Are Lone Mothers Responsive to Policy ... - Wiley Online Library

Scand. J. of Economics 114(4), 1129–1159, 2012 DOI: 10.1111/j.1467-9442.2012.01722.x

Are Lone Mothers Responsive to Policy Changes? Evidence from a Workfare Reform in a Generous Welfare State∗ Magne Mogstad Statistics Norway, NO-0317 Oslo, Norway [email protected]

Chiara Pronzato Bocconi University, 20136 Milano, Italy [email protected]

Abstract There is a heated debate in many developed countries about how to design a welfare system that moves lone mothers off welfare and into work. We analyze the consequences of a major Norwegian workfare reform of the generous welfare system for lone mothers. The reform imposed work requirements and time limits on welfare receipt, while raising in-work benefits. Our difference-in-differences estimates show that the reform was successful in improving labor-market participation and in increasing the earnings of lone mothers. However, the reform was associated with income loss and increased poverty among a sizeable subgroup of lone mothers, who were unable to offset the loss of out-of-work benefits with gains in earnings. Keywords: Disposable income; earnings; labor-market participation; lone mothers; poverty; workfare reform JEL classification: C23; I32; I38; J 00

I. Introduction Most of what we know about how lone mothers respond to policy changes that increase their incentives to move off welfare and into work comes from program evaluations carried out in Canada, the UK, and the US. The results are striking: the welfare reforms generally have positive effects on employment, earnings, and income, and they reduce program caseloads

∗ We thank Rolf Aaberge, Tony Atkinson, Richard Blundell, Andrea Brandolini, Ugo Colombino, John Ermisch, Marco Francesconi, Tarjei Havnes, Terje Skjerpen, and two anonymous referees, as well as participants at a number of seminars and conferences, for helpful comments. The Norwegian Research Council and the European Research Council have provided financial support for this project.

C The editors of The Scandinavian Journal of Economics 2012. Published by Blackwell Publishing, 9600 Garsington

Road, Oxford, OX4 2DQ, UK and 350 Main Street, Malden, MA 02148, USA.

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M. Mogstad and C. Pronzato

and government expenditure.1 The Anglo-Saxon experience has fueled an increasing interest among European policy-makers and researchers in policy changes that provide stronger work incentives for lone mothers. Yet, caution must be applied when drawing lessons from the Anglo-Saxon experience, as emphasized in several recent review articles (e.g., Blank, 2002; Moffitt, 2007; Brewer et al., 2009). In particular, the impact of welfare reforms might depend heavily on the broader institutional context and the economic environment in which they are implemented. With this background, Brewer et al. (2009) have concluded that evidence from countries with different economic and institutional structures from the Anglo-Saxon countries has significant value for research and policy. In this paper, we examine the consequences of a major Norwegian workfare reform of the generous welfare system for lone mothers. The reform imposed and enforced work requirements and time limits on welfare receipt, while raising in-work benefits. The aim of the reform was to improve the labor-market attachment of lone mothers, and thereby to reduce welfare dependency and to alleviate poverty. Our study provides the first evidence of the effects of the Norwegian workfare reform. To this end, we make use of a unique household panel dataset, based on administrative registers covering the entire resident population. To identify the effects of the reform on lone mothers, we use a difference-in-differences (DD) approach. Our baseline specification compares the outcome of interest of lone mothers to that of a comparison group of married mothers (with children of the same age), before and after the reform.2 To increase confidence in our identification strategy, we perform several specification checks. Unlike most previous studies of welfare reforms that target lone mothers, we investigate the impacts not only on earnings and labor-market participation, but also on disposable income and poverty. The insights can be summarized in two broad conclusions. First, the workfare reform was successful in increasing the average earnings and labor-market participation of lone mothers. Second, the reform was associated with the side effects of income loss and increased poverty among a sizeable subgroup of lone 1 Lone-parent benefits in the US underwent a major workfare reform in 1996, when time limits and work requirements were imposed, the funding for childcare increased and, in many states, the benefit reduction rates were lowered. Moffitt (2007) has summarized the evidence found from this much-studied reform. In addition, there are a number of program evaluations of in-work benefit reforms of the Earned Income Tax Credit reform in the US (Eissa and Liebman, 1996; Meyer and Rosenbaum, 2001), of the Working Families’ Tax Credit reform in the UK (Brewer and Gregg, 2001; Blundell et al., 2005; Francesconi and van der Klaauw, 2007), and of the Canadian Self-Sufficiency Project (Michalopoulos et al., 2005; Card and Hyslop, 2006; Bitler et al., 2008). Brewer et al. (2009) have reviewed the evidence from the reform of the Working Families’ Tax Credit. 2 Throughout this paper, we have included mothers who are cohabiting in the “married” category.

C The editors of The Scandinavian Journal of Economics 2012.

Are lone mothers responsive to policy changes? 1131

mothers, who were unable to offset the loss of out-of-work benefits with gains in earnings. Because theory predicts systematic heterogeneity in the impact of the reform, we also look closely into whether the policy affected different groups differently. The remainder of the paper is organized as follows. In Section II, we describe the Norwegian welfare system for lone mothers, and we discuss the expected effects of the workfare reform. In Section III, we outline our empirical strategies, and we describe the data in Section IV. In Section V, we report on the empirical results, and we conclude in Section VI.

II. Background In this section, we describe in detail the workfare reform of the Norwegian welfare system for lone mothers, before commenting on predictions from labor supply theory.

Policy Changes Historically, the transitional benefit scheme has been a generous out-ofwork welfare program targeted exclusively at lone mothers. A workfare reform of the transitional benefit scheme was introduced on January 1, 1998. There were three important changes. First, work requirements were imposed, although only for lone mothers whose youngest child was three years old or more. Second, for eligibility, the age limit on the youngest child was lowered, and time limits on welfare participation were introduced. Third, in-work benefit levels were raised. Table 1 provides more details of the transitional benefit scheme and the changes made in the 1998 reform. Because the aim of this paper is to evaluate the workfare aspects of the reform, we focus attention on lone mothers whose youngest child is older than three years, because these were the parents faced with work requirements. Another reason for not focusing on lone mothers with younger children is that, in 1998, the Norwegian government introduced a cash-forcare reform, which is a cash transfer to married and lone mothers with children aged one or two who do not make or only partly make use of government-subsidized daycare centers.3

3 To obtain consistent estimates of the transitional benefit reform on lone mothers with their youngest child less than three years of age, it would have been necessary to assume that the cash-for-care reform had the same impact on married and lone mothers with young children. Schøne (2003) and Naz (2004) find that the reform reduced employment among married mothers, in particular among those with high education. See also Havnes and Mogstad (2011a, 2011b) for a discussion of government-subsidized childcare arrangements in Norway, and their relationship to maternal labor supply and child development.


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Table 1. Key features of the transitional benefit reform (€ – 1998) Characteristic Maximum benefit level Benefit reduction rate

Before the reform

€695 per month

Work requirements

40 percent of earnings exceeding a threshold of €215 per month None

Time limit

None

Age limit

Youngest child less than 9–10 years old (fourth grade of primary school) None

Means-testing of benefits depending on assets

After the reform

€855 per month

40 percent of earnings exceeding a threshold of €230 per month If youngest child is three years old or more, the lone parent has to work half time Maximum three years of welfare receipt Youngest child less than eight years old None

Economic Impacts of the Workfare Reform Figure 1 shows the budget constraint before and after the welfare reform in order to illustrate how the work requirements and the increase of inwork benefits affected work incentives. Specifically, the figure shows how disposable income on the vertical axis varies with working hours per week on the horizontal axis; the earnings and welfare components (after tax) are above the zero line, while the taxes and childcare costs are below it. For brevity and with minimal loss of generality, we present only the work incentives for a lone mother with one child who has an hourly wage equal to 75 percent of the average wage in the labor force. To assess the expected effects of the workfare reform, we begin with the usual static labor supply model assumptions: the woman can freely choose hours of work at the given offered wage, and offered wages are constant. In particular, we ignore any human capital, search-theoretic, or related issues. We also assume that there is no time limit. Later, we relax these assumptions. Consider, first, the case in which a lone mother prefers to work zero hours and therefore she receives the maximum transitional benefit payment, when exposed to the pre-reform rules. Depending on the lone mother’s preferences, the reform could lead to either of two outcomes. First, she might continue not to work and to receive zero transitional benefits. Second, she might enter the labor market; transitional benefit payment remains zero until she works half-time. At this point, she satisfies the work requirement, which makes her eligible for in-work benefits and generates a sharp jump in disposable income. Hence, the reform provides strong incentives for lone mothers to enter the labor market and, especially, to work at least half-time. C The editors of The Scandinavian Journal of Economics 2012.


