Her money, her time: Women’s earnings and their housework hours

Social

Social Science Research 35 (2006) 975–999

Science

RESEARCH

www.elsevier.com/locate/ssresearch

Her money, her time: WomenÕs earnings and their housework hours Sanjiv Gupta Department of Sociology and Social and Demographic Research Institute (SADRI), W34A Machmer, University of Massachusetts, 240 Hicks Way Amherst, MA 01003-9278, USA Available online 12 September 2005

Abstract Do husbands and wives divide housework on the basis of who makes more money? Much of the recent literature has focused on the effects of individualsÕ earnings relative to their partnersÕ on their housework. By contrast, this paper analyzes the effects of womenÕs own earnings on the time they spend doing housework in the context of heterosexual couple households. A conservative estimate using the second wave of the National Survey of Families and Households (NSFH) is that the negative association of womenÕs housework with their own earnings is two to three times greater than that with their partnersÕ; in the full model, the association with partnersÕ earnings is not statistically significant. The importance of womenÕs own earnings in housework models is highlighted by the comparable effect of income on housework among single women. It appears that so far as housework is concerned, women do not benefit greatly from their male partnersÕ incomes. The finding emphasizes the gender segregation of domestic labor, and underscores the importance of income differences among women in explaining their housework behavior. It shows that the difference between the mean housework hours of the women with the lowest and highest earnings is as large as the difference between the mean housework hours of women and men.  2005 Elsevier Inc. All rights reserved. Keywords: Housework; Gender display; Relative resources

E-mail address: [email protected] 0049-089X/$ - see front matter  2005 Elsevier Inc. All rights reserved. doi:10.1016/j.ssresearch.2005.07.003

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1. Introduction Do husbands and wives divide housework on the basis of who makes more money? The question cuts to the heart of the economics of heterosexual households and has received considerable scholarly attention (e.g., Bittman et al., 2003; Blair and Lichter, 1991; Brines, 1994; Coverman, 1985; Evertsson and Nermo, 2004; Farkas, 1976; Greenstein, 2000; Ross, 1987). Posing it like this implies that what is important is not how much money each partner in a couple makes, but how much she/he makes compared to the other partner, i.e. their relative earnings. Two variants of this argument have come to dominate the quantitative housework literature. The ‘‘economic exchange’’ or ‘‘relative resources’’ hypothesis proposes that the partner with greater earnings has more bargaining power in couplesÕ negotiations over the division of domestic labor. The second, the ‘‘gender display’’ or ‘‘deviance neutralization’’ hypothesis, suggests that individuals with relative earnings that are unusually high or low for their gender compensate by exaggerating their gender-normative housework performance: men with unusually low relative earnings spend less time on housework than other men, and women with very high relative earnings spend more time on housework than other women. Despite their contrary predictions, both of these theories derive their explanatory power from the notion that housework is affected by the earnings of one partner relative to the otherÕs, or by a partnerÕs proportionate contribution to the coupleÕs total earnings. These models make the important but rarely tested assumption that the unit associations of womenÕs and menÕs earnings with housework are equal, so that what matters is their relative earnings. The analysis presented here uses the National Survey of Families and Households (NSFH) to test this assumption. Focusing on womenÕs housework, this paper analyzes the separate effects of womenÕs own absolute earnings as well as their male partnersÕ. This step, omitted by virtually all studies to date, is crucial to determining the validity of the commonly used measures of relative resources in housework models, such as one partnerÕs share of the coupleÕs total earnings. These measures may not be sound if there is a substantial difference between the separate associations of womenÕs and menÕs earnings on housework time. In this paper I test explicitly for the difference in the coefficients of menÕs and womenÕs incomes in models of womenÕs time spent on housework. As an additional test of the appropriateness of the measures employed by the relative resources and gender display models, I determine the relationship between earnings and housework hours for single women, and compare it to the association for partnered (i.e. married or cohabiting) women. By definition, single women do not have to bargain with or display gender to male partners. Therefore if the relationship between income and housework is really a function of one partnerÕs earnings relative to the otherÕs, as both the exchange and display models suggest, then the relationship between their earnings and time spent on housework for partnered women should be quite different from that for single women. On the other hand, if the relationship between earnings and housework time is similar for the two groups of women, it would suggest that the relative earnings measures employed by the exchange and display models are not very useful in explaining womenÕs housework performance.

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This analysis of the independent effects on womenÕs housework of their own and their partnersÕ absolute earnings has two substantive motivations. First, it seeks to determine whether household labor is a gender-segregated activity not only in terms of the time men and women spend doing it, but also in the degree to which their respective earnings are associated with its performance. Second, this study calls attention to womenÕs absolute incomes as an important source of variation in womenÕs housework. In contrast to the usual focus of the quantitative housework literature on womenÕs economic resources relative to menÕs, the results presented here highlight the importance of economic stratification among women in determining their time spent on domestic labor.

2. Background Two theories have come to dominate the literature on the relationship between earnings and housework, the ‘‘economic exchange’’ and ‘‘gender display’’ frameworks. The first of these has also been referred to as the ‘‘economic dependence’’ or ‘‘relative resources’’ perspective; other terms synonymous with the second include ‘‘doing gender’’ and ‘‘gender deviance neutralization.’’ 2.1. Economic exchange The economic exchange theory of the allocation of domestic labor in couple households posits that the more money one partner makes relative to the other, the less housework she/he does. The notion that there is an inverse relationship between labor market performance and housework has a long intellectual lineage. In an application of DurkheimÕs concept of organic solidarity, the early functionalists argued that households are maintained through an agreement between husbands and wives to specialize in paid and unpaid work, respectively. (See Lopata (1993) for a succinct account of these theories.) In a similar spirit, Becker (1991) formulated a model in which partners specialize in home or market production based on their productivity in each sector, with the aim of maximizing output for the household as a whole. Becker argued that the gender gap in housework is a consequence of the voluntary adaptation of husbands and wives to menÕs higher labor market wages and womenÕs biological advantages in childrearing. More recent feminist and social structural perspectives have challenged the functionalist and new home economics accounts. They view couple households as sites of contention and bargaining between the two partners on questions such as the division of domestic labor. The outcome of the bargaining process is a function of the relative economic resources of the partners. If both partners view housework as an unpleasant necessity, the person with fewer resources will do more of it, because she/he has more to lose if the union dissolves (England and Farkas, 1986). Because men typically have an advantage with respect to economic resources, they have greater power in the negotiation over the division of household labor (Blumberg and Coleman, 1989; Huber and Spitze, 1983). Game-theoretic approaches in

