A Signal to End Child Marriage: Buchmann, Field, Glennerster, Nazneen & Wang (2023)

What this is. Core results, the signaling model of marriage timing, and the main empirical specifications from a 44-page paper: enough to know what it found and how, without reading the full source. To replicate or extend, read the original at the DOI.

TL;DR

The paper runs a clustered RCT in rural Bangladesh (2007-2017) testing two policy approaches to child marriage: (i) a conditional financial incentive (cooking oil, approximately US$16/year) paid to families while their adolescent daughters remain unmarried, and (ii) a standard adolescent empowerment program (Kishore Kontha). Girls age 15 eligible for two years are 19 percent less likely to marry underage. Among girls in school at program start, the age-15 group is also more likely to remain enrolled. The study detects no statistically significant incentive penalty in marriage-market outcomes. In the main parents’ sample, the empowerment program has no statistically significant marriage effect and raises dowry; a smaller young-women sample offers suggestive evidence of earlier marriage. The paper develops a signaling model in which early marriage signals socially conservative bride type (valued by grooms). Its predicted spillover appears among less socially conservative women near incentive communities; the estimate for more conservative women is not statistically significant.

Core results

Magnitudes as reported; *** = 1%, ** = 5%, * = 10%. ITT estimates from OLS with Huber-White SEs clustered at the community level and union fixed effects (Tables 2-4). Table 5 uses a proximity-based spillover design within non-incentive communities.

#	Result	Locator	Magnitude
R1	Incentive reduces underage marriage (< 18)	Table 2, cols. 1-2, p. 2661	-7.4 ppts*** (age 15, 19% reduction); -4.9 ppts*** (age 15-17, 17%); control means 38.5% and 29.3%
R2	Incentive increases average age at marriage	Table 2, cols 6-7, p. 2661	+0.32 years*** (3.9 months) for age 15; +0.21 years*** (2.5 months) for age 15-17
R3	Incentive increases school enrollment and attainment	Table 3, cols. 2, 4, p. 2664; text p. 2663	+8.6 ppts*** in school at midline for age 15; +5.0 ppts** in school at endline for age 15; control means 48.2% (midline) and 27.8% (endline) for age-15 cohort
R4	No marriage-market penalty detected for incentive recipients	Table 4, cols. 1, 3, p. 2666	Dowry: US$17.6 (SE 24.3, not sig.); husband education: -0.166 (SE 0.119, not sig.); control mean US$888
R5	Empowerment raises dowry with no marriage improvement	Table 4, col. 1, p. 2666; Table 2, col. 1, p. 2661	Dowry +US$57.3*** (6% above control); marriage rate coefficient -0.007 (SE 0.008, not sig.)
R6	Spillovers on untreated nearby women; null for ineligible ages	Table 5, cols 1-2, p. 2678	Age 15-17 within 500m: -2.9 ppts* married <18; placebo age 7-14: -0.9 ppts (not sig.)
R7	Spillovers are specific to less socially conservative (nonpreferred-type) women	Table 5, cols 3-4, p. 2678	Low SC: -14.7 ppts** married <18; high SC: +9.3 ppts (not sig.); control means 36.0% and 47.9%
R8	Incentive reduces marriage before age 16 among girls eligible for two years	Table 2, col. 3, p. 2661	-2.0 ppts* (18% relative to control mean 11.3%)
R9	Incentive reduces teenage childbearing, supporting the marriage-timing result with a less bias-prone birth-history outcome	Table 2, cols. 8-9, p. 2661; text p. 2663	-1.6 ppts* for ages 15-17; -3.9 ppts** for age 15
R10	Incentive increases schooling attainment, measured by last class passed	Table 3, cols. 5-6, p. 2664; text p. 2664	+0.17 years* for ages 15-17; +0.25 years (not significant) for age 15
R11	Empowerment has no statistically significant effect on marriage timing in the main parents’ sample	Table 2, cols. 1-7, p. 2661	Married <18: -0.007 (SE 0.008), ages 15-17; -0.005 (SE 0.015), age 15; no significant effect across reported marriage outcomes
R12	Incentive recipients show no marriage-market penalty in denmeher or additional husband characteristics	Table 4, cols. 2, 4-5, p. 2666	Denmeher +US$95.741 (SE 124.305); husband salaried -0.019 (SE 0.013); husband outside union +0.017 (SE 0.017); all not significant
R13	Parent-reported marriage age closely matches young women’s reports, with no treatment-arm reporting bias	Text p. 2659	In the 1,222-woman subsample, parents reported marriage age 1.5 months higher on average; difference balanced across treatment arms
R14	Empowerment raises schooling attainment despite no main-sample marriage effect	Table 3, col. 5, p. 2664; text p. 2665	Last class passed +0.174 (SE 0.091), p<0.10, for ages 15-17
R15	No complementarity between incentive and empowerment for marriage or schooling outcomes	Tables 2-3, pp. 2661, 2664; text p. 2665	Interaction coefficients are statistically insignificant throughout; last class passed interaction -0.166 (SE 0.243)
R16	No proximity spillover appears for girls observably ineligible for the incentive	Table 5, col. 2, p. 2678	-0.9 ppts (SE 1.5), not significant, for girls age 7-14 within 500m of incentive communities

