Property Rights and Gender Bias:
Evidence from Land Reform in West Bengal *
12 March 2016
Sonia Bhalotra
University of Essex
Dilip Mookherjee
Boston University
Abhishek Chakravarty
University of Essex
Francisco J. Pino
University of Chile
Abstract:
Formalization of land rights can increase productivity and welfare, but may also worsen gender
inequality when men have stronger property rights than women. We document male-biased
fertility stopping, sex selection, and infant survival improvements flowing from land reforms in
the Indian state of West Bengal. The reforms increased infant survival in both Hindu and NonHindu families. However, boys’ survival rates increased far more relative to girls in Hindu
families, which exhibit greater son preference than Non-Hindu families. The effects persist
despite conditioning on post-reform increases in agricultural yields. Regional income growth
therefore does not mitigate increased gender inequality following reforms.
JEL Classification: I14, I24, J71, O15
Keywords: Land reform, property rights, infant mortality, gender bias
*We are grateful to S. Anukriti, Maitreesh Ghatak, Giacomo de Giorgi, Tarun Jain, Stephan Litschig, Pushkar Maitra,
Giovanni Mastrobuoni, Patrick Nolen, Imran Rasul, Debraj Ray, Sanchari Roy, Alessandro Tarozzi, and participants
at several conferences, seminars, and workshops for their comments. Funding from the MacArthur Foundation
Inequality Network, NSF Grant No. 0418434, and the Centre for Social Conflict and Cohesion Studies
(CONICYT/FONDAP/15130009) is greatly acknowledged. All errors remain our own.
1
1. Introduction
Secure property rights are considered a cornerstone of economic development. Developing
countries with large populations dependent on agriculture have devoted much attention to land
rights. During 1955-2000 a billion people and nearly as many hectares were affected by land
reforms (Lipton, 2009). Extensive research demonstrates the importance of land security in
increasing agricultural productivity, facilitating access to credit, and reducing poverty (E.g. see
Banerjee et al., 2002; Besley & Burgess, 2000; Besley, 1995; Besley & Ghatak, 2010; Besley et al.,
2012; Goldstein & Udry, 2008; Hornbeck, 2010). However, there is little research on the impact
of improved land rights on the gender difference in child survival in developing countries, which
is significantly biased in favour of males (Klasen & Wink, 2002; Sen, 2003). Greater land security
may increase incomes, reducing both male and female child mortality. It may however also cause
the returns to parental investments in children to diverge in favour of sons when rights to land
ownership, cultivation, and inheritance are institutionally male-biased. Increased gender
inequality in child survival is therefore a potential consequence of land reform, which may help
explain why income growth in poor countries may worsen gender inequality.
There is growing evidence that the gender difference in human capital responds to changes in
women’s earnings opportunities and inheritance rights in developing countries (Deininger et al.,
2013; Jensen, 2010; Roy, 2015), and that economic development does not reduce gender
discrimination if male-biased cultural norms persist (Duflo, 2012; Jayachandran, 2014). In this
paper we examine how sharecropper registration during the land reform programme Operation
Barga affected infant mortality by gender, the sex ratio at birth, and gender-biased fertility
stopping. Operation Barga was initiated by the state government of West Bengal in the 1970s to
provide secure and heritable rights to sharecropping tenants by registering their tenancy and
guaranteeing them a minimum share of sharecropped rents. We show that the male advantage in
infant survival and the sex ratio at birth increased after this land reform. We further show that
these results obtain even after conditioning on post-reform increases in regional agricultural
yields, thereby indicating that the results reflect gender bias in property bequest motives rather
than wealth effects. 1 The increased gender gap in survival is concentrated in the Hindu
community, which historically shows greater daughter-aversion in health investments than NonHindu communities (E.g. see Bhalotra, et al., 2010; Bhalotra & Zamora, 2009). Together these
While we condition on the effects of increased post-reform regional productivity, and thereby regional income, the
institutional mechanisms we identify include increases in household returns to the labour of male children relative to
female children, which may evolve independently of regional aggregate income growth.
1
2
findings indicate that institutional and cultural norms may cause the formalisation of property
rights to widen gender inequality while increasing incomes.
To the best of our knowledge, Almond et al. (2013) is the only other study regarding the impact
of property rights formalisation on gender inequality in the presence of son preference,
specifically examining the impact of land reforms in China on male bias in the sex ratio at birth.
Their paper also finds an increased proportion of male births in response to improved land
rights. Our findings