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economics also treat the performance of housework as the result of a bargaining process in which labor market earnings are the currency of negotiation. (See Bittman et al., 2003 for a discussion.) 2.2. Gender display Whether we view the division of domestic labor as a consensual or contested process, the central prediction of economic exchange theory is the same: the more money individuals make relative to their partners working outside the home, the less work they do inside it. This prediction is gender neutral; the inverse relationship between relative earnings and housework obtains for both men and women. By contrast, some researchers have theorized gender asymmetries in the relationship. They argue that the compulsion to affirm gender identity is so strong that individuals will violate the logic of economic exchange in the face of gender-atypical economic circumstances. For example, women who earn more than their husbands will compensate for the resulting threat to their partnersÕ masculine identity by doing more, not less, housework than women who earn less than their partners. This view of housework as a demonstration of gender identity was first articulated by West and Zimmerman (1987). They argued that men and women enact their gender through their daily activities, especially in situations involving interactions with individuals of the opposite gender. Several studies have since explored the empirical implications of this idea for the division of housework in heterosexual coresidential unions (Bittman et al., 2003; Brines, 1994; Evertsson and Nermo, 2004; Gupta, 1999; South and Spitze, 1994). In one of the first tests of the gender display hypothesis, Brines (1994) used the Panel Study of Income Dynamics (PSID) to reveal a gender asymmetry in housework responses to relative earnings. While womenÕs housework time did vary inversely and linearly with their relative earnings, that of men was described by an upside-down parabola: husbands who contributed less than their wives to the coupleÕs income did less housework than husbands who contributed equally. Brines concluded that these husbands were practicing gender display. Because their economic situation was gender-atypical, they did less housework than other husbands in order to affirm their normative gender identities. Subsequent attempts to test the gender display hypothesis have yielded conflicting results, especially for women. The recent history of research on the relationship between womenÕs relative earnings and their housework is summarized in Table 1. Though Brines (1994) did not find evidence of gender display in that relationship for women, Evertsson and Nermo (2004) reported a gender display pattern in multiple waves of the same longitudinal survey that Brines used (though not in the same year). However, there is no such tendency in the relationship between relative earnings and housework among Swedish women, whose behavior is consistent with the exchange model. In another test using the first wave of the survey employed in the present analysis, Greenstein (2000) reported a gender ‘‘deviance neutralization’’ pattern in the effect of womenÕs share of couplesÕ earnings on their share of total housework. But Bittman et al. (2003), using the same data, did not find this curvilinearity for women. In addition, they showed that Australian womenÕs housework followed

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Table 1 Evidence for the economic exchange (or relative resources) and gender display (or deviance neutralization) perspectives in recent research using relative earnings as predictor of housework hours Study

Women

Men

Data

Bittman et al. (2003)

US: exchange

US: display

Australia: both

Australia: neither

National Survey of Families and Households, 1987–1988 Australian Time-Use Survey, 1992

Brines (1994)

US: exchange

US: display

Panel Study of Income Dynamics, 1985

Evertsson and Nermo (2004)

US: neither US: display

US: display US: exchange

US: display Sweden: exchange

US: neither Sweden: exchange

Panel Study of Income Dynamics, 1973 Panel Study of Income Dynamics, 1981, 1991 Panel Study of Income Dynamics, 1999 Swedish Level of Living Survey, 1974, 1981, 1991, 2000

US: display

US: display

Greenstein (2000)

National Survey of Families and Households, 1987–1988

Note. The dependent variable in the studies by Bittman, Brines, and Evertsson and Nermo is absolute housework hours. Greenstein uses both absolute and share measures of housework hours; above based on distributional measure.

the predictions of economic exchange up to the point where their earnings were equal to their husbandsÕ, but deviated in the direction suggested by the gender display hypothesis if their earnings exceeded their husbandsÕ. They argued that the discrepancy between their U.S. and Australian findings reflected genuine national differences. 2.3. Untested assumption of the exchange and display models I suspect that the conflicting findings summarized in Table 1 are due to the questionable validity of womenÕs relative earnings as a predictor of their housework. Consider the models used to test the exchange and display hypotheses, which have the following form: Y i ¼ a þ b1 X i þ b2 X 2i þ ei : Here Yi is an individualÕs housework hours and Xi is her or his share of the coupleÕs total earnings, a commonly used measure of relative resources. (Some studies have used the equivalent ratio of the difference between partnersÕ earnings and the sum.) The linear term represents the exchange effect. Its coefficient is expected to be negative if partnersÕ housework hours are inversely related to their shares of couplesÕ total earnings. The second-order term captures the curvilinearity in the relationship between relative earnings and housework that characterizes gender display. In the case of women, for example, the coefficient will be positive if women with unusually high relative earnings do more housework than other women. A key assumption of this specification is that the unit effects of the two partnersÕ incomes on housework hours are the same. A dollar is a dollar, with respect to its

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association with time spent on housework, regardless of who earns it; what matters is how many dollars one partner earns relative to the other. But the unit associations between the two partnersÕ earnings and the time spent on housework by either of them may or may not be comparable. If they are not, ratio or share measures involving the two partnersÕ earnings will distort the relationship between earnings and housework. Take the limiting case in which there is no association between one partnerÕs earnings and housework time, while there is a non-zero relationship between the other partnerÕs earnings and housework hours. In this situation, what would matter is only how much money the second partner earns, not how much she/he makes compared to the first. To see the implications of unequal unit effects for the commonly employed share measure, compare two couples, one in which the womanÕs and manÕs earnings equal 10 and 30 thousand dollars respectively, and another in which they equal 30 and 90 thousand dollars. The womanÕs share of the coupleÕs total earnings is equal to onefourth in both couples. Models employing her share as an explanatory variable do not distinguish between these two women. But consider the possibility that her partnerÕs earnings are not associated with her housework, or affect it much less than do her own. In that case, the woman in the second couple will do less housework than the woman in the first (assuming that the effect of womenÕs income on housework is negative), even though their earnings relative to their partnersÕ are equal. The share measure disguises possible differences in the effects of the absolute incomes. If the assumption of equal effects of both partnersÕ incomes is violated, the womanÕs share of a coupleÕs total earnings will not be a meaningful predictor of her housework hours. This problem is not solved simply by controlling for total couple income to the exchange and display models, as previous studies have done, for two reasons. First, the addition of couple (or household) income to the model makes the same untested assumption of equality between the unit associations of each partnerÕs earnings with housework hours. Let Ti = Wi + Mi denote the total income of couple i, where Wi and Mi are the female and male partnerÕs earnings, respectively. If we add this to the model, we get: Y i ¼ a þ b1 X i þ b2 X 2i þ bT T i þ ei ¼ a þ b1 X i þ b2 X 2i þ bT ðW i þ M i Þ þ ei ¼ a þ b1 X i þ b2 X 2i þ bT W i þ bT M i þ ei This model forces the coefficients of the two partnersÕ earnings to be equal to the same value, bT. Second, this standard model ignores the empirical relationship between womenÕs relative earnings and couplesÕ total earnings. Fig. 1 depicts the relationship in my sample between womenÕs share of couplesÕ total incomes, or relative earnings, on the horizontal axis and couplesÕ total earnings on the vertical axis, using median splines. Note that couplesÕ total earnings increase with womenÕs relative earnings up to shares of about 40%, but decline after that. Women with the highest shares actually belong to poorer couples than women with the lowest shares. If they spend

own/partner/total earnings (thousands) 10 20 30 40

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0

own earnings partner earnings total earnings 0

20

40 60 share of couple's total earnings

80

100

Fig. 1. WomenÕs own income, partner earnings, and coupleÕs total earnings by womenÕs percentage share of total earnings.