Overall (paper’s conclusion). The pattern is inconsistent with child marriage reflecting strong cultural preferences for young brides or unraveling in a thin marriage market. A signaling model in which early marriage signals socially conservative (preferred) bride type explains why: a small financial incentive orthogonal to type can shift the equilibrium away from universal early marriage: treated women and untreated nonpreferred types delay, while untreated preferred types marry early in the semiseparating equilibrium. The empowerment program’s failure, and its associated dowry increase, is also predicted by the model: making more women appear nonpreferred strengthens the incentive for conservative women to signal type by marrying early.

Theory / model

The model is a two-period marriage-market signaling game (Section IV, pp. 2666-2675). Women have measure $|W| = 1$ and men measure $|M| > 1$. Women are heterogeneous in an unobservable type $\Theta \in \{\Theta_H, \Theta_L\}$, where $\Theta_H$ (preferred, socially conservative) has population fraction $f \in (0,1)$ and is privately known. Men are homogeneous and desire $\Theta_H$ women.

If a woman marries in period $t_1$ she has education $E_L$; if she delays to $t_2$ her education rises to $E_H > E_L$. The total transferable utility of a union between a man and woman of type $\Theta_j$ and education $E_j$ is $\mu(\Theta_j, E_j)$, increasing in both arguments. Two conditions govern equilibrium (p. 2669):

(i) Preferred type is first-order in desirability:

$$\mu(\Theta_H, E_H) \;>\; \mu(\Theta_H, E_L) \;>\; \mu(\Theta_L, E_H) \;>\; \mu(\Theta_L, E_L)$$

(ii) Single-crossing, type and education are substitutes:

$$\mu(\Theta_H, E_H) - \mu(\Theta_H, E_L) \;<\; \mu(\Theta_H, E_L) - \mu(\Theta_L, E_H) \;<\; \mu(\Theta_L, E_H) - \mu(\Theta_L, E_L)$$

Condition (ii) implies preferred types gain less from delayed marriage (higher education) than nonpreferred types. Dowry $D$ can be conditioned only on observable characteristics (marriage timing, education); men compete for available women and receive their outside options in equilibrium.

Result 1 (p. 2671, Section IV.B): Under a liquidity constraint $Y < \omega_M + \mu(\Theta_H, E_H) - 2\mu(\Theta_H, E_L)$, no separating equilibrium exists: nonpreferred types cannot afford the dowry $D|t_2$ required to separate as nonpreferred types.