Fig. 1. Work incentives before and after the reform for a lone mother with one child 4–5 years of age Notes: Figure 1 considers a lone mother with one child who has an hourly wage equal to 75 percent of the average wage in the labor force. The figure is based on an exact representation of the Norwegian tax-benefit system. The exceptions are social assistance and housing benefits, which are granted at the discretion of social security office staff, supplementary to other social policies as last resorts of assistance. How these benefits vary with earnings is therefore imputed from the data. Children start school the year they turn six. To reflect childcare costs, the figure assumes that the child is between 4 and 5 years of age. Primary school is compulsory and free of charge.

Consider next a lone mother who prefers to work positive hours. Therefore, she receives less than the maximum transitional benefit payment, when exposed to the pre-reform rules. Depending on the preferred hours of work, the expected reform effects are qualitatively different. If she prefers to work less than half time, the substitution and income effects go in the same direction, inducing her to increase hours of work. In contrast, if she prefers to work half-time or more, the income effect from the increase of in-work benefits reduces labor supply. Overall, the work requirements and the increase of in-work benefits imply a subsidy to part-time work. The effect of the reform on average earnings will depend on the size of the different responses, weighted by the relative numbers of lone mothers at different points along the budget constraint. The effect of the reform on disposable income depends on the extent to which the loss of out-of-work benefits is offset by gains in earnings and higher in-work benefits for part-time workers. In particular, the disposable income will fall and the poverty rate will rise among lone mothers facing insurmountable employment barriers, who after the reform have lost eligibility to transitional benefit. C The editors of The Scandinavian Journal of Economics 2012.

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Fig. 2. Distribution for hours of work of lone and married mothers with the youngest child 4–9 years of age Notes: Panel A shows the distribution for hours of work in 1997 of lone and married mothers with the youngest child 4–9 years of age. Panel B shows the differences in population shares by hours of work between the lone mothers and married mothers in 2001, subtracting the differences in population shares by hours of work between the two groups in 1997. To compute the population shares of non-working mothers, we use administrative data on earnings for the entire population: zero hours of work is defined in accordance with our measure of labor force participation (see Section IV). To compute the distribution of hours of work among working mothers, we use information from the Wage Statistics Survey. This survey covers all employees in the public sector. For employees in the private sector, the data are based on an annual stratified random sampling of enterprises; all large enterprises, 40 percent of medium-sized enterprises, and less than 20 percent of small enterprises are sampled. Hours of work are defined as the contractual number of working hours per week, excluding meal breaks.

To get a perspective on the relative numbers of lone mothers at different points along the budget constraint, Panel A of Figure 2 shows the distributions for hours of work of lone and married mothers. However, it should be noted that the information on hours of work comes from survey C The editors of The Scandinavian Journal of Economics 2012.


data on an unrepresentative sample of women. With this caveat in mind, it seems that as many as 35 percent of all lone mothers were not participating in the labor market in 1997. However, we also see that the vast majority of employed lone mothers were working fairly long hours. In fact, as many as 88 percent of those who participated in the labor market were working at least half-time (19 hours per week), and almost 45 percent worked full-time (38 hours per week). In comparison, married mothers were participating in the labor market more often than lone mothers, although typically working fewer hours. Figure 1 does not reflect the introduction of welfare time limits and the reduction of the upper age limit for the youngest child to the welfare eligibility criteria. The long-term static effect of these measures is to eliminate welfare completely for certain lone mothers. This should increase labor supply for the same reasons that welfare decreases labor supply in the first place. In addition, there are some dynamic effects that unambiguously go in the same direction. First, lone mothers on welfare are expected to anticipate the date when benefits will run out and to begin to intensify their job search or even to accept job offers at an increasing rate when approaching this date.4 This implies that the time limits and the upper age limits do not have to be binding to affect the labor supply of welfare recipients. If there is uncertainty in terms of job opportunities or randomness in wage offers, it might be better to accept an offer that is, in the short run, less attractive than remaining on welfare, even if it arrives in advance of the date when benefits will run out. Furthermore, the introduction of time limits should provide incentives for recipients who might need welfare in the future to delay the use of welfare benefits, or to leave welfare as rapidly as possible, in order to preserve future eligibility.5 The total impact of the change in time and age limits on disposable incomes and poverty depends on whether the increases in earnings outweigh the loss of benefits. To get a sense of the likely responses to the workfare reform, we can compare the differences in the distributions for hours of work of lone mothers and married mothers, before and after the reform. Panel B of Figure 2 displays the differences in population shares by hours of work between lone mothers and married mothers in 2001, subtracting the differences in population shares by hours of work between the two groups in 1997. In line with the predictions from labor supply theory, lone mothers appear to have 4 Moffitt (1985) and Røed and Zhang (2005) find this behavior for the recipients of unemployment insurance when approaching the time their benefits will run out. 5 Grogger (2002), Grogger and Michalopoulos (2003), and Swann (2005) find that the introduction of time limits reduces welfare receipt substantially and that a significant part of this reduction occurs because recipients are forward-looking.


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Table 2. Participation rates and average benefit amount for the transitional benefit scheme, 1993–2001 Year

Participation rate on transitional benefits (percent)

Average monthly transitional benefit payment per recipient(€ – 1998)

1993 1994 1995 1996 1997 1998 1999 2000 2001

66 65 65 65 65 66 64 61 36

477 469 460 465 470 524 496 492 449

Notes: In each year, the sample consists of all lone mothers with the youngest child 4–9 years of age. The participation rate is defined as the fraction of the sample that receives transitional benefits. The transitional benefit amount is defined as the average monthly transitional benefit payment per recipient of transitional benefits.

increased their labor force participation but they have reduced part-time work with longer hours (31 hours per week), relative to married mothers. At the same time, the prevalence of both half-time work (19 hours) and full-time work (38 hours) seems to have increased. The relatively large increase in full-time work is consistent with institutional constraints on the choices for hours of work (e.g., because of the fixed costs of working).

Phase-in Provisions A final important feature of the transitional benefit reform is that phasein provisions were introduced so that a subgroup of lone mothers, who were entitled to and had applied for benefits by January 1, 1998, could continue to receive transitional benefits under the pre-reform rules. From January 1, 2001, benefits were paid exclusively according to the postreform rules. Note that women becoming lone mothers after January 1, 1998 were not entitled to the phase-in provisions. Table 2 displays the participation rates and average benefit payments for the transitional benefit scheme over the period 1993–2001. The time trends clearly mirror the fact that the reform was phased in gradually. The participation rates for lone mothers declined gradually after the reform in 1998, with a substantial drop in 2001 when the phase-in provisions were terminated. Furthermore, the average benefit payment declined in 2001; this conforms to intuition, because lone mothers were then exposed to work requirements and benefit payments are reduced when earnings increase. In our estimations, we pay close attention to the phase-in issue. C The editors of The Scandinavian Journal of Economics 2012.


III. Empirical Strategies In this section, we outline the empirical strategies that we have used to evaluate the effects of the workfare reform. First, we discuss the main empirical strategy, before describing the methods used to investigate heterogeneity in the effects of the reform. Finally, we describe how our estimations deal with the phase-in issue.