more time on housework than other women, as the display model predicts, is that because their relative earnings are high or because their couplesÕ total earnings are low? The standard model shown above cannot answer this question because it does not account for the curvilinearity in the relationship between relative and total earnings. The most parsimonious solution to these problems is to specify the effects of womenÕs and menÕs earnings separately: Y i ¼ a þ bw W i þ bm M i þ ei : If the two coefficients bm and bw are of comparable size—that is, if their sample estimates bm and bw are not significantly different—we can then proceed to test the standard exchange and display models. If they have substantially different values, however, the standard models may not be valid. 2.4. Possible origins of unequal unit effects of menÕs and womenÕs earnings on housework Substantively, why might the association between womenÕs housework time and their own earnings be different from its relationship with their male partnersÕ? One possibility is that womenÕs earnings have a greater association than their male

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partnersÕ on expenditures on market substitutes for housework. More generally, the literature on intrahousehold resource allocation has documented gender differences in the use of earnings for particular kinds of household consumption. Perhaps the best-known recent study on this issue is by Lundberg et al. (1997), who found that government cash payments to mothers in the UK in the late 1970s were associated with greater expenditures on womenÕs and childrenÕs clothing, compared to expenditures on menÕs. Other studies have found that men withhold more of their incomes for personal use than do women, and that womenÕs earnings may have larger associations with expenditure on household goods such as food (Pfeiffer et al., 2001; Treas, 1993). Further, womenÕs non-wage incomes have larger associations with childrenÕs health and nutrition than do menÕs (Thomas, 1990). Though much of this literature has focused on households in developing countries, Brandon (1999) showed that U.S. mothersÕ own earnings increase the odds that they will choose market child care over parental care; fathersÕ incomes have an effect on child care choices only when husbands and wives pool their incomes. Taken together, these findings are a caution against the assumption that husbandsÕ and wivesÕ earnings will have the same associations with household outcomes and expenses. In particular, there is evidence that womenÕs and menÕs earnings have differing associations with expenses for housework substitutes. Cohen (1998) and Oropesa (1993) found that the effect of wivesÕ incomes on housekeeping expenses was larger than that of husbandsÕ incomes. This suggests that women and men not only spend different amounts of time on housework but may also tap into different sources of money to reduce it, even when they are members of the same couples. Hanson and Ooms (1991) reported that dual-earner couples spent more than twice the amounts spent by single-earner couples on domestic services and day care. Given that the wives in dual-earner couples in the sample likely earned less than the husbands on average, this implies that wivesÕ earnings were deployed to a greater extent for the purpose of housework reduction than their husbandsÕ. Despite these findings, the possibility of separate effects of womenÕs own earnings and their partnersÕ on time spent doing housework has received scant attention in the housework literature over the last two or three decades. In perhaps the only study to date to interrogate the assumption of equal effects of own and partner income, Ross (1987) found that husbandsÕ and wivesÕ incomes had equal but opposite effects in a model of husbandsÕ participation in housework. This would immediately invalidate the share or ratio measures of relative resources commonly employed in the housework literature. However, RossÕs dependent variable was an index of the mean responses to five questions about who does various chores rather than a measure of the actual time spent on them. The metric of this variable makes it impossible to determine the associations between womenÕs and menÕs incomes with actual time spent doing housework. Still, RossÕs study is valuable for its explicit test for the separate effects of partnersÕ incomes. No other study to date has performed such a test, though two analyses found incidentally that womenÕs wages have independent and negative associations with their time spent on housework (Maret and Finlay, 1984; Shelton and John, 1993).

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3. Data and method 3.1. Analytic strategy My analysis proceeds in two steps. 3.1.1. Multivariate model with separate own and partner incomes First, I test the hypothesis of equal effects of womenÕs and menÕs incomes on partnered womenÕs housework using the following specification: Y i ¼ a þ bw W i þ bm M i þ bc C i þ ei : Here Yi denotes the housework hours of woman i, Wi her own annual earnings, and Mi those of her male partnerÕs. The null hypothesis of primary interest is H0 : bw ¼ bm : The controls are collectively represented by the term Ci. The most important of these are the womenÕs and their partnersÕ employment hours, which we would expect to be collinear with incomes. To ensure that the earnings effects in these models are not due to variation in employment hours, I estimate the model separately for women working full time with partners also working full time. The analyses are weighted and standard errors corrected for the complex survey design of the NSFH. 3.1.2. Comparison of married and single women It could be argued that the model described above is just a variant of the exchange and display models, but one that emphasizes absolute rather than relative earnings. Perhaps women with higher earnings spend less time on housework because they have more bargaining power in negotiations over housework than women with lower earnings. Another possibility is that womenÕs earnings, by enhancing their negotiating power, lead to greater involvement in domestic labor by their partners, thereby allowing them to spend less times on it themselves. (It should be noted, however, that this possibility is difficult to distinguish from the potential selectivity of the male partners of women with high earnings.) Or, consistent with the gender display hypothesis, women with high earnings may spend more time on housework than women with lower earnings, not because their earnings are high compared to their male partnersÕ, but simply because they are high in absolute terms. To assess the merit of these possibilities, I compare the effects of married and cohabiting womenÕs earnings on their housework with that of single womenÕs earnings on theirs. By definition, single women do not have to bargain with male partners over housework. Therefore, any association between their earnings and their housework time would be due to processes other than bargaining. If the effect of own earnings on housework is similar for the two groups of women, it would suggest that the relationship between own earnings and housework for partnered women is not driven solely by bargaining, but also by other phenomena common to the two groups of women, for example their use of earnings to purchase market substitutes for housework. The model is specified as follows:

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Y i ¼ a þ bw W i þ bs S i þ bI ðS i W i Þ þ bc C i þ ei : Here Wi represents the earnings of woman i, Si her marital status, and (SiWi) the interaction between those two variables. The null hypothesis of interest is H0 : bI ¼ 0: Unlike the model for partnered women, this model does not contain partner characteristics. 3.2. Sample I use data obtained from the second wave of the National Survey of Families and Households (NSFH), which employed a national probability sample of housing units; one adult per household was randomly selected as the main respondent (Sweet et al., 1988). Members of racial and ethnic minorities were oversampled, as were single-parent families, cohabiting couples, and members of some other types of family. The survey was initiated in 1987; the second wave used in the present study was conducted in the period 1992–1994.1 The first wave of the survey had a response rate of 74%, with 13,007 respondents. The second wave retained 10,005 of these original respondents. The NSFH provides individual case weights that adjust for sampling stratification, nonresponse and attrition; these are described in detail at ftp://elaine.ssc.wisc.edu/pub/nsfh/cmapp_o.001. All models in this analysis employ these weights. The models also incorporate the complex survey design of the NSFH in computing standard errors for the regression coefficients reported here. My sample consists of 2283 women in heterosexual unions at the time of the second wave of the NSFH.2 Of these, 2119 women were married and 164 were cohabiting. The sample is a subset of the 3127 married and cohabiting women between 18 and 65 years of age, inclusive. (Most of the studies described in Table 1 used similar age restrictions.) The single largest reduction of cases from this number is due to the nonresponse of the womenÕs spouses/partners: information on partner characteristics is available for 2553 of the 3127 women eligible to be in the sample. Of these 2553 women, 2283 have non-missing data on all the variables used in the analysis.

