rafmacalaba/data-use-mentions-extended huggingface dataset:
−0.003|−0.072|
||(0.170)|(0.143)|(0.131)|(0.147)|
|Number of friends|0.068***|0.006|0.044***|0.052***|
||(0.017)|(0.015)|(0.016)|(0.016)|
|Control for baseline value of dep. var.|Yes|Yes|Yes|Yes|
|_P_-value: treatment * (% of friends)|0.129|0.279|0.270|0.061|
|=spillover * (% of friends)|||||
|Observations|3,856|3,882|3,840|3,891|

_Source_ : Authors’ analysis from student surveys. _Note_ : Dependent variable is an index of related questions. All components of the indices are standardized to mean 0 and standard deviation 1, based on the sample frame at baseline. Standard errors clustered at the school level in parentheses. *** _p <_ 0.01, ** _p <_ 0.05, * _p <_ 0.1. _Condom use attitudes subindex_ : Respondent disagrees with statements (a) “It’s not right to carry a condom because people may think that I planned to have sex”; (b) “If a woman wants to have sex without a condom, the man must not refuse”; (c) “Only women are responsible for unwanted pregnancies”; respondent is (d) confident if requesting that a condom be used; (e) willing to delay sex if condoms are unavailable; respondent thinks (f) he/she will use a condom in his/her next sexual relationship. _Sexually conservative attitude subindex_ : Respondent thinks that (a) it is not right when people of
. We thus have direct observation data for 339 of the 1,363 health workers in our experiment. The observation was conducted using a structured, quantitative checklist and the data were collected by enumerators trained in the direct observation of antenatal care provision. Enumerators recorded whether the health worker performed actions listed in the standard WHO protocol, and whether she performed five standard screening tasks to assess the woman’s risk for developing serious pregnancy complications. The latter are closely related to our experimental task of identifying emerging complications and guiding interventions. Moreover, like in our task, the screening can be performed in any setting whereas the physical actions may require supplies or equipment, such as a stethoscope or lab kits for 14
Policy Research Working Paper 10605

# **Abstract**

This paper analyzes gender gaps in the Colombian labor market, with a particular focus on the impact of the COVID-19 shock and the economic recovery. Using household survey and administrative data, the analysis finds significant and persistent gender gaps in favor of men in terms of participation, unemployment, and income. These gaps are heterogeneous at the regional level in Colombia, and are exacerbated among women with children, particularly young children, and those with low levels of education. The COVID-19 pandemic widened these gaps, including from higher female concentration in the most affected sectors and potentially associated with the disproportionate burden of care on women. Moreover, using decomposition techniques, the paper finds that the decline observed in the hourly gender wage gap results from two offsetting effects: the explained gap, which favors women due to their endowments, primarily education level, and the unexplained gap, which favors men and may be associated with discriminatory biases. Moreover, the gender wage gap widens significantly when considering monthly income, showing variations in hours worked between men and women. Given the poorer labor market outcomes among women with children, policies toward the reduction and redistribution of care work within households could contribute to increase women’s opportunities in the labor market.

This paper is a product of the Poverty and Equity Global Practice. It is part of a larger effort by the World Bank to provide open access to its research and make a contribution to development policy discussions around the world. Policy Research Working Papers are also posted on the Web at http://www.worldbank.org/prwp. The authors may be contacted at mdavalos@ worldbank.org. A verified reproducibility package for this paper is available at http://reproducibility.worldbank.org, click **here** for direct access.

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conditions typical of poor countries, make it harder to convert their notional SP demands into public spending outcomes—in short, that poorer countries face higher costs in effectively implementing this type of public spending. Prices of competing human development services (health and education) may also play a role, to the extent that these services are relatively expensive in richer countries, encouraging substitution toward SP transfers within public budgets.

Testing the luxury good hypothesis can be readily recognized as a classic issue in causal inference. The idea of omitted variable bias (OVB) provides a natural framework for structuring our effort. We formulate a series of alternative hypotheses to the luxury good hypothesis. These suggest specific country characteristics that might be expected to influence the SPEC while also being correlated with national income. (We strive to avoid a “kitchen sink” approach to controls, but instead to carefully formulate hypotheses based on the literature or our reasoning.) We interpret these characteristics as reflecting differences in the implicit prices faced in providing SP; for example, that the implicit price of SP is higher when government effectiveness or technology access in general is lower. We do not claim that our list of potential confounders is exhaustive, but (as we will see) it is enough to be confident about the answer to the main question posed by this paper.