Main Empirical Strategy Our main empirical strategy is the following. We start by comparing the change in the outcome of interest from 1997 to 2001 for women (with children of the same age) who were either lone mothers or married mothers in both years. As discussed below, the reason for using 2001 as the last year of the comparison is that, although the reform was undertaken in 1998, it was three years before the policy changes were fully implemented. Because married and lone mothers might react differently to time-specific factors, such as changes in the labor-market conditions, we add a comparison of the change in the outcome from 1993 to 1997 for women who were either lone mothers or married mothers in both years. This gives us a trend-adjusted DD estimator, exploiting the fact that the reform creates variation along three dimensions: (a) between lone and married mothers; (b) between time periods before and after the reform (1997–2001 versus 1993–1997); (c) between the first year (1997 for the 1997–2001 period and 1993 for the 1993–1997 period) and the second year (2001 for the 1997–2001 period and 1997 for the 1993–1997 period). The trend-adjusted DD estimator can be defined as β = [(Y¯ 01 − Y¯ 97 |LONE = 1) − (Y¯ 01 − Y¯ 97 )|LONE = 0] −[(Y¯ 97 − Y¯ 93 |LONE = 1) − (Y¯ 97 − Y¯ 93 |LONE = 0)]

(1)

where Y¯ t is the average outcome of interest in year t, and LONE is a binary assignment indicator equal to 1 if the woman is a lone mother in both years and to 0 if she is a married mother in both years. The term in the first set of square brackets compares the time change in the average outcome for lone and married mothers in the time period after the reform. The term in the second set of square brackets makes the same comparison for the time period before the reform. Whereas the first term would be the standard DD estimator of the effect of the reform, the second term accounts for the possibility that lone and married mothers have some (unobserved) characteristics that make them react differently to time-specific factors. The identifying assumption is that the relative outcome of lone and married mothers would, on average, have changed in the same way in the period C The editors of The Scandinavian Journal of Economics 2012.

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after the reform as in the period before the reform, in the absence of the reform. A trend-adjusted DD regression can be expressed as Yi jt = α1 + α2 REFORM j + α3 SECONDt + α4 LONEij + α5 (REFORM j × SECONDt ) + α6 (REFORM j × LONEij ) + α7 (SECONDt × LONEij ) + β(REFORM j × SECONDt × LONEij ) + θ X i jt + εi jt ,

(2)

where i, t, and j are the indices for mother, year (1 = second; 0 = first), and time period (1 if the period is 1997–2001; 0 if the period is 1993–1997), respectively, X is the local unemployment rate for the area where the mother resides, and εijt is a composite residual consisting of an individual-specific fixed effect and a standard error term.6 The dummy variable SECOND is equal to 1 if the year is 2001 for the time period 1997–2001 or to 1997 for the time period 1993–1997, and to 0 if the year is 1997 for the time period 1997–2001 or 1993 for the time period 1993–1997. The dummy variable REFORM is equal to 1 if the time period is 1997–2001, and to 0 if the time period is 1993–1997. Equation (2) allows for different intercepts and time-specific effects for lone and married mothers. The effect of the reform is given by β, identified from the time change in the average outcome of lone mothers relative to married mothers, in the post-reform period relative to the pre-reform period. In comparison, the conventional DD regression imposes α 2 = α 5 = α 6 = β = 0, and it takes α 7 to be the effect of the reform. Previous studies of welfare reforms have generally applied a DD regression to repeated cross-sectional data.7 Because we have access to panel data, we improve on this by restricting the sample to the same lone and married mothers in the first and second years of each time period; this removes biases from comparison over time within each group because of unobserved compositional changes from the first year to the second year.

Heterogeneous Effects As discussed above, theory predicts systematic heterogeneity in the responses to the reform. This raises two concerns with the empirical strategy discussed above. 6 Bertrand et al. (2004) have shown that the standard errors in DD regressions might be mis-stated in the presence of a serial correlation of outcomes for the same mother over time. Although our fixed effects specification directly accounts for time-invariant unobserved heterogeneity, the fixed effects do not capture time-varying dependence. However, by limiting the time dimension to a few years, we reduce the problem of serial correlation considerably. 7 A notable exception is Francesconi and van der Klaauw (2007).



The first concern is that the estimated mean impacts might average together the effects of different sign and magnitude. In particular, lone mothers with little or no labor-market attachment are likely to be affected differently by the reform, compared to those with stronger attachments to the labor market. On the one hand, in order to gain eligibility to benefits, these mothers need to increase labor supply considerably. On the other hand, if they are unable to find work and thus lose eligibility, their disposable income will fall substantially. The most common way to address heterogeneous responses is to estimate mean impacts for subgroups thought to be likely to respond differently to the intervention. In line with this tradition, we estimate equation (2) separately according to the characteristics of the lone mothers, such as educational attainment, labor-market experience, and local labor-market conditions. The second concern is that the above empirical strategy might not necessarily produce a representative picture of the average effect for the entire population of lone mothers. In particular, equation (2) identifies the mean impact of the reform on women who had endured at least four years as lone mothers by 2001, hereafter labeled “lasting lone mothers”, and it is silent on the responses of women with shorter spells as lone mothers. In the spirit of Heckman (1991), there are two possible explanations for why the responses to the reform are likely to vary systematically with the length of time spent as a lone mother. The first explanation is that as a consequence of experiencing a long spell as a lone mother, preferences or the budget constraint relevant to future choices (or outcomes) are altered. In this case, time as a lone mother has a genuine behavioral effect, in the sense that a woman who has experienced a short spell would behave differently in the future than an otherwise identical woman who has experienced a longer spell as a lone mother. There are a number of different mechanisms that might explain such a difference. One is that being the sole caregiver for a child might, over time, result in a loss of motivation or a depreciation of human capital, which could make it less likely for the parent to find work. A similar mechanism is at work if longer spells as a lone mother aggravate social exclusion or stigma, thereby reducing labor-market opportunities. A second explanation is that women might differ in certain characteristics that influence their probability of experiencing a long spell as a lone mother, but these characteristics are not influenced by the length of the spell itself. In this case, time spent as a lone mother might affect the estimated effects of the reform because women who remarry (quickly) differ from those who remain single (longer). For example, in Becker’s influential model of the marriage market, it is less likely that low-skilled women (re)marry because they are less attractive as spouses. Hence, the C The editors of The Scandinavian Journal of Economics 2012.

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fraction of low-skilled lone mothers might be expected to rise with time spent in lone motherhood. Both these explanations suggest that the responses to the reform vary systematically with the length of time spent as a lone mother. Also, if the differences in the responses are not captured by some observable characteristic, the subsample estimation of equation (2) will fail to reveal the heterogeneous effects associated with time spent as a lone mother. Therefore, to complement the subsample estimation of equation (2), we use the following empirical strategy. We start by comparing the change in the outcome of interest from 1997 to 1999 of mothers who are married in 1997, split up in 1998, and are lone mothers in 1999, to the change in the outcome from 1995 to 1997 of mothers with children of the same age who are married in 1995, split up in 1996, and are lone mothers in 1997. We label these women as “newly lone mothers”, as opposed to the women who have experienced at least four years as lone mothers when faced with the post-reform rules.8 Because this strategy samples from the flow of lone mothers, it avoids oversampling those with long spells as lone mothers (i.e., length-biased sampling). The reasons for considering the year immediately after the married mother splits up and becomes a lone mother, and not the year of change itself, are that we have annual data only on the outcomes, and that we want to allow the women some time to adjust to their new situation. In order to take into account time-specific factors, such as economic fluctuations, which might otherwise confound the estimates of the effects of the reform, we add a comparison of the change in the outcome of married mothers who remain married from 1997 to 1999 with that of married mothers who remain married from 1995 to 1997. A DD estimator of the reform effect on newly lone mothers can be defined as ¯ 97 |NEW = 1) − (Y ¯ 97 − Y ¯ 95 |NEW = 1)] ¯ 99 − Y δ = [(Y ¯ 97 |NEW = 0) − (Y ¯ 97 − Y ¯ 95 |NEW = 0)] ¯ 99 − Y −[(Y

(3) where NEW is a binary assignment indicator equal to 1 if a mother who is married in the first year of the comparison (1997 or 1995) makes the transition to lone mother before the last year of the comparison (1999 or 1997), 8 Another difference between newly and lasting lone mothers is that the former group consists only of lone mothers who were formerly married or cohabiting. Because models of the marriage market predict that it is less likely that low-skilled women marry, it is likely that lone mothers who were formerly married or cohabiting have higher education and earnings potential. It should be noted, however, that the vast majority of Norwegian lone mothers were formerly married or cohabiting (e.g., Noack and Keilman, 1993), indicating that a large fraction of the lasting lone mothers were also formerly married or cohabiting. Unfortunately, our data do not allow us to identify the lasting lone mothers who were formerly married or cohabiting.