1 A third wave became available recently, but it followed up a restricted subset of the sample from the first two waves. (See the NSFH website, http://www.ssc.wisc.edu/nsfh/home.htm, for a complete description of the three waves.) Though the time spent by both men and women on housework has been declining over the last few decades, there is no evidence to date that the determinants of housework time have changed. Accordingly I have chosen the second wave, which has already been extensively used in the housework literature, and which has data on a larger sample than does the third. 2 The earlier studies described in Table 1 were restricted to married women. I include cohabiting women because they also have male coresidential partners whose earnings may affect their housework time. The partner nonresponse rate is higher for cohabiting women (33%) than for married women (17%), but 89% of women in both groups have valid data on all the variables. Though there is some evidence of differences in the housework behavior of married and cohabiting individuals (e.g. Shelton and John, 1993), cohabitation is not statistically significant in any of the analyses presented here, and excluding cohabiting women from the sample does not affect the main findings.

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The implications of missing data for the analysis are discussed at the end of the results section. To ensure that the estimates for the income effects are not unduly influenced by employment hours (which are controlled for in the model), I perform an additional test on a sub-sample of 944 women working full time with partners also working full time. The comparison with single women employs a sample of 1818 women who were not in marital or cohabiting unions. This sample is a subset of the 1865 single women under age 65 for whom valid data is available on all variables. For the comparison, the sample of single women is pooled with the sample of partnered women. 3.3. Variables Following most recent research, this analysis uses absolute number of housework hours as the dependent variable. A few studies have used relative measures of housework time, such as individualsÕ share of couplesÕ total housework hours (e.g. Blair and Lichter, 1991; Greenstein, 2000). However, Bittman et al. (2003) pointed out that measures of housework share do not allow us to determine whether some women do a greater proportion of couplesÕ total housework than others because they spend more time on housework or their partners spend less, or both. Accordingly, the measure used here is the total number of hours women reported spending per week on four tasks: cooking, washing the dishes, housecleaning and laundry. It has become customary in the empirical housework literature to focus on time spent on these particular tasks because they are the staples of household labor. Others, such as yard work, are understood as being relatively discretionary and are typically excluded from measures of routine housework.3 The main independent variables are womenÕs own annual labor market earnings and their partnersÕ, measured in thousands of dollars. The most important controls are both partnersÕ employment hours. Because of the highly clumped nature of the employment hours data, I collapse them into three categories: not employed, employed part time (greater than zero and fewer than 35 h), and employed full time (35 h or more). For the same reason, both partnersÕ educational levels are also categorized into less than high school, at least high school, and some college. Household composition is captured by three indicators for the presence of adult women other than the respondent, adult men other than her partner, and children under the age of 18. For all these variables, using interval-level measures rather than categorical ones makes no difference to the substantive results. Age is included as a continuous variable. The analyses control for race/ethnicity in the form of an indicator

3 To account for the implausibly high values for housework hours reported by some respondents I adopt a procedure used by South and Spitze (1994). Values higher than the 95th percentile are recoded to that percentile for each of the four chores before summing them to obtain the dependent variable. To maximize the number of usable cases, the mean number of hours for each task is imputed for men who do not specify or do not know how many hours they spend on that task. Also, zeros are substituted for men who do not answer the survey question for a particular task but report hours for at least five other tasks.

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variable equal to 1 for white women. Cohabiting women are identified with an indicator variable. It should be noted here that respondents are known to overestimate time spent on housework in retrospective surveys like the NSFH, compared to time diary data (e.g., Bianchi et al., 2000; Juster and Stafford, 1991). However, no study to date has documented systematic variations in this tendency by gender, income, or other important substantive variables. Moreover, the NSFH contains a richer set of covariates than most contemporaneous sources of time diary data; crucially, it contains matched data on respondentsÕ partners. Accordingly, the first two waves of the NSFH have been used widely in studies of domestic labor; a partial listing is available at http://www.ssc.wisc.edu/nsfh/bib.htm#householdtask.

4. Results Summary statistics for all variables used in the multivariate analyses are shown in Table 2. All analyses reported here employ individual case weights, and use the survey regression routines in Stata to compute standard errors that are corrected for the complex survey design of the NSFH. (For most variables, these corrected standard errors were larger than standard errors in regression models that did not incorporate survey design effects.) The table shows weighted and design-adjusted means, standard errors for those means, and standard deviations corresponding to the original unweighted means. In most cases the means shown are not very different from the unweighted means. One of the few exceptions is partnerÕs income, for which the weighted and design-corrected mean is a little over $1000 larger than the uncorrected mean. Focusing on the income and employment variables, we note that among all women, the mean of womenÕs own income is less than half that of partnerÕs income; however, there is greater variation in the latter. One-third of the women in this group are not employed, and almost half are employed full time. For the sake of brevity, descriptive statistics on the subsample of women working full time are omitted, and are available from the author. Women in this subsample, who also have partners working full time, spend about 18% fewer hours on housework than all women. Not surprisingly, the gap between the mean own and partner incomes is smaller, as is the gap between own and partner employment hours. 4.1. Descriptive findings: Tables 3a and 3b The associations between womenÕs housework hours, and their own and their partnersÕ earnings can be seen in Tables 3a and 3b. These tables report means of womenÕs housework cross-tabulated by quartiles of their own and their partnersÕ earnings. Table 3a reports means for all women; Table 3b is restricted to women working full time (35 h or more) with partners also working full time. The rows in both tables correspond to quartiles of womenÕs own earnings and the columns to quartiles of their partnersÕ earnings. The tables show the greater association of womenÕs housework and their own earnings, compared to its relationship with their

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Table 2 Summary statistics, all women—survey design-corrected mean, standard error of mean, and uncorrected standard deviation (N = 2283) Variable

Design-corrected mean

S.E. of mean

Non-design S.D.