The specific hypotheses we consider relate to relative prices for human development services, political economy (governmental accountability, the capacity for public service provision more broadly, and redistributive policy making), the distributional impacts of economic growth, access to appropriate technologies, population ageing, and the selection processes for observed public spending data. We argue that some of the hypotheses we consider can already be dismissed from what we know, based on the literature. Among the rest, each hypothesis points to one or more potential confounders in estimating the SPEC, which we use in testing for OVB. We also point to some likely theoretical ambiguities in how the confounders can influence SP spending; for example, having a more capable
more female workers but exhibit a larger gender wage gap than domestic firms, based on a census of firms. The lower wages offered to females are reflective of their assignment to low-tech and low-training jobs and their lower productivity, so they do not indicate gender discrimination. Coniglio et al. (2017) show that foreign affiliates in Vietnam create more employment opportunities for female workers than domestic firms, based on a cross-sectional representative sample of firms. But most of those jobs are in low-skilled occupations whereas job opportunities for high-skill female workers created by foreign firms are limited, likely due to Vietnam’s comparative advantage in labor-intensive low-tech manufacturing. The wage premia of foreign affiliates decrease if firms have a higher share of female workers. Rocha and Winkler (2019) <mark>show a female labor share premium for firms engaged in global value chains, and in particular for foreign-owned firms, based on a cross-section of more than 29,000 manufacturing firms in 64 developing countries from the World Bank's Enterprise Surveys. The female labor share premium is much higher for production workers compared to non-production workers, implying that women specialize in low-skill production. The evidence also shows that while average wage rates are lower for firms with higher female labor shares, this negative correlation is smaller for foreignowned firms. K</mark> odama et al. (2018) show that foreign affiliates are more gender-equal in Japan in the sense that they exhibit higher proportions of females among workers, managers, directors, and board members than domestic firms of comparable size operating in the same industry in the same year, based on repeated cross-sections of firms over the period 2004-2014. Foreign affiliates create more female-friendly conditions in the workplace relative to domestic firms in that they offer 5
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.

## IV.1 Definition

The analysis in this paper relies on a comprehensive UNESCAP-World Bank dataset of bilateral trade costs. Following Novy (2013), Arvis et al. (2013) derive measures of annual trade costs for the period 1995-2018. For any given country pair i and j, trade costs are obtained as geometric averages of trade flows between countries i and j. They are computed according to the formula: ( _Xii Xjj_ )/( _Xij Xji_ )<sup>1/2 (σ-1)</sup> , where _Xij_ represents trade between countries _i_ and _j_ (goods produced in _i_ and sold in _j_ ) and σ refers to the elasticity of substitution. This measure assumes that international trade costs relative to domestic trade costs are reflected in international trade flows relative to domestic trade flows: when international trade costs are higher than the costs of domestic trade, countries will trade more domestically than internationally, i.e. the ratio ( _Xii Xjj_ )/( _Xij Xji_ ) will be higher. In the application of this methodology, domestic trade is proxied by the difference between gross output and total exports. Trade costs thus estimated are expressed as a proportion of the value of traded goods (comparable with an ad valorem tariff rate) and can be computed for the economy as a whole, or
all country dyads and identified whether that dyad is governed by a treaty. For country dyads with a treaty, we pay close attention to when the first treaty was negotiated. The time-period after the first treaty is signed is

> 3 Due to the most recent update of the TFDD treaty database (Giordano et al. 2014) and of the water event data set (De Stefano et al. 2010), this paper included 220 treaties and 755 water events more than the one by Brochmann (2012).

5
, approximately 65 percent of households had consumed some type of tobacco product during the month previous to the interview. Comparing across time, the share of households consuming tobacco has been decreasing during the last decade. Using previous rounds of the HIES and comparable measures, one may see that the share of households consuming tobacco was 82 percent in 2005, 71 percent in 2010, and 68 percent in 2016. Yet, this has been accompanied by an increase in the share of households consuming cigarettes from about 23 percent in 2005 to 30 percent in 2016.

> 2 Bidis are popular hand rolled tobacco products smoked in South Asia.

> 3 Note that the true smoking prevalence among women may be greater than estimated because evidence suggests that women (and those who proxy-respond for them in surveys) tend, for social reasons, to underreport risky behaviors.

> 4 Data on tobacco consumption vary in terms of coverage, period, and quality. These numbers come from the 2009 Global Adult Tobacco Survey Bangladesh.

2
Agriculture Questionnaire, which solicited information on land areas, labor and non-labor input use, crop cultivation and production at the plot<sup>5</sup> -level, separately for the reference rainy and dry seasons.