and to 0 if she remains married. The term in the first set of square brackets compares the time change in the average outcome of newly lone mothers in the time period before and after the reform. The term in the second set of square brackets makes the same comparison for married mothers remaining married. The impact of the reform on newly lone mothers is then identified as the effect of becoming a lone mother after the reform, relative to the effect before the reform. The identifying assumption is that the relative outcome of married mothers making the transition to lone motherhood and those remaining married would, on average, have changed in the same way in the period after the reform as in the period before the reform, in the absence of the reform. A DD regression of the reform effect on newly lone mothers can be expressed as Yi jt = γ1 + γ2 REFORM j + γ3 SECONDt + γ4 N E Wij + γ5 (REFORM j × SECONDt ) + γ6 (REFORM j × N E Wij ) + γ7 (SECONDt × N E Wij ) + δ(REFORM j × SECONDt × N E Wij ) (4) + λX i jt + u i jt , where i, t, and j are indices for mother, year (1 = second; 0 = first), and time period (1 if the period is 1997–1999; 0 if the period is 1995–1997), respectively, X is the local unemployment rate in the area where the mother resides, and uijt is a composite residual consisting of an individual-specific fixed effect and a standard error term. Taking advantage of our panel data, in each time period, we sample only mothers who are married in the first year and either remain married or make the transition to lone motherhood before the second year. The dummy variable SECOND is equal to 1 if the year is 1999 for the time period 1997–1999 or 1997 for the time period 1995–1997, and to 0 if the year is 1997 for the time period 1997–1999 or 1995 for the time period 1995–1997. The dummy variable REFORM is equal to 1 if the time period is 1997–1999, and to 0 if the time period is 1995–1997. Equation (4) allows for different intercepts and time-specific effects for married mothers making the transition into lone motherhood and those remaining married. The reform effect is given by δ, identified from the time change in the average outcome of the two groups, in the period after the reform relative to the period before the reform.

Phase-in Period A complicating issue in many program evaluations using a DD approach is that welfare reforms are seldom retroactive, so temporary provisions are often introduced during a phase-in period. In this phase-in period, some or C The editors of The Scandinavian Journal of Economics 2012.

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all welfare recipients might continue to receive benefits according to prereform rules or to receive some form of compensation for inconveniences caused by the reform. This blurs the before and after distinction that forms the basis for the DD methods. While several past program evaluations employing the DD approach have simply ignored the potentially confounding effects of a gradual phase-in of reforms, Blundell et al. (2005) decided to exclude observations from a one-year phase-in period in their evaluation of the reform of the Working Families’ Tax Credit in the UK. In our case, phase-in provisions were introduced so that lone mothers who had applied for and who were entitled to benefits before 1998 could continue to receive these under the pre-reform rules for up to three years. As suggested by Table 2, the phase-in period provides limited information about the incentive effects of the reform on lasting lone mothers. To circumvent the phase-in issue, we exclude the observations from the phase-in years 1998, 1999, and 2000 when estimating the effects of the reform on lasting lone mothers. In comparison, the empirical strategy for newly lone mothers has the advantage of avoiding the phase-in issue, as newly lone mothers in 1999 will not be entitled to phase-in provisions.

IV. Data and Definitions The empirical analysis is based on administrative registers covering the entire resident population of Norway in the period 1993–2001. The register panel dataset with household and demographic information is merged with detailed income data from the Tax Assessment Files through unique individual identifiers. The income data are collected from tax records and other administrative registers rather from than interviews and self-assessment methods. The coverage and reliability of Norwegian register data are considered to be exceptional, as is documented by the fact that the quality of such national datasets received the highest rating in a data quality survey in the Luxembourg Income Study database (Atkinson et al., 1995). Our empirical analysis focuses on lone mothers who are at least 18 years old and not more than 55, whose youngest child is between 4 and 9 years of age. In our baseline specification, our comparison group consists of married mothers who are between 18 and 55 years of age, whose youngest child is between 4 and 9 years of age. In one of the robustness checks, we also consider single women in the same age range. Following closely the previous body of literature, students and the self-employed, as well as individuals receiving permanent disability benefits, are dropped from the analysis.9

9

Eissa and Liebman (1996) and Francesconi and van der Klaauw (2007) have used similar sample selection criteria in their reform evaluations of lone-parent benefits. C The editors of The Scandinavian Journal of Economics 2012.


To account for variations in local labor-market conditions, we make use of data on local unemployment rates for 90 economic regions. Specifically, the economic regions constitute a regional level between country and municipality. The main criteria used for defining the economic regions are labor-market, trade, and service patterns, as well as commuting and internal migration patterns. Allowing economic regions rather than municipalities form the basis for measuring unemployment rates could provide a better predictor of local labor-market conditions. The dependent variables are defined as follows. Our measure of earnings is defined as the woman’s annual gross earnings. Disposable income is defined in close agreement with international recommendations (see Expert Group on Household Income Statistics, 2001) and it incorporates the woman’s (but not her spouse’s) annual wages, capital income, and all public cash transfers, less taxes. To evaluate the effects of the reform on annual gross earnings and disposable income, we use the consumer price index to make earnings and incomes from different periods comparable; throughout this paper, the reference year is 1998, and €1 is set equal to NOK 8.4. The fixed time-specific effects account for general income and earnings growth. In addition, we construct a measure of labor-market participation based on the basic amount thresholds for earnings used by the Norwegian Social Insurance Scheme to determine labor-market status (in order to determine eligibility for unemployment benefits, disability benefits, and old-age pension). Specifically, a mother is defined as working if her annual earnings exceed one basic amount (in 1998, about €5,300). Our final outcome is poverty, in which case we follow common practice and we define the annual poverty thresholds as 50 percent of the median in the distribution of household equivalent disposable income. To enable comparison of household disposable income between individuals belonging to households of varying size and composition, the OECD equivalence scale is applied; the weight of the first adult in the household is set to 1, each additional adult receives a weight of 0.7, and each child receives a weight equal to 0.5. In a given year, a woman is defined as poor if her household equivalent income is lower than the poverty threshold. The choices of poverty threshold and equivalence scale correspond to Norwegian official poverty statistics, as well as the 2002 Poverty White Paper (Ministry of Social Affairs, 2002).

Descriptive Statistics Figure 3 shows labor-market participation and mean earnings by year, among lone mothers, married mothers, and single women without children. As expected, labor-market participation and mean earnings are considerably lower for lone mothers, compared to married mothers and single women. C The editors of The Scandinavian Journal of Economics 2012.

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Fig. 3. Labor-market participation and mean earnings of lone mothers, married mothers, and single women Notes: In each year, the sample consists of lone and married mothers with the youngest child between 4 and 9 years of age, as well as single women without children.

We also see a fairly good coherence between the time trends of lone and married mothers before the reform, whereas single women display a substantially different time trend in labor-market participation. For this reason, married mothers appear to be the most suitable comparison group. Nevertheless, to take into account that the pre-reform trend in labor-market participation is steeper for lone mothers, we apply trend-adjusted DD estimators. In this regard, it is reassuring that the pre-reform data establish a clear trend that can be extrapolated to the post-reform period. From Figure 3, we further see that the gap in labor-market participation between lone and married mothers is reduced significantly after the reform, whereas the change in the relative earnings of the two groups is less distinct. This conforms to the theoretical predictions because the reform subsidizes part-time work; therefore, labor-market participation should C The editors of The Scandinavian Journal of Economics 2012.

Are lone mothers responsive to policy changes? 1145 Table 3. Pre-reform and post-reform descriptive statistics of married and lasting lone mothers Difference Lone mothers – Married mothers

Outcomes (mean) Earnings (€ – 1998) Labor-market part. (%) Disposable income (€ – 1998) Poverty (%) Characteristics (mean) Age Years of schooling Labor-market experience Non-western immigrant (%) Number of children Age of youngest child Unemployment rate (%) Observations Pop. share of lone mothers

Level Lone mothers Pre-reform First year (1993)

First year (1993)

Second year (1997)

First year (1997)

Second year (2001)

DD Estimate (No controls)

9,958

−4,101

−4,220

−4,523

−3,514

1,128∗ ∗ ∗

52.6

−22.8

−17.2

−20.2

−10.3

4.3∗ ∗ ∗

17,676 4.7

5,123 2.0

5,447 2.8

6,222 3.3

5,720 7.6

−826∗ ∗∗ 3.5∗ ∗ ∗

29.9

−3.4

−3.4

−2.9

−2.9

0.0

11.1

−1.0

−1.0

−1.1

−1.1

0.0

7.7

−3.3

−4.1

−3.7

−4.2

0.3

2.6

0.4

0.4

1.0

1.0

0.0

1.6

−0.7

−0.7

−0.6

−0.6

0.0

3.0

0.2

0.3

0.4

0.4

4.3 10,992 0.122

0.1 90,171 0.122

0.1 90,171 0.113

0.1 93,116 0.113

0.1 93,116 0.117

Pre-reform

Post-reform

−0.1 0.0 366,574

Notes: The pre-reform sample consists of women who were either lone mothers or married mothers in both 1993 (first year) and 1997 (second year). The post-reform sample consists of women who were either lone mothers or married mothers in both 1997 (first year) and 2001 (second year). Level refers to means for lone mothers in the first year of the pre-reform period. Difference refers to differences in means between lone and married mothers in the first/second year in the pre-reform/post-reform period. Labor market experience is defined as years of employment in previous years, according to the standard of the National Insurance Administration. DD estimate is based on equation (1). ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

unambiguously increase, whereas the mean impact on earnings could be averaging together positive effects on low-earning lone mothers and possibly negative effects on high-earning lone mothers. Table 3 displays detailed descriptive statistics for the sample of lasting lone mothers, which forms the basis for the estimation of equation (2). We can see that lone mothers have substantially lower earnings and labormarket participation than married mothers, but at the same time they have higher individual disposable income, simply because of the generous welfare system for lone parents. Yet, lone mothers are more prone to poverty (based on equivalent household disposable income, in line with common C The editors of The Scandinavian Journal of Economics 2012.