Housework hours Own income (thousands) PartnerÕs income (thousands)

24.93 13.51 30.71

0.29 0.36 0.61

14.32 14.57 26.11

0.32 0.20 0.48

0.01 0.01 0.01

0.46 0.40 0.50

40.73 0.86

0.26 0.01

9.13 0.37

Education Less than high school At least high school Some college

0.10 0.65 0.26

0.01 0.01 0.01

0.30 0.47 0.43

PartnerÕs employment status Not employed Part-time Full-time

0.12 0.04 0.84

0.01 0.01 0.01

0.32 0.19 0.36

PartnerÕs education Less than high school At least high school Some college

0.13 0.58 0.29

0.01 0.01 0.01

0.34 0.49 0.45

Household composition Children less than 18 years old Adult males (other than partner) Adult females (other than self)

0.58 0.14 0.12

0.01 0.01 0.01

0.48 0.34 0.32

Union type Cohabiting

0.06

0.00

0.26

Employment status Not employed Part-time Full-time Sociodemographic characteristics Age (years) Race (1 = white)

partnersÕ earnings. In Table 3a we see that the mean housework hours for women in the top quartile of own earnings is 19, compared to 32 h for women in the lowest quartile. Though women in the highest quartile of partner earnings also do less housework than those in the lowest quartile of partner earnings, the contrast is less striking. To look at it another way, women in the poorest quartile of own earnings and partnered with the highest-earning men spend more time on housework on average (31 h) than women in the highest quartile of own earnings and partnered with the lowest-earning men (20 h). Other gross differences among women based on earnings are also apparent. Women in the highest quartiles of own and partner earnings have the lowest mean housework hours, 18, and women in the lowest quartiles of both have the highest, 33 h.

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Table 3a Means of womenÕs housework hours, by own and partner income, with cell frequencies Quartiles of own earnings

Quartiles of partner earnings 1

2

3

4

1.

Mean n

33 147

33 133

33 103

31 197

Total 32 580

2.

Mean n

28 164

28 132

27 154

26 106

27 556

3.

Mean n

23 147

23 197

22 160

22 94

23 598

4.

Mean n

20 109

19 118

19 146

18 176

19 549

Total

Mean N

26 567

26 580

25 563

24 573

25 2283

Table 3b Means of womenÕs housework hours by own and partner income, both partners working full time (cells with n < 30 highlighted) Quartiles of own earnings

Quartiles of partner earnings 1

2

3

4

Total

1. First quartile omitted because nÕs < 10 2.

Mean n

29 50

26 46

22 53

30 15

26 164

3.

Mean n

21 69

23 146

21 127

21 54

22 396

4.

Mean n

21 48

19 90

19 117

16 116

18 371

Total

Mean N

23 173

22 283

20 300

19 188

21 944

Because Table 3a includes all women, those in the lowest quartile of own earnings include homemakers, unemployed women, and other women with zero incomes. Likewise, those in the lowest quartile of partner earnings include women whose male partners are not employed or for other reasons have zero incomes. The relationship between earnings and housework depicted in it could therefore be affected by variation in employment hours. Hence Table 3b is restricted to women working full time with partners who also are employed full time; here both partners have non-zero earnings and there is much less variation in employment hours. Because there are very few women employed full time in the row corresponding to the lowest quartile

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of own earnings, the table excludes that quartile. The basic pattern, though less dramatic, is similar to that observed in Table 3a. The association of womenÕs own income with their housework hours is stronger than that of their partnersÕ. If minimizing housework time is the objective, women are better off if their own earnings are high than if their partnersÕ are, and they do best if both their own and their partnersÕ earnings are high. 4.2. Multivariate results: Tables 4a and 4b Table 4a presents the results of a multivariate model for womenÕs housework hours that accounts for the possibility of distinct effects of own and partner income. The first panel in the table shows a restricted model that includes only own and partnerÕs employment statuses as controls. We observe that both womenÕs own and their partnersÕ incomes reduce their housework hours, but that the coefficient for own income is more than three times greater than that of partner income. The contrast

Table 4a Multivariate model with separate effects of own and partner income, all women (N = 2283) b

S.E.

p > Œt Œ

b

S.E.

p > Œt Œ

Own income PartnerÕs income

0.17 0.05

0.028 0.012

0.000 0.000

0.12 0.02

0.027 0.012

0.000 0.167

Own employment status Part-time Full-time

4.65 8.23

0.866 0.958

0.000 0.000

4.15 7.37

0.811 0.944

0.000 0.000

3.33 3.68

2.532 0.982

0.193 0.000

3.45 2.86

2.368 0.938

0.150 0.003

Sociodemographic characteristics Age White

0.13 1.53

0.038 1.053

0.001 0.150

Own education At least high school Some college

2.10 3.46

1.598 1.721

0.194 0.049

Partner education At least high school Some college

3.45 2.86

2.368 0.938

0.150 0.003

5.85 0.11 0.56

0.645 1.022 0.928

0.000 0.914 0.546

0.14

1.219

0.906

28.27 0.21

2.627

0.000

Partner employment status Part-time Full-time

Household composition Children present Adult male present Adult female present Union type Cohabiting Constant R2

30.53 0.14

1.084

0.000

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between the two coefficients becomes even sharper upon the addition of all controls, as shown in the second panel of Table 4a. Here the coefficient for partner income loses statistical significance altogether. The coefficient for own income is smaller than that in the restricted model but remains significant, with every increment of 8.3 thousand dollars corresponding to 1 h less of housework. Even though the models in Table 4a control for womenÕs own and their partnersÕ employment statuses, the models in Table 4b isolate income effects further by restricting the sample to women working full time with partners also working full time. The coefficient for own income is 2.3 times bigger than that for partner income, a smaller difference than that observed in the entire sample but still large. The coefficient for partner income once again loses statistical significance in the full model. Note also that the coefficient for own income is comparable in size to that in the full model shown in Table 4a. Though women working full time (and with partners working full time) spend less time on housework than all women, the effect of their own income on their housework is about the same as it is for all women. The coefficients of the control variables are generally consistent with previous research on womenÕs housework. Employed women spend less time on domestic labor than women who are not employed; in addition, women working full time do significantly less housework than women working part time. They spend more time on Table 4b Multivariate model with separate effects of own and partner income, both partners employed full time(N = 944) b

S.E.

p > Œt Œ

b

S.E.