Result 2 (p. 2672, Section IV.B): If $\mu(\Theta_H, E_H) - \mu(\Theta_H, E_L) < \mu(\Theta_H, E_L) - \bar{\mu}(E_H) < \mu(\Theta_L, E_H) - \mu(\Theta_L, E_L),$ the unique equilibrium is pooling on $t_1$ (all women marry early), with dowries:

$$D|t_1 = \omega_M - \bigl[f\,\mu(\Theta_H, E_L) + (1-f)\,\mu(\Theta_L, E_L)\bigr], \qquad D|t_2 = \omega_M - \mu(\Theta_L, E_H)$$

This pooling equilibrium survives the intuitive criterion. Pooling on $t_2$ fails that criterion because preferred types would profitably deviate to $t_1$ (where grooms know with certainty that early entrants are preferred type). Pooling on $t_1$ survives because neither type can credibly signal itself by deviating to $t_2$. This is the child-marriage equilibrium of Wahhaj (2018) but generated here by differential returns to education rather than preference for young brides.

Corollary 2 (p. 2675): A decrease in $f$ (more nonpreferred types) makes pooling more likely. This is the mechanism by which the empowerment program can worsen early marriage: by reducing the perceived fraction of preferred types, it strengthens the signaling incentive for conservative women.

Result 3 (p. 2673): A random incentive independent of type cannot sustain a separating equilibrium; it relaxes liquidity constraints for preferred and nonpreferred types alike. Result 4 (pp. 2673-2674): If a conditional incentive $C$ satisfying

$$C \;>\; \mu(\Theta_H, E_L) - \bar{\mu}(E_H) - \bigl[\mu(\Theta_H, E_H) - \mu(\Theta_H, E_L)\bigr]$$

is offered randomly to fraction $\tau$ of women (treatment orthogonal to type), a semiseparating PBE exists and Pareto-dominates the pooling equilibrium. Untreated preferred types marry at $t_1$; all other women delay to $t_2$. Equilibrium dowries under the incentive:

$$D|t_1 = \omega_M - \mu(\Theta_H, E_L), \qquad D|t_2 = \omega_M - \bar{\mu}(E_H)$$

The paper defines $\bar{\mu}(E_H)$ as expected match quality among $t_2$ entrants. Written out from that definition, it is $\bar{\mu}(E_H) \equiv \frac{\tau f\,\mu(\Theta_H, E_H) + \tau(1-f)\,\mu(\Theta_L, E_H) + (1-\tau)(1-f)\,\mu(\Theta_L, E_H)}{\tau f + \tau(1-f) + (1-\tau)(1-f)}$ when the fraction of preferred types among $t_2$ entrants is $f' < f$. Crucially, untreated nonpreferred types also delay (the spillover): because treatment status is unobservable, delaying marriage is no longer a certain signal of being the nonpreferred type (Prediction 4, p. 2676).

Method

Experimental design. A clustered randomized trial in 460 communities across five subdistricts in rural Bangladesh, randomized in ratio 1:2:1:2 to (i) incentive only, (ii) empowerment only, (iii) both, (iv) control. Randomization was stratified by union (administrative grouping of approximately 10 communities) and community size. All girls aged 15-17 and unmarried at program start (January 2008) were eligible for the conditional incentive; 92 percent received it at least once. The empowerment program (Kishore Kontha) reached 93 percent of eligible girls.

The incentive program was implemented January 2008 to August 2010. Main outcome data come from an endline survey conducted 2016-2017, approximately 4.5 years after program completion.

Estimation. ITT effects are estimated via OLS linear probability models (panel-regression) with Huber-White robust standard errors clustered at the community level (the unit of randomization). The theoretical mechanism is formalized in the signaling-game-pbe model (Results 1-4 above). Field and Ambrus (2008) establish the education channel in Bangladesh; Duflo, Dupas, and Kremer (2015) and Baird, McIntosh, and Ozler (2011) provide comparable evidence from transfer programs in other contexts.