1146


practice and Norwegian official poverty statistics), because lone parent benefits do not appear to fully compensate for being in a single-earner family rather than a dual-earner family. More importantly, it is evident from Table 3 that there are distinct differences before and after the reform in the changes over time in the outcomes of lone and married mothers. In our trend-adjusted DD framework laid out above, this suggests substantial effects of the reform on lasting lone mothers (see Column 6). For example, the difference in labor-market participation between married and lasting lone mothers is considerably smaller in the second year than in the first year, in the period after the reform relative to the period before the reform. Specifically, equation (1) suggests that the workfare reform increased the labor-market participation of lasting lone mothers by 4.3 percentage points. Table 3 also shows descriptive characteristics of married and lone mothers before and after the reform. As is clear from Table 3, married mothers are, on average, older and have more children, higher education, and more labor-market experience than lasting lone mothers. However, we are not too concerned with differences in the characteristics of these two groups, per se. Our primary concern is that there could be differences in how timevarying characteristics of lone and married mothers change over time, in the period after the reform relative to the period before the reform. Therefore, it is reassuring that none of the variables listed in Table 3 raises such concerns. Consistent with this, there are no significant effects of the reform when the time-varying characteristics are used as the dependent variable in the DD estimation (see Column 6). Table 4 shows the same descriptive statistics as Table 3, except that newly lone mothers replace lasting lone mothers. Again, there are signs of substantial reform effects (see Column 6). For example, equation (3) suggests that the workfare reform increased labor-market participation of newly lone mothers by 2.2 percentage points. It is also evident that married mothers are, on average, older, and that they have slightly more children and somewhat higher labor-market experience and education than the newly lone mothers. However, the time-varying characteristics of lone and married mothers change very little over time, in the period after the reform relative to the period before the reform. Consistent with this, there are no significant effects of the reform when the time-varying characteristics are used as the dependent variable in the DD estimation (see Column 6). When comparing Tables 3 and 4, it is clear that the lasting lone mothers are younger and that they have considerably lower education and less labormarket experience than the newly lone mothers. In fact, the newly lone mothers have quite similar labor-market participation and earnings to those of married mothers (see Column 3, which displays the second-year averages before the reform but after the woman has become a lone mother). C The editors of The Scandinavian Journal of Economics 2012.

Are lone mothers responsive to policy changes? 1147 Table 4. Pre-reform and post-reform descriptive statistics of married mothers and newly lone mothers

Outcomes (mean) Earnings (€ – 1998) Labor market part. (%) Disposable income (€ – 1998) Poverty (%) Characteristics (mean) Age Years of schooling Labor market experience Non-western immigrant (%) Number of children Age of the youngest child Unemployment rate (%) Observations Pop. share of lone mothers

Difference Lone mothers – Married mothers

Level Lone mothers Pre-reform First year (1995)

First year (1995)

Second year (1997)

First year (1997)

Second year (1999)

DD Estimate (No controls)

16962

972

368

486

289

407∗

82,1

1,9

−2,4

1,8

−0,3

2,2∗ ∗

15762

1528

8530

225

9674

2447∗ ∗ ∗

2,3

−0,1

2,3

0,7

2,2

−0,9

33

−2,3

−2,3

−2,2

−2,2

0,0

11,8

−0,3

−0,3

−0,4

−0,4

0,0

11

−1,5

−1,6

−1,8

−1,7

0,2

2,4

0

0

0,1

0,1

0,0

2,1

−0,3

−0,3

−0,2

−0,2

0,0

6,6

−0,2

−0,2

−0,2

−0,2

0,0

3,6 2,276

0 83,186 0,027

0 83,186 0,027

0,1 87,402 0,027

0 87,402 0,027

−0,1 341,176 0,027

Pre-reform

Post-reform

Notes: The pre-reform sample of lone mothers consists of women who were married in 1995 (first year), split up in 1996, and are lone mothers in 1997 (second year). The pre-reform sample of married mothers consists of mothers who stay married from 1995 (first year) to 1997 (second year). The post-reform sample of lone mothers consists of women who were married in 1997 (first year), split up in 1998, and are lone mothers in 1999 (second year). The post-reform sample of married mothers consists of mothers who stay married from 1997 (first year) to 1999 (second year). Level refers to the means of lone mothers in first year of the pre-reform period. Difference refers to differences between the means of lone and married mothers in first/second year in pre-reform/post-reform periods. Labor market experience is defined as years of employment in previous years, according to the standard of the National Insurance Administration. DD estimate is based on equation (3). ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

V. Empirical Results In this section, we investigate the responses of lone mothers to the reform. First, we discuss our main results, before reporting estimates of the effects of the reform during the phase-in period and the results from several robustness checks. C The editors of The Scandinavian Journal of Economics 2012.

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Table 5. Estimated reform effects on newly and lasting lone mothers Panel A: Lasting lone mothers

Panel B: Newly lone mothers

Dependent variable

Estimate

Mean

Obs.

Estimate

Mean

Obs.

Earnings (€ – 1998) Labor-market part.(perc. points) Disposable income (€ – 1998) Poverty (perc. points)

1,116∗∗∗ (128) 4.3∗∗∗ (0.7) −844∗∗∗ (114) 3.5∗∗∗ (0.4)

13,872

366,574

18,557

341,176

67.0

366,574

82.1

341,176

21,974

366,574

25,053

341,176

4.7

366,574

406∗ (207) 2.2∗∗ (1.0) 2,447∗∗∗ (205) −0.9 (0.7)

4.3

341,176

Notes: All estimations include individual-specific and time-specific fixed effects, as well as controls for the local unemployment rate in the area where the mother resides. Outcomes are defined in Section IV. The standard errors in parentheses are robust to heteroskedasticity and they are clustered at the individual level. Mean refers to average outcome of lone mothers in the second year in the pre-reform period (1997). In Panel A, estimations are based on OLS on equation (2), with 1993–1997 as the pre-reform period and 1997–2001 as the post-reform period. In Panel B, estimations are based on OLS on equation (4), with 1995–1997 as the pre-reform period and 1997–1999 as the post-reform period. ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

Main Results Table 5 shows the estimated mean impact of the workfare reform on earnings, labor-market participation, disposable income, and poverty of lasting and newly lone mothers. As discussed above, the reform is expected to stimulate the earnings of lone mothers as long as the positive effects of the time and the age limits, as well as the work requirements, dominate the negative effect induced by the increase of in-work benefit levels for those working at least part-time. Indeed, Table 5 shows positive and significant effects of the reform on earnings and labor-market participation. Specifically, the reform led to an increase of 2.2 percentage points in the labor-market participation of newly lone mothers, and to a €406 increase in their earnings. In comparison, the reform is estimated to have had a substantially larger impact on lasting lone mothers. Specifically, the reform increased their earnings by €1,116, and their labor-market participation by 4.3 percentage points. To get a perspective on the magnitudes, these parameter estimates suggest that the workfare reform reduced the gaps between married mothers and lasting lone mothers in labor-market participation and in earnings by around 25 percent. The differences in the labor-market responses of newly and lasting lone mother are consistent with the theoretical predictions. Because the reform essentially subsidizes part-time work, the estimated mean impacts depend on the relative number of lone mothers at different points along the budget constraint (as well as their elasticity). Because the lasting lone mothers had a relatively weak labor-market attachment before the reform, on average, C The editors of The Scandinavian Journal of Economics 2012.