p > Œt Œ

Own income PartnerÕs income

0.15 0.05

0.039 0.020

0.000 0.011

0.12 0.03

0.036 0.023

0.001 0.192

Own employment hours Partner employment hours

0.14 0.02

0.111 0.061

0.213 0.749

0.17 0.03

0.102 0.059

0.102 0.578

Sociodemographic characteristics Age White

0.20 1.96

0.056 1.085

0.001 0.075

Own education At least high school Some college

1.12 2.55

2.045 2.168

0.587 0.243

Partner education At least high school Some college

0.85 3.23

1.903 2.116

0.655 0.131

Household composition Children present Adult male present Adult female present

4.88 0.74 0.05

0.956 1.348 1.275

0.000 0.584 0.970

0.13

1.218

0.915

12.24 0.11

6.981

0.084

Union type Cohabiting Constant R2

20.23 0.05

5.410

0.000

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housework if their male partners are employed full time, but partnersÕ part-time employment does not significantly affect their housework hours. Both their own and their partnersÕ educational levels affect their housework, and in the same direction, though the effect of their own education is just significant at the 5% level. The difference between the effects of high school and college education for partners is significant. These results are certainly consistent with the bargaining perspective— women with some college education may have greater power in housework negotiations, and better-educated partners may be more flexible in these negotiations, or perhaps be willing to do more housework themselves. But the effect of education may also be due in part to its collinearity with income. In separate models excluding own and partnerÕs education, the effects of own and partnerÕs income are larger, and the coefficient for partnerÕs income is significant (results not shown). This suggests that education is absorbing some of the variance in housework that would otherwise be explained by income, particularly in the case of partners. That possibility is lent further credence by the lack of significance of education among women working full time with full-time partners. Finally, the presence of a child increases womenÕs housework by nearly 6 h. To summarize: in the limited models, the effect of womenÕs own income is two to three times larger than that of their partnersÕ, and in the full models with all controls, partnersÕ income has no statistically significant effect on womenÕs housework. The evidence against the null hypothesis of equal associations of womenÕs own earnings and their partnersÕ with womenÕs housework is strong. It appears that the associations between womenÕs housework and their earnings relative to their male partnersÕ observed in earlier studies are driven largely by womenÕs own earnings. 4.3. Comparison with single women: Fig. 1, Table 5 Fig. 1 illustrates the relationship between own income and housework hours for both single and partnered women, smoothed with a set of median splines. The actual data points plotted are for partnered women. The pattern looks quite similar for the two groups of women, though for any given income married women spend more time on housework than single women. The biggest impact of womenÕs earnings on their housework hours occurs up to about 20 thousand dollars, which corresponds to the 75th percentile of income for both single and partnered women, and it tapers off after that. Thus it appears that there is a slowing of the effect of womenÕs income on their housework at higher incomes; it may pick up again at very high incomes for single women, but there are too few women in the sample with such incomes to make robust inferences. Possibly there are diminishing returns to income beyond a certain point in terms of market substitutes for housework. Or there may be a latent floor in womenÕs housework because they are unwilling or unable to reduce housework below a certain point. Whatever mechanism leads to the weakening of the income association seems to operate for both single and partnered women. If it is gender display, partnered women do not appear to be more affected by it than are single women (Fig. 2).

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single

0

housework hours 20 40

60

partnered

0

20 40 own earnings (thousands)

60

Fig. 2. Housework hours of single and partnered women, by own income.

Table 5 presents a formal test of the difference of the associations between womenÕs own earnings and their housework for single and partnered women. Partner characteristics are excluded from this model, because single women by definition do not have partners. The interaction of marital status and own earnings does not achieve statistical significance. That is, the negative association between earnings and housework time is not appreciably different in magnitude for single women compared to women living with male partners. Further, note that the coefficient of own earnings is not substantially different from that reported in the full model in Table 4a for all partnered women, which includes partner characteristics. As with the model for all partnered women in Table 4a and 4b, I performed a separate comparison for single and partnered women working full time. Here also the interaction term did not achieve significance (results omitted for brevity). This is not to say that there are no differences between the determinants of single and partnered womenÕs housework time. Table 5 shows that partnered women benefit more from education than single women in terms of spending less time on housework. On the other hand, single women benefit more from the presence of other adult women in their households, most often their mothers. But regardless of these differences, the associations between own earnings and housework time appear to be the same for both groups of women. That is, partnered women do not appear to benefit greatly from their access to an additional source of income, their partnersÕ,

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Table 5 Associations of own income for single and partnered women (N = 4101) b

S.E.

p > Œt Œ

Single Own income Single · own income

5.32 0.13 0.02

3.577 0.027 0.042

0.142 0.000 0.615

Own employment status Part-time Full-time

4.27 6.99

0.833 0.932

0.000 0.000

Sociodemographic characteristics Age White

0.11 2.31

0.036 1.099

0.004 0.039

Own education At least high school Single · high school Some college Single · college

3.96 4.13 7.27 4.85

1.535 1.903 1.522 2.173

0.012 0.034 0.000 0.029

Household composition Children present Adult male present Adult female present Single · adult female

6.00 0.03 0.39 3.04

0.623 1.068 0.976 1.221

0.000 0.976 0.690 0.015

Constant R2

29.36 0.21

2.360

0.000

Note. In addition to the interactions shown here, the model includes interactions between marital status and all other independent variables. The ones not shown are not statistically significant at the 5% level.

compared to single women, who do not have access to the income of another wageearner. 4.4. Robustness of results To establish the robustness of the association between womenÕs own earnings and their housework time, and its difference from the association of their housework with their male partnersÕ earnings, I performed several checks on the model reported in Table 4a and 4b. First, to account for the leveling off of the slope of the relationship depicted in Fig. 1, I estimated the model excluding women whose annual earnings exceeded 25 thousand dollars. In this model the coefficient of their own earnings was twice as large as it appears in the full model in Table 4a, i.e. 0.24; their partnersÕ earnings remained not significant. As a further test I excluded women with zero earnings, with the same result—the association of their housework with their own earnings was about twice as large as reported in Table 4a and 4b, and that of their partnersÕ incomes was not significantly different from zero. An additional model with splines to allow for different slopes at different points in the earnings distribution yielded the same general conclusion, namely that the coefficient is largest up to about

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the 75th percentile of womenÕs earnings. The coefficient reported here is, therefore, a lower bound. To confirm my suspicion that the relative earnings associations documented in previous research were driven mostly by womenÕs absolute incomes, I estimated another model that adds womenÕs share of couplesÕ total earnings to the model in Tables 4a and 4b, but omits male partnersÕ earnings to avoid multicollinearity. The relative earnings variable does not achieve statistical significance even at the 10% level (results not shown).4 Further, the results were not affected by the choice of functional forms for key controls such as employment hours and educational level. Substituting the continuous measures for the categorical ones used here made no difference. Finally, a number of checks were performed to assess the impact of sample restrictions and missing data on the results. The largest reduction in the potential sample size of 3127 is due to the nonresponse of the spouses/partners of nearly 20% of these women. An additional model using all these women, but excluding the partner variables, yielded a coefficient of 0.13 for womenÕs own earnings, nearly identical to the one reported in Table 4a and 4b. Among the 2553 women with spouse/partner data, imputing values for missing data on the model variables, using a variety of techniques, did not alter the results shown in Table 4a and 4b. Also, because the earlier studies shown in Table 1 were limited to married women, I estimated another model that excluded the cohabiting women who constitute 7% of my sample. This accentuated the association of housework with womenÕs own earnings, which suggests that the association may be weaker for cohabiting women than for married women. However, because the number of cohabiting women in the sample is relatively small, it is difficult to generalize from this finding.

