate that about 28% of gay men in the GSS (1988–1996) and NHSLS are cohabiting with a partner. In the sample here, about 30% of gay men are cohabiting with a partner.

  11. Results for indicator variables are interpreted as exponents of the estimated coefficients because the dependent variable is a natural log.

  12. As noted above in the text, the prevalence of gay men in traditionally female occupations may contribute to the gay earnings penalty. The small sample size for gays precludes more detailed analysis of this possibility. The estimated coefficient of the gay indicator variable is −0.124 (0.061) when controls for 5 of 6 broad occupational categories and 8 of 9 industries are included. Only 1 of the cohabiting gay men and 9 of the non-cohabiting gay men report having any children, which is not enough variation to accurately measure any effects of children.

  13. However, an F-test shows that the coefficients shown in column 4 of Table 3 are jointly significantly different from 0 (F-statistic = 3.97, p-value = 0.00).

  14. Education results for the stratified sample of gay men are not monotonically increasing, but only 3 gay men—all non-cohabiting—are not high school graduates.

  15. The sample sizes for columns 1–3 in Table 4 are less than the 4,913 observations in the other tables because the 1988–1990 GSS surveys did not ask the gender of sexual partners during the last 5 years and the 1988 survey did not ask the number of partners of each gender since age 18.

  16. The mean age of these men is 38, which is between the mean ages of cohabiting and non-cohabiting gay men.

  17. Standard techniques for examining the role of selection along unobservable characteristics, such as individual fixed effects or siblings comparisons, cannot be implemented in the GSS/NHSLS data.

  18. The results of either a Small-Hsiao or a Hausman test of the independence of irrelevant alternatives (IIA) assumption required by the multinomial logit model fail to reject the IIA assumption.

  19. However, Loh (1996) find a positive association between wives’ labor market experience and husbands’ earnings.

  20. Such results are also consistent with negative assortative mating and with employer discrimination against men with working wives (Jacobsen and Rayack 1996). Studies typically find that the negative effect of wives’ hours of work on men’s earnings lessens when controlling for selection and endogeneity using instrumental variables techniques (e.g., Blackaby et al. 1998; Jacobsen and Rayack 1996).

  21. The Tobit regressions include the spouse/partner’s age (a quartic) and race/ethnicity (black, other race, and Hispanic) as well as all of the variables included in the earnings regressions. Separate Tobit regressions were estimated for married women, cohabiting women, and gay men. The analysis does not use the presence and age of children (other identifying variables frequently used in the literature) because so few gay male couples have children and because children may not be exogenous with respect to men’s earnings.

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Acknowledgments

The author thanks Kitt Carpenter, Shoshana Grossbard, Lisa Jepsen, and session participants at the 2006 Southern Economic Association meetings for helpful comments on an earlier version.

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  1. Department of Economics, Agnes Scott College, 141 E. College Ave, Decatur, GA, 30030, USA

    Madeline Zavodny

  2. IZA, Bonn, Germany

    Madeline Zavodny

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Correspondence to Madeline Zavodny.

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Zavodny, M. Is there a ‘marriage premium’ for gay men?. Rev Econ Household 6, 369–389 (2008). https://doi.org/10.1007/s11150-007-9022-1

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