they have stronger incentives to increase their labor supply in response to the policy changes, compared to newly lone mothers. By computing the weighted average of the estimated effects of the reform on lasting and new lone mothers, we find that the aggregate mean effects are around €985 for earnings and 3.9 percentage points for labor-market participation. The relatively large overall effects reflect that there are nearly five times as many lasting lone mothers as newly lone mothers. In comparison to the evidence from the Anglo-Saxon countries, our results on labor-market participation are fairly similar. For example, Francesconi and van der Klaauw (2007) have estimated that the reform of the Working Families’ Tax Credit in the UK increased the employment of lone mothers by around 5 percentage points. Moffitt (2007) has reported that employment increased by about 4 percent as a result of the US workfare reform in 1996. However, our estimated effects of the reform on earnings appear to be somewhat lower than is typically found in studies of similar reforms in Canada, the UK, and the US (e.g., Moffitt, 2007). Turning our attention to disposable income and poverty, the last two rows of Table 5 suggest a qualitative difference in the effects of the reform on newly and lasting lone mothers. On the one hand, the policy changes led to a substantial increase in the disposable income of newly lone mothers, in part because of higher earnings but also because of higher in-work benefit levels. On the other hand, the workfare reform caused a considerable decrease in disposable income and an economically significant increase in poverty among lasting lone mothers. The reason for this is that a sizeable group was unable to offset the loss of out-of-work benefits with gains in earnings, even though lone mothers were offered highly subsidized childcare, after-school programs were widely available, and the economy was fairly strong. This suggests that the desired effects of the workfare reform on earnings and labor-market attachment are associated with income loss and increased poverty among a subgroup of lasting lone mothers who face overwhelming employment barriers. The difference in the estimated effects of the reform on poverty and disposable income align well with the fact that lasting lone mothers have considerably less labor-market experience and lower education than newly lone mothers. The weighted averages of the effects of the reform on lasting and newly lone mothers are around €–241 for income and 2.7 percentage points for poverty. In contrast, studies of workfare reforms in Anglo-Saxon countries have suggested positive effects on income, and reductions in poverty.10 10 For example, see Blank and Schoeni (2003), who investigated changes in the distribution of family income in the US during the 1990s. Their findings suggest that the welfare reforms led to a rise in family income and a decline in poverty, and they are consistent with the study by Meyer and Sullivan (2004) of the consumption pattern of lone mothers during this period.


1150


In Table 6, we investigate to what extent the differences in their observable characteristics capture the differences in the effects of the reform between lasting and newly lone mothers. We start by estimating equation (2) separately according to characteristics of the lasting lone mothers, including educational attainment, labor-market experience, and local unemployment rates. The subsample results reported in Columns 1–6 demonstrate that there is considerable heterogeneity in the responses of lasting lone mothers to the reform, in line with the theoretical predictions. Moreover, they indicate that at least some of the differential effects between newly and lasting lone mothers could be attributable to differences in observable characteristics. Specifically, the increases in labor-market participation and earnings are strongest among lasting lone mothers with relatively low education or little labor-market experience. At the same time, these subgroups of lone mothers have experienced the largest income loss and rise in poverty incidence, most likely because a considerable proportion could not find work and thus they lost benefits. Also, as expected, the reform was most successful in areas of low unemployment, where the rise in labor-market participation and earnings were the largest and the income loss and increased poverty were the smallest. To overcome the curse of dimensionality that we would face if we were to use several characteristics to define subgroups, we use a standard survey weighting method to make lasting lone mothers similar to newly lone mothers in all observable characteristics (e.g., Yansaneh, 2005). The sampling weights are equal to the inverse probability of sampling a lasting lone mother (or a married mother) with a given set of characteristics (reported in Tables 3 and 4), in the period before and after the reform. The last column of Table 6 estimates equation (2) in the reweighted sample of lasting lone mothers. The results show that the differential effects between newly and lasting lone mothers are reduced, but they are far from eliminated by the reweighting procedure. In particular, the evidence continues to suggest that the reform led to income loss and a rise in the incidence of poverty among lasting lone mothers.

Phase-in Issue As discussed in detail above, the workfare reform was gradually phased in, so that lone mothers who had applied for, and were entitled to, benefits before 1998 could continue to receive them under pre-reform rules for up to three years. While several past program evaluations have ignored the potentially confounding effects of a gradual phase-in of reforms, here we have followed Blundell et al. (2005) in discounting the observations from the phase-in years when estimating the effects of the reform on lasting lone mothers. In comparison, the empirical strategy for newly lone mothers C The editors of The Scandinavian Journal of Economics 2012.

Newly lone mothers 11.8 12.6 2.6 35 2.4 2.1 6.6

1,324∗∗∗ (186) 5.3∗∗∗ (1.1) −1,228∗∗∗ (143) 5.1∗∗∗ (0.7) 118,594

(1) Low educated

Low educated 8.9 10.5 2.4 33.8 6.3 2.3 6.9

885∗∗∗ (180) 3.4∗∗∗ (0.9) −560∗∗∗ (177) 2.4∗∗∗ (0.5) 247,980

(2) High educated

High educated 13.7 13.9 2.3 35.9 1.2 2.2 6.8

1,197∗∗∗ (187) 4.2∗∗∗ (1.2) −1,133∗∗∗ (197) 5.8∗∗∗ (0.8) 97,592

(3) Low labormarket exp. 1,110∗∗∗ (155) 4.7∗∗∗ (0.9) −852∗∗∗ (124) 3.5∗∗∗ (0.5) 274,594

(5) Low local unempl. rate

Low labormarket exp. 10.6 5.2 2.4 32.1 8.3 2.4 6.7

High labormarket exp. 12.7 15.5 2.3 36.3 0.9 2.2 6.9

Lasting lone mothers

672∗∗∗ (179) 1.7∗∗ (0.8) −673∗∗∗ (135) 2.0∗∗∗ (0.4) 268,982

(4) High labormarket exp.

Low local unempl. rate 12.2 13 2.1 35.3 2.8 2.3 6.8

1,205∗∗∗ (301) 2.8 (1.7) −1,234∗∗∗ (243) 4.8∗∗∗ (1.1) 91,980

(6) High local unempl. rate

High local unempl. rate 12 12.2 3.1 34.9 3 2.3 6.8

975∗∗∗ (181) 3.0∗∗∗ (0.8) −661∗∗∗ (165) 2.6∗∗∗ (0.5) 369,574

(7) Reweighted sample

Reweighted sample 11.9 12.1 2.8 35.8 2.2 2.1 6.6

Notes: In the upper panel, each cell reports the effect of the reform from a separate regression. The regressions are based on OLS on equation (2), with 1993–1997 as the pre-reform period and 1997–2001 as the post-reform period. Columns 1–6 report subsample results. For each characteristic, the sample is split into two groups, above and below the median in the distribution of lone mothers in the first year in the pre-reform period. Column 7 estimates equation (2) on the reweighted sample. All estimations include individual-specific and time-specific fixed effects, as well as controls for the local unemployment rate in the area where the mother resides. Outcomes are defined in Section IV. The standard errors in parentheses are robust to heteroskedasticity and they are clustered at the individual level. The lower panel reports the characteristics in 1997 of newly lone mothers, the subsamples of lasting lone mothers, and the reweighted sample of lone mothers. ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

Characteristics (mean) Years of schooling Labor-market experience Unemployment rate (%) Age Non-western immigrant (%) Number of children Age of the youngest child

Obs.

Labor market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points)

Earnings (€ – 1998)

Dependent variable

Table 6. Heterogeneity in the effects of the reform among lasting lone mothers



1152


Table 7. Estimates of the effects of the reform on newly and lasting lone mothers in the phase-in period Lasting lone mothers Last year in post-reform period Panel A: 1998 Earnings (€ – 1998) Labor-market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points) Panel B: 1999 Earnings (€ – 1998) Labor-market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points) Panel C: 2000 Earnings (€ – 1998) Labor-market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points)

Newly lone mothers

Estimate

Mean

Obs.

Estimate

Mean

Obs.