5. Discussion Much of the literature to date on the relationship between earnings and housework has focused on the impact of individualsÕ earnings relative to their partnersÕ. By contrast, this paper analyzes the independent effects of each partnerÕs absolute incomes on womenÕs housework. A conservative characterization of my findings is that the effect of womenÕs own incomes on their housework hours is two to three times larger than that of their male partnersÕ incomes. With controls for household composition and education, the association of their housework time with their partnersÕ incomes loses statistical significance entirely. This is the case even among women working full time and whose partners work full time. This finding has been replicated 4 In an overspecified model that includes all three variables—both partnersÕ earnings and womenÕs share—the share measure is significant in the model for all women, but its coefficient is quite fragile. It loses significance when women with either zero incomes or the highest 1% of own earnings are excluded from the sample. And it is not significant among the women working full time whose partners also work full time. In other words, it appears to be driven by threshold effects of extremely low and high earnings. By contrast, the coefficient of womenÕs own earnings is resilient across all these specifications.

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in other datasets, including the German Social and Economic Panel (GSOEP), the Swedish Level of Living survey, and multiple waves of the US Panel Study of Income Dynamics (PSID). In each case, the association between womenÕs housework and their husbandsÕ earnings is either substantially weaker than its relationship with their own earnings, or not statistically significant (results available from author). This finding, therefore, resolves the discrepancies across data sources described in Table 1. The results reported here challenge the implicit assumption of the relative resources and gender display models that the unit effects of both partnersÕ incomes are equal. An immediate implication is that models employing ratio or share measures as independent variables should first determine the magnitudes of the associations of each of the components of the measure with the dependent variable. If these associations are not of comparable size, the validity of these measures may be in doubt, regardless of their theoretical soundness. If a woman makes half as much money as her husband, but each dollar she earns has twice the effect on her housework time as a dollar earned by her male partner, her share of the coupleÕs total earnings is effectively 100%, with respect to its association with housework time. And if it is the case that their male partnersÕ earnings are not at all associated with womenÕs housework time, as in the full model reported in Tables 4a and 4b, the models need account only for the level of womenÕs own earnings. 5.1. Interpretation of unequal associations How do we interpret substantively the finding of greater association of womenÕs own earnings with their housework hours, compared to that of their partnersÕ? One possibility is that women draw from their own earnings to a greater extent than they do from their partnersÕ in order to purchase market substitutes for housework, such as dining out and domestic services. This strategy may also ease the friction associated with negotiations over the allocation of domestic labor documented in the classic study by Hochschild and Machung (1989). WomenÕs use of their own earnings to substitute for their housework may mitigate what Treas (1993) described as the ‘‘transaction costs’’ of decision-making in households, or the efforts required to coordinate, negotiate and monitor their membersÕ activities. Treas found that couples organized their finances in ways that minimized these transaction costs. For some couples this meant separate bank accounts, while others pooled their financial assets. Although TreasÕs study did not deal directly with the organization of expenses, it raises the possibility that partners do not pool their earnings for all purposes. Because the NSFH does not have expense data, I cannot directly test this hypothesis. However, I conducted a separate analysis using data from the Consumer Expenditure Survey (CES) conducted by the Bureau of Labor Statistics, 1998–2000. The association between married womenÕs own earnings and household expenditures on two such substitutes, dining out and domestic services, was indeed significantly greater than that between their husbandsÕ earnings and these expenses, controlling for the employment hours of both spouses (results available from the author). This result accords with earlier studies of household expenses such as Cohen (1998) and Oropesa (1993).

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Another explanation for the finding is that women may feel freer to buy out their housework if they themselves earn more. That is, couples may pool their earnings for the purpose of housework substitution expenses, but women may draw from the pool in proportion to their own incomes rather than their partnersÕ. It is also possible that the effect of their own earnings simply reflects a reduced sense of obligation to do housework on the part of higher-earning women, even if they do not use their earnings to substitute for their housework. Further, it might be that the male partners of higher-earning women pick up the time that would be spent on housework by those women if their earnings were lower. However, there is no significant association between womenÕs earnings and their partnersÕ housework hours (results not shown). To determine exactly what lies behind the larger association of womenÕs housework time with their own earnings compared to their partnersÕ would require a dataset with information on earnings, time use, expenses and couplesÕ organization of finances. Unfortunately, virtually no such data exist. Whatever the correct interpretation, it would need to account for the second principal finding of this study, namely that the association of partnered womenÕs housework time with their own earnings is similar to that of single women. The quantitative literature on the association between womenÕs earnings and housework has focused solely on married and cohabiting women. But if, as reported here, the relationship between own earnings and housework time is comparable for single and partnered women, we should consider the possibility that women in heterosexual couples use their earnings as a source of purchasing power, not just bargaining power. Likewise, any explanation involving gender display should account for the association between own earnings and housework hours among single women, who presumably do not have the same imperative as partnered women to display gender or neutralize gender deviance. It is possible that the association between womenÕs own earnings and housework involves two processes, one that is similar to that for single women, and another that involves bargaining with their male partners. Although their share of couplesÕ total earnings is not significant in models that include their absolute earnings (not shown), there may be some kind of joint effect not captured by the present analysis. 5.2. Importance of earnings variation among women In addition to documenting a difference in the associations of womenÕs own earnings and their partnersÕ with womenÕs housework, my findings emphasize the importance of earnings-based variation in womenÕs domestic labor. The focus on the relative earnings of men and women in the recent quantitative literature has had the unintended consequence of obscuring this class dimension of womenÕs housework. Table 4b shows that among women working full time with partners working full time, those in the lowest category of own and partner earnings spent an average of 13 more hours on housework than women in the highest category. That income gap is larger than the difference between the mean housework hours of these women and their male partners. From the multivariate results for the same group of women in Table 4b, we see that those at the 90th percentile of own earnings, or 34 thousand