While the data do not include direct measures of either intra-household cooperation or wives’ empowerment, the IHPS included over 40 questions asked at varying levels of observation across all income domains that could be used to identify up to 2 household members that decide on the use of each unique source of income. Using this information, we calculate total gross income<sup>6</sup> that is controlled solely by females; solely by males; and jointly by females and males; and use the share of joint controlled income in total gross income as a proxy for intra-household cooperation, and the share of female controlled income in total disjoint gross income as a proxy for wives’ empowerment. Using the share of female controlled income in total disjoint gross income avoids difficulties in interpreting the coefficient that would arise if we used female share in total income. Conceptually, it also allows for a cleaner distinction between cooperation leading to joint control of income versus wives’ bargaining power leading to greater control over disjoint income. Appendix II provides the details on the computation of intra-household income shares.

Additionally, we would like to allow for non-linear impacts of income shares on our outcomes of interest. However, including a squared term may magnify attenuation bias if there are measurement errors in our income shares variables (Griliches, 1986). Income shares are, of course, based on measures of income, which suffer from measurement error. As noted by Bound et al. (1994), fixed effects can compound measurement error, particularly when the variable of interest is highly correlated over time, as it is here. Therefore, instead of using intra-household income shares as calculated above, we divided both income shares of interest into five categories. Three categories map to income share terciles calculated in each round among households where income shares
8 _Figure 5: Progressivity of health financing sources in Japan and the Asian Tigers_ * <!-- Start of picture text --> 0.40<br>0.30<br>0.20<br>Taxes<br>0.10<br>Soc. Ins.<br>Priv. Ins.<br>OOP<br>0.00<br>Total<br>-0.10<br>-0.20<br>-0.30<br>Hong Kong, Japan Korea, Rep. Taiwan, China<br>China<br>Kakwani index<br><!-- End of picture text --> The differences in the progressivity of total payments reflect the different shares raised from the various sources (Figure 3) and the progressivity of each source (Figure 5).<sup>†</sup> Taxes are most progressive in Hong Kong (China), which is also the health system that relies most on them in the financing of health care. They are regressive in Japan (largely because of the regressiveness of indirect taxes there<sup>4</sup> ), but progressive—albeit less so than in Kong Hong (China)—in Korea and Taiwan (China).

Social insurance contributions are regressive in all systems where they are used. This is not unusual and typically reflects the effects of contributions ceilings.<sup>13,14</sup> Any exemptions or reduced contribution rates among the poor would reduce the degree of regressiveness, and could in principle make contributions progressive. In Korea<sup>15</sup> and Taiwan (China)<sup>11</sup> there are exemptions or reduced contribution rates for the poor or otherwise disadvantaged (in the case of Taiwan (China) the poorest 1% of households are fully exempt, and a further 4.5% have their contributions paid in full or subsidized
Policy Research Working Paper 10337

# **Abstract**

This paper studies the impacts of the COVID-19 pandemic on Indonesia’s labor market by exploiting the exogeneous timing of the pandemic in a seasonal difference-in-differences framework. The analysis uses multiple rounds of Indonesia’s National Labor Force Survey from 2016 to 2020 to establish a pre-pandemic employment trend and then attribute any excess difference in employment outcomes from this trend as the estimated effect of the pandemic on individual employment outcomes. The results suggest that the pandemic has had mixed effects on the Indonesian labor market. While the pandemic has narrowed the gender gaps in employment participation through the “added worker effect” among women, it has also lowered the overall employment quality among both gender groups. The findings show that the increase in female employment is primarily driven by women in rural areas without high school education, entering informal work, agricultural employment, or unpaid family work. For men, the pandemic has had adverse impacts on their employment across the board in all sub-populations. Consistent with findings from other studies, steeper employment declines are observed in urban areas, particularly among males. Among those employed, both women and men work fewer hours and earn lower wages.

This paper is a product of the Social Sustainability and Inclusion Global Practice, the Poverty and Equity Global Practice, and the Gender Global Theme. It is part of a larger effort by the World Bank to provide open access to its research and make a contribution to development policy discussions around the world. Policy Research Working Papers are also posted on the Web at http://www.worldbank.org/prwp. The authors may be contacted at dhalim@worldbank.org and rpurnamasari@worldbank.org.

_The Policy Research Working Paper Series disseminates the findings of work in progress to encourage the exchange of ideas about development issues. An objective of the series is to get the findings out quickly, even if the presentations are less than fully polished. The papers carry the names of the authors and should be cited accordingly.
not accessible for safety reasons in the 2007 survey (in NTB province), we have the total of 98 villages (in 39 kabupatens and 48 kecamatans).

# **4.2 Descriptive evidence**

# **Agricultural production**

Using the survey data, summary statistics provide basic information on agricultural households producing rice, maize, cassava, estate plantation crop, and horticulture crop in 2007 (N=762). The agricultural sector in Indonesia is characterized by smaller size of land (mean is 1.4 Ha in 2007) and multiple-crop farming. Many farmers cultivated three or four crops in Central and East Java.

There was a dramatic increase in agricultural production from 2007 to 2010, which seems to be accompanied by large investments in production capital (i.e. owning farm building; having machineries such as tractor, thresher, sprinkler, irrigation pump, sprayer, and dryer; buying agricultural tools; see Table 2 for a detailed breakdown of investments).

Table 3 shows the regional variation of the average agricultural production revenues in 2007 for rice, maize, cassava, estate plantation crops, and horticulture crop from IMDG 2007. Based on food crop statistics by the Badan Pusat Statisti (BPS), Java is the major producing area of rice, maize, and cassava as well as farming multiple crops including various vegetables. In remote areas in South Kalimantan and NTB, farmers primarily produce rice, whereas farmers in Lampung, and North and South Sulawesi are specialized in estate plantation crops.

16
education of the mother, and state of residence. Note that the effect of each of these covariates is allowed to change across cohorts.<sup>17</sup> Table 4 shows estimated coefficients on a selective group of variables of equation (1). Panel A is a copy of Table 3, and presents the expected impact of the program as a function of the cohort of the child, the year height was measured and the type of locality of residence. Panel B shows the estimated results for rural areas, and Panel C presents the results for urban areas. Each cell presents the estimated coefficient corresponding to the i1 , i2 , and i3 terms in equation (1). In Panel B, for example, the cell corresponding to the older cohort in community-group 1 in 2002 (-0.52) αα αα αα shows the estimated value of 1 , when equation (1) is restricted to the rural sample. oo02 αα In the absence of selective program placement, the coefficients in equation (1) would already represent impact effects. But we know that is not the case. Looking at the rural areas, results in Table 4 are consistent with the fact that Oportunidades was first implemented in the poorest communities of Mexico, and reinforces descriptive analysis done with census data on a broader set of socioeconomic variables. As expected, rural communities incorporated to Oportunidades before 2005 are statistically different from communities incorporated later (or never incorporated) in terms of average children’s height: 2 and 3 are negative and significant at the 5 and 10 oo02 oo02 percent level, when they should be zero in the absence of selection in program placement. This αα αα analysis cannot provide direct evidence of placement selection for the first group of communities but the literature documents they are among the poorest communities in Mexico. As a result, we see that Oportunidades was introduced first to communities where children’s nutritional status was

> 17 We tried specifications that allow for the covariates to be cohort-time specific and they essentially
# **1. INTRODUCTION AND REVIEW OF THE LITERATURE**

**Transport planning should not be gender neutral** (Dominguez Gonzalez _et al_ ., 2020). Even if transport related improvements seem to benefit women and men equally, women and men experience transport and infrastructure differently in terms of affordability, acceptability, and availability (Alam _et al_ ., 2022). Lack of safe transport can reduce women’s access to economic opportunities (International Labor Organization, 2017). For example, global evidence suggests that the lack of proper infrastructure heavily influences women’s decision to cycle (Garrard _et al_ ., 2008), contributing to gender inequity in access to jobs and other opportunities in parts of the urban areas where such infrastructure is unavailable. In addition, there is a recognition that there are other, non-transport-related barriers that can constrain women’s mobility, such as social norms regarding gender, unequal division of household responsibilities, and other more personal elements influencing women’s agency, such as self-efficacy and aspirations (Dominguez Gonzalez _et al_ ., 2020). Unfortunately, regardless of these differences, transport planning has mainly focused on unidirectional travel from home to work, paying less attention to, for instance, shorter trips associated with “mobility of care”, which are typically carried out by women (Sánchez de Madariaga, 2013).

**Gender influences the decision to travel in the first place and, specifically, mode choices and trip patterns.** It was not until the last decade that data on mobility in Latin America started to be genderdisaggregated in origin-destination surveys, for instance, in Chile, Colombia, Uruguay, and Mexico City. Evidence on women’s mobility from Latin America seems to resemble that from some developed countries in terms of modal choice, distance, and travel purpose. Data shows that women travel shorter distances, have more complex multimodal trips, and walk and rely on public transport more than men. Women have less access to cars and motorcycles, while