193∗∗∗ (57) 0.6∗ (0.3) 260 (309) −0.2 (0.2)

15,810

425,732

18,557

338,360

73.5

425,732

82.1

338,360

23,183

425,732

25,053

338,360

4.5

425,732

445∗∗ (217) 1.6 (1.0) 2,029∗∗∗ (206) −1.1 (0.7)

4.3

338,360

337∗∗∗ (84) 1.3∗∗∗ (0.5) 112 (78) −0.3 (0.3)

15,116

396,746

18,557

341,176

71.1

396,746

82.1

341,176

22,633

396,746

25,053

341,176

4.3

396,746

406∗ (207) 2.2∗∗ (1.0) 2,447∗∗∗ (205) −0.9 (0.7)

4.3

341,176

413∗∗∗ (105) 2.4∗∗∗ (0.6) −247∗ (130) 0.1 (0.3)

14,510

379,836

18,557

340,856

68.9

379,836

82.1

340,856

22,396

379,836

25,053

340,856

4.4

379,836

428∗∗ (207) 1.5∗ (0.9) 2,193∗∗∗ (247) −0.1 (0.7)

4.3

340,856

Notes: All estimations include individual-specific and time-specific fixed effects, as well as controls for the local unemployment rate in the area where the mother resides. Outcomes are defined in Section IV. The standard errors in parentheses are robust to heteroskedasticity and they are clustered at the individual level. Mean refers to the average outcome of lone mothers in the second year in the pre-reform period (1997). Lasting lone mothers: all estimations are based on OLS on equation (2). In Panel A, 1996–1997 is the pre-reform period and 1997–1998 is the post-reform period. In Panel B, 1995–1997 is the pre-reform period and 1997–1999 is the post-reform period. In Panel C, 1994–1997 is the pre-reform period and 1997–2000 is the post-reform period. Newly lone mothers: all estimations are based on OLS on equation (4), with 1995–1997 as the pre-reform period. In Panel A, 1996–1998 is the post-reform period. In Panel B, 1997–1999 is the post-reform period. In Panel C, 1998–2000 is the post-reform period. ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

sidesteps these issues, as they were not entitled to phase-in provisions if they split up after the reform. Table 7 reports estimates of the effects of the reform on lasting lone mothers in 1998, 1999, and 2000. There is a clear suggestion that we C The editors of The Scandinavian Journal of Economics 2012.


would seriously underestimate the impact of the policy changes if we were to disregard, or if we were unaware of, the gradual phase-in of the reform. As expected, the effects of the reform on earnings and labor-market participation increase as the end of the phase-in period (January 1, 2001) approaches. From a policy point of view, it is also interesting to see that there are no detrimental effects of the reform on poverty and disposable income among lasting lone mothers during the phase-in period, when they had the choice to self-select into the new system with higher inwork benefits or to remain in the old system without work requirements. Table 7 also displays the effects of the reform on newly lone mothers in the phase-in years; there is strong consistency over time in the estimates. This finding illustrates that it is unnecessary to wait until the phase-in period is over to evaluate the effects of the reform on newly lone mothers. This can be attractive from a policy perspective. Also, the consistency in the estimates over time increases our confidence in this empirical strategy.

Robustness Analysis In Table 8, we investigate the robustness of our baseline results, reported in Table 5, to changes in the specification of control variables. It is reassuring to find that the results from these specification checks are qualitatively the same as the baseline results and, moreover, that the point estimates are generally quite similar. Column 2 reports the estimates when dropping the control for local unemployment rates, and Column 1 reports our baseline results from Table 5 for comparison. In Column 3, we address the concern that other changes in the local environment might bias our estimates of the effects of the reform. For each municipality in every year, we have included detailed demographic controls for age and gender composition, local government spending, disposable income per capita, and the population share of full-time employed. Column 4 demonstrates that our results are robust to relaxing the linearity restriction in local unemployment rates, controlling for local unemployment rate and local unemployment rate squared. In Column 5, we also control for the interaction between local unemployment rate and lone mother status, as well as interactions between the local unemployment rate and time fixed effects. This specification check means that the effect of being a lone mother is allowed to differ, depending on local unemployment rates; also, the effect of economic shocks in the country is allowed to differ, depending on local labor-market conditions. Finally, Column 6 controls for interactions between the local unemployment rate and the age, education, and labor-market experience of individuals. This specification check allows unemployment rates to affect different groups in different ways. C The editors of The Scandinavian Journal of Economics 2012.

1154


Table 8. Robustness analysis of changes in the set of control variables (1) (2) Baseline No specification controls Dependent variable Earnings (€ – 1998) Labor-market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points)

(6) Controls: unempl. interactions

Panel A: lasting lone mothers 1,116∗∗∗ (128) 4.3∗∗∗ (0.7) −844∗∗∗ (114) 3.5∗∗∗ (0.4)

Dependent variable Earnings (€ – 1998) Labor-market part. (perc. points) Disposable income (€ – 1998) Poverty (perc. points)

(4) (5) (3) Quadratic Lone mother: More unempl. unempl. controls rate interactions

1,128∗∗∗ 1,125∗∗∗ 1,089∗∗∗ (128) (129) (128) 4.3∗∗∗ 4.2∗∗∗ 4.3∗∗∗ (0.7) (0.7) (0.7) −826∗∗∗ −832∗∗∗ −880∗∗∗ (114) (114) (114) 3.5∗∗∗ 3.5∗∗∗ 3.5∗∗∗ (0.4) (0.4) (0.4)

1,113∗∗∗ (136) 4.4∗∗∗ (0.7) −898∗∗∗ (116) 3.8∗∗∗ (0.4)

1,059∗∗∗ (125) 5.4∗∗∗ (0.7) −896∗∗∗ (114) 3.3∗∗∗ (0.4)

124 (220) 1.1 (1.1) 2,135∗∗∗ (217) −1.0 (0.7)

294 (204) 2.0∗∗ (1.0) 2,430∗∗∗ (205) −0.9 (0.7)

Panel B: newly lone mothers 406∗ (207) 2.2∗∗ (1.0) 2,447∗∗∗ (205) −0.9 (0.7)

407∗ 405∗ (208) (207) 2.2∗∗ 2.2∗∗ (1.0) (1.0) 2,447∗∗∗ 2,448∗∗∗ (205) (205) −0.9 −0.9 (0.7) (0.7)

402∗ (207) 2.2∗∗ (1.0) 2,444∗∗∗ (205) −0.9 (0.7)

Notes: Each cell in every panel reports the effect of the reform from a separate regression. Column 1 repeats the results from Table 5. Column 2 drops the control for local unemployment rate. Column 3 adds municipality–year controls for age (five-year age groups) and gender composition, the proportion of population employed full-time, mean disposable income, and mean local government expenditure. Column 4 adds a quadratic term in the local unemployment rate. Column 5 controls for the interaction of the local unemployment rate with lone-mother status and time fixed effects. Column 6 controls for the interaction of the local unemployment rate with age, education, and labor-market experience. The standard errors in parentheses are robust to heteroskedasticity and they are clustered at the individual level. ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

To further increase the confidence in our results, in Table 9, we perform several other specification checks. In Column 2, we drop the trend adjustment in the DD regressions (i.e., the second bracket of equations (1) and (3)). In Panel A, this specification check examines how our estimates of the effects of the reform are affected by assuming that married and lone mothers react in the same way to time-specific factors, such as changes in macroeconomic conditions. In Panel B, this specification check corresponds to restricting the time-specific effects to be the same for lone mothers before and after the reform. In both panels, we see that the results are qualitatively the same as the baseline results. However, the point estimates differ somewhat, in particular for the labor-market participation of lasting lone mothers. The reason for this is that our baseline specification, unlike the standard DD approach, takes into account the fact that the C The editors of The Scandinavian Journal of Economics 2012.