Empirical specifications

Main ITT regression (Equation 1, p. 2659):

$$Y_{icu} = \alpha + \beta_1 I_c + \beta_2 E_c + \beta_3 (I_c \times E_c) + \beta_4' \mathbf{X}_{ic} + \varepsilon_{icu} \tag{1}$$

where $Y_{icu}$ is outcome for person $i$ in community $c$ and union $u$; $I_c$ = assignment to the incentive program; $E_c$ = assignment to the empowerment program; $\mathbf{X}_{ic}$ includes village population tercile and union fixed effects, age indicators, household size, older unmarried sister in household, school enrollment, mother’s education, and public transport access (a proxy for remoteness). The interaction $I_c \times E_c$ tests complementarity. Tables 2, 3, and 4 report $\hat{\beta}_1$, $\hat{\beta}_2$, $\hat{\beta}_3$. The main analysis sample covers 15,576 women in the parents’ survey.

Spillover specification (written out from Section V; Table 5, p. 2678): Estimated on nonincentive communities only, comparing women within 500 meters of an incentive community center to women farther away:

$$Y_{icu} = \alpha + \gamma \cdot \text{Close}_c + \delta' \mathbf{W}_{ic} + \varepsilon_{icu}$$

where $\text{Close}_c = 1$ if the community center is within 500 meters of the closest incentive community center. $\mathbf{W}_{ic}$ includes union fixed effects, baseline characteristics, and distances to the closest community center and safe space (to rule out urbanicity confounds). Placebo test: same regression for girls aged 7-14 who are observably ineligible. Type heterogeneity: equation (4) is estimated separately for girls with above-median and below-median social conservatism index (columns 3-4 of Table 5), testing Prediction 4 of the signaling model. Huber-White standard errors are clustered at the community level. The parents’ survey specification has 10,544 observations for ages 15-17 and 40,136 for the ineligible ages 7-14 placebo; the young women’s survey specification has 682 low-conservatism and 417 high-conservatism observations.

Datasets used

Dataset	Role in paper	Wiki page
Authors’ RCT survey data (parents’ and young women’s surveys)	Main analysis: marriage outcomes, schooling, dowry, husband characteristics; 460 communities, 24,095 girls at baseline, 15,576 in endline parents’ sample	No page yet (hand-collected; replication data at https://doi.org/10.3886/E192114V1)
Bangladesh Demographic and Health Surveys (DHS, waves 2004-2017)	Descriptive context: trends in marriage age, education, and child mortality 2004-2017 (Figure 1, p. 2646)	No page yet

Sample scope: rural Bangladesh. The initial sample covered six subdistricts (Daulatkhan, Babuganj, Muladi, Patuakhali Sadar, Bauphal, Bhola Sadar); Muladi was dropped in 2011, excluding 2,061 girls in 84 communities, leaving five subdistricts in the analyzed sample. Program period January 2007 to September 2017. Main endline wave: 2016-2017.

When to read the full paper

Read the original if you are: (i) extending the signaling model to other cultural contexts with bride-type heterogeneity or dowry; (ii) designing a conditional transfer program for child marriage reduction; (iii) examining the cost-effectiveness calculation of US$1,010 NPV per US$1,000 invested (Buchmann et al. 2021 companion paper); or (iv) testing Predictions 1-4 of the signaling model in a new setting. Tables 2, 3, and 5 contain the core empirical results; Section IV formalizes the model; Appendix A carries the proofs.

Attribution and rights

Source: peer-reviewed, American Economic Review 113(10), October 2023. This distillation was extracted by an LLM on 2026-10-04 and re-verified against the source PDF on 2026-10-04; it has not been independently reproduced. The article is freely accessible after the AEA 12-month embargo (expired October 2024) and via NIH PMC; no CC license applies, so redistribution is extract-only.

Buchmann, Nina, Erica Field, Rachel Glennerster, Shahana Nazneen, and Xiao Yu Wang. “A Signal to End Child Marriage: Theory and Experimental Evidence from Bangladesh.” American Economic Review 113, no. 10 (October 2023): 2645-2688. DOI: 10.1257/aer.20220720.