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dollars, spend 4.3 fewer hours on housework than those at the 10th percentile earning 8 thousand dollars (4.3 = (0.25 · 20 + 0.09 · 14)  0.25 · 8). That almost compensates for the additional housework associated with the presence of children. The pattern shown in Table 3a and 3b may help explain the curvilinear relationship between womenÕs relative earnings and their housework observed in some earlier studies, which have found that women with exceptionally high shares of couplesÕ total earnings spend more time on housework than other women. Fig. 1 depicts the relationship in the sample between womenÕs share of couplesÕ total incomes on the horizontal axis and their own earnings, partnersÕ earnings, and couplesÕ total earnings on the vertical axis. The relationships between all of the earnings variables and womenÕs share of total incomes are indicated by median splines. We note that womenÕs own earnings increase almost linearly up to shares of about 60%, plateau beyond that value, and actually decline at the highest values of earnings share. This means that higher shares correspond to higher own earnings up to about the point that womenÕs earnings equal their partnersÕ, but higher shares beyond that point correspond to steady or declining own incomes. Further, women with income shares below 20% have the highest-earning male partners, and women with shares of 80% and above have the lowest-earning partners. Thus, women in the highest range of earnings share belong to the poorest couples. They are poorer than women in the equality range, and poorer even than women in the sub-equal range. Their spending more time on housework than other women may therefore reflect their lower purchasing power rather than gender display. There might also be class norms regarding appropriate amounts of time spent doing housework, with the normative requirements being higher for poorer women. This does not appear to be the most important factor, however, given that low-earning women with high-earning male partners spend more time on housework than women with high own earnings. It matters crucially what womenÕs own earnings are, and not just their SES as measured by couplesÕ total earnings. In a separate analysis, I confirmed that the predicted values from the separate incomes model reported here overlap almost completely with those of the gender display model (results available from author). These results echo CohenÕs (1998, p. 228) caution against an ‘‘undifferentiated description of gender experience’’ in the matter of domestic labor. It is not my intent to deny that couples bargain over the division of domestic labor, nor that womenÕs relative earnings may play a role in the display of gender identities. Rather, the aim of this analysis is to call attention to the impact of womenÕs own absolute earnings on their housework hours. One might think that because womenÕs domestic labor benefits the entire household, both partnersÕ earnings would count equally towards explaining variation in it, but in fact womenÕs own earnings matter much more than their partnersÕ. This emphasizes the gender segregation of unpaid labor in the home, and the disadvantages of heterosexual coresidential partnership for women where domestic labor is concerned. Not only do partnered women spend more time on housework than single women, but they also do not benefit greatly from their partnersÕ incomes in terms of housework reduction. Because of data limitations, I cannot determine the reasons for this. If it is due to the greater negative association of womenÕs own earnings on household expenses for housework

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substitutes, it would suggest that women are free to spend their own earnings to reduce the burden of housework. But this would be a peculiar kind of freedom. It belies the cooperative ideal whereby partners pool resources and then redistribute them for the maximum benefit of each partner or the household as a whole. It would be another way in which the economic reality of marriage violates the cultural norms of the unified relationship.

References Becker, G.S., 1991. A Treatise on the Family. Harvard University Press, Cambridge. Bianchi, S.M., Milkie, M.A., Sayer, L.C., Robinson, J.P., 2000. Is anyone doing the housework? Trends in the gender division of household labor. Social Forces 79 (1), 191–228. Bittman, M., England, P., Folbre, N., Sayer, L., Matheson, G., 2003. When does gender trump money? Bargaining and time in household work. American Journal of Sociology 109 (1), 186. Blair, S.L., Lichter, D.T., 1991. Measuring the division of household labor: gender segregation of housework among american couples. Journal of Family Issues 12 (1), 91–113. Blumberg, R.L., Coleman, M.T., 1989. A theoretical look at the gender balance of power in the American couple. Journal of Family Issues 10 (2), 225–250. Brandon, P.D., 1999. Income-pooling arrangements, economic constraints, and married mothersÕ child care choices. Journal of Family Issues 20 (3), 350–370. Brines, J., 1994. Economic dependency, gender, and the division of labor at home. American Journal of Sociology 100 (3), 652–688. Cohen, P.N., 1998. Replacing housework in the service economy: gender, class, and race-ethnicity in service spending. Gender and Society 12 (2), 219–231. Coverman, S., 1985. Explaining husbandsÕ participation in domestic labor. Sociological Quarterly 26 (1), 81–97. England, P., Farkas, G., 1986. Reproduction and production in the household. In: Households, Employment, and Gender: A Social, Economic and Demographic View. Aldine, New York, pp. 73–101. Evertsson, M., Nermo, M., 2004. Dependence within families and the division of labor: comparing Sweden and the United States. Journal of Marriage and the Family. Farkas, G., 1976. Education, wage rates, and the division of labor between husband and wife. Journal of Marriage and the Family 38 (3), 473–483. Greenstein, T.N., 2000. Economic dependence, gender, and the division of labor in the home: a replication and extension. Journal of Marriage and the Family 62 (2), 322–335. Gupta, S., 1999. Gender display? A reassessment of the relationship between menÕs economic dependence and their housework hours. In: Paper Presented at the Annual Meeting of the American Sociological Association, Chicago, IL. Hanson, S.L., Ooms, T., 1991. The economic costs and rewards of two-earner, two-parent families. Journal of Marriage and the Family 53 (3), 622–634. Hochschild, A., Machung, A., 1989. The Second Shift: Working Parents and the Revolution At Home. Viking, New York. Huber, J., Spitze, G., 1983. Sex Stratification: Children, Housework, and Jobs. Academic Press, New York. Juster, F.T., Stafford, F.P., 1991. The allocation of time: empirical findings, behavioral models, and problems of measurement. Journal of Economic Literature 29 (2), 471–522. Lopata, H.Z., 1993. The interweave of public and private: womenÕs challenge to American society. Journal of Marriage and the Family 55 (1), 176–190. Lundberg, S.J., Pollak, R.A., Wales, T.J., 1997. Do husbands and wives pool their resources? Evidence from the United Kingdom child benefit. Journal of Human Resources 32 (3), 463–480.

S. Gupta / Social Science Research 35 (2006) 975–999

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Maret, E., Finlay, B., 1984. The distribution of household labor among women in dual-earner families. Journal of Marriage and the Family 46 (2), 357–364. Oropesa, R.S., 1993. Using the service economy to relieve the double burden: female labor force participation and service purchases. Journal of Family Issues 14 (3), 438–474. Pfeiffer, J., Gloyd, S., Li, R.L., 2001. Intrahousehold resource allocation and child growth in Mozambique: an ethnographic case-control study. Social Science and Medicine 53 (1), 83–97. Ross, C.E., 1987. The division of labor at home. Social Forces 65 (3), 816–833. Shelton, B.A., John, D., 1993. Does marital status make a difference? Journal of Family Issues 14 (3), 401–420. South, S.J., Spitze, G., 1994. Housework in marital and nonmarital households. American Sociological Review 59 (3), 327–347. Sweet, J., Bumpass, L., Call, V., 1988. The design and content of the national survey of families and households (NSFH Working Paper #1). Center for Demography and Ecology, University of Wisconsin at Madison. Thomas, D., 1990. Intra-household resource allocation: an inferential approach. Journal of Human Resources 25 (4), 635–664. Treas, J., 1993. Money in the bank: transaction costs and the economic organization of marriage. American Sociological Review 58 (5), 723–734. West, C., Zimmerman, D.H., 1987. Doing gender. Gender and Society 1 (2), 125–151.