Are lone mothers responsive to policy changes? 1155 Table 9. Specification checks (2) (3) (4) No control Additional comp. Alternative comp. (5) (6) (1) for pregroup: lone group: single Pre-reform LogBaseline reform mother, child aged women, placebo dependent specif. trend 10 − 14 no child test variable Dependent variable

Panel A: lasting lone mothers ∗∗∗

Earnings 1,116 (€ – 1998) (128) Labor market 4.3∗ ∗ ∗ part. (perc. points) (0.7) Disposable −844∗ ∗ ∗ income (€ – 1998) (114) Poverty (perc. 3.5∗ ∗ ∗ points) (0.4)

∗∗∗

1,004 (99) 10.1∗ ∗ ∗ (0.5) −530∗ ∗ ∗ (89) 4.3∗ ∗ ∗ (0.3)

Dependent variable

1,192∗ ∗ ∗ (128) 4.8∗ ∗ ∗ (0.7) −800∗ ∗∗ (113) 3.4∗ ∗ ∗ (0.4)

1,421∗ ∗ ∗ (139) 4.5∗ ∗ ∗ (0.7) −797∗ ∗ ∗ (124) 2.8∗ ∗ ∗ (0.4)

116 (95) 0.9 (0.6) 703 (717) 0.4 (0.3)

0.045∗ ∗ (0.021) −0.132∗ ∗∗ (0.006)

Panel B: newly lone mothers ∗

Earnings 406 (207) (€ – 1998) Labor market 2.2∗ ∗ part. (perc.points) (1.0) Disposable 2,447∗ ∗ ∗ income (€ – 1998) (205) Poverty (perc. −0.9 points) (0.7)

∗∗∗

636 (244) 3.3∗ ∗ ∗ (1.0) 1,622∗ ∗ ∗ (278) −0.3 (0.6)

406∗ (208) 2.2∗ ∗ (1.0) 2,433∗ ∗ ∗ (205) −0.9 (0.7)

573∗ ∗ ∗ (211) 2.7∗ ∗ ∗ (1.0) 2,287∗ ∗ ∗ (481) −1.2∗ (0.7)

−107 (196) 0.2 (1.0) 346 (214) 1.2 (0.7)

0.040 (0.026) 0.061∗ ∗ ∗ (0.017)

Notes: Each cell in every panel reports the effect of the reform from a separate regression. Column 1 repeats the results from Table 5. Column 2 of Panel A drops the observations of women who were lone and married mothers in 1993 and 1997 from the estimation of equation (2), thereby imposing α 2 = α 5 = α 6 = β = 0 and taking α 7 to be the effect of the reform (see first bracket, equation (1)). Column 2 of Panel A drops the observations of married mothers from the estimation of equation (4), thereby imposing γ 4 = γ 6 = γ 7 = δ = 0 and taking γ 5 to be the effect of the reform (see first bracket, equation (3)). Column 3 adds lone mothers with the youngest child aged 10–14 years as an additional comparison group. Column 4 uses single women without children as the comparison group, instead of married mothers. Column 5 of Panel A estimates equation (2) with 1993–1996 as the pre-reform period, and 1994–1997 as the post-reform period. Column 5 of Panel B estimates equation (4) with 1993–1995 as the pre-reform period, and 1995–1997 as the post-reform period. Column 6 measures the continuous dependent variables in logs rather than levels, excluding observations with zero earnings or zero income. The standard errors in parentheses are robust to heteroskedasticity and they are clustered at the individual level. ∗∗∗ , ∗∗ , and ∗ denote significance at the 1, 5, and 10 percent levels, respectively.

pre-reform trend in labor-market participation is steeper for lone mothers than for married mothers. In Column 3, we perform a triple-difference analysis, adding lone mothers with older children (aged 10–14) as an additional comparison group; we exploit the fact that these were not affected by the reform. The comparison group of married mothers enables us to control for time-varying changes common to mothers with the youngest child aged 4–9 years, whereas the comparison group of lone mothers with older children is used to control for time-varying changes common to lone mothers. The fact that our results are robust to adding the comparison group of lone mothers with older children indicates that changes in, for example, the macroeconomic conditions that affected married and single mothers differently are an C The editors of The Scandinavian Journal of Economics 2012.

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unlikely source of bias in our estimates. In Column 4, we use single women without children as the comparison group, instead of married mothers. It is encouraging to find that the estimated effects of the reform are quite similar. As a further specification check, in Column 5, we pretend that the workfare reform took place before the actual implementation, and then we estimate the effects of this placebo reform using data from the pre-reform period. If we find an effect of the actual timing of the reform, but no effect of the placebo reform, then we will be more confident in our empirical strategies. Furthermore, the sign of the effect of the placebo reform might give an indication concerning the direction in which the estimates of the effects of the welfare reform could be biased. To perform the placebo reform, we change the definition of the dummy variables SECOND and REFORM in equations (2) and (4), so that only observations from the pre-reform period are used. We see that the estimates are insignificant and that they are generally rather small. If anything, the sizeable positive point estimate for the disposable income of lasting lone mothers indicates that the actual impact of the reform reported in Table 8 understates the reduction in disposable income. As pointed out by Athey and Imbens (2006), even if the identifying assumption in the DD analysis is satisfied in levels, this does not imply that it is satisfied in logs, and vice versa. Therefore, in Column 6 of Table 9, we examine the sensitivity of our results to the specification of the dependent variable in logarithmic form for our two continuous outcomes, earnings and income.11 It is reassuring to find that the results are qualitatively the same and quantitatively quite similar, when measuring earnings and income in logs rather than levels. Another concern is that the transition to lone motherhood (i.e., the inclusion in our treatment group) could be endogenous to the policy reform. The workfare reform might influence the incentives to become a lone mother in two conflicting ways. On the one hand, the work requirements and the time and age limits might make it more costly to be a lone mother, thereby providing weaker incentives for divorce. On the other hand, the increase of in-work benefit levels might strengthen the incentives for divorce. Our DD approach controls for this in so far as the fixed effects or the observable time-varying covariates capture the source of endogeneity. To investigate this further, we have examined the time trend in the probability of becoming a lone mother for married mothers with the youngest child aged 10–14 years and for married mothers with the youngest child aged 11

When running the regression with the dependent variable logged, we exclude a substantial number of observations with zero earnings (and a handful of observations with zero income), because log of zero is not defined.



4–9 years. Only the latter group was affected by the reform. We find that there is a very good coherence between the time trends of the groups before the reform. More importantly, it is encouraging to find that there is no change in the relative probability of lone motherhood at the time of or after the reform. This conforms well to the vast evidence from program evaluations carried out in the US that have shown that there are insignificant effects of welfare reforms on family composition (Moffitt, 2007).

VI. Conclusion The generous Nordic model of welfare is commonly viewed as an exceptional success, in terms of equality, economic growth, and female labormarket participation.12 However, recently, it became evident that subgroups of the population with weak labor-market attachment and high welfare dependency, such as lone mothers, were overrepresented among the poor. This prompted a major workfare reform of the Norwegian welfare system for lone mothers: work requirements were established, time limits imposed, and in-work benefit levels increased. In this paper, we have used a DD approach to produce the first evidence of the effects of this workfare reform. A main finding is that the policy changes were successful in increasing the labor-market participation and earnings of lone mothers. However, the desired effects of the workfare reform were associated with the side effects of income loss and increased poverty among a subgroup of lone mothers who were unable to offset the loss of out-of-work benefits with gains in earnings. Because most of what we know about the responses to such policy changes comes from Canada, the UK, and the US, evidence from a generous welfare state should be of particular interest. With regards to labormarket participation and earnings, our results correspond well with the Anglo-Saxon experience. However, the finding of income loss and increased poverty among a subgroup of lone mothers with overwhelming employment barriers stands in contrast to the evidence from a similar workfare reform implemented in the US in 1996. In a survey of the research on this reform, Moffitt (2007, p. 31) concluded “that the 1996 welfare reform was a success in overall terms and, on average, is almost universally accepted by policy analysts and researchers”. It is still an open question whether the presence of detrimental effects of the workfare reforms in Norway and their absence in the US are related to cross-country differences in preferences, 12 For example, the EU summit at Hampton Court in October 2005 on globalization and social models had an agenda that centered on the Nordic model. An article appearing in the International Herald Tribune in September 2005 says that “European leaders want to know how Sweden and its Nordic neighbors, so heavily laden with cradle-to-grave welfare systems, float high above the struggling economies of much of the rest of the Continent”.


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to institutional factors, such as labor-market regulations and unions, or to the design of the tax system.13 However, the disparity in the responses of lone mothers to workfare reforms across two highly differentiated worlds of welfare capitalism, as found by Esping-Andersen (1990), emphasizes the fact that policy-makers in other developed countries should be cautious when drawing lessons from the Anglo-Saxon experience.

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Whereas Blanchard (2004) emphasizes the preference argument for why labor supply is lower in Europe than in the US, Olovsson (2009) argues that differences in taxes can account for the discrepancy, and Alesina et al. (2005) stress the role of labor regulations and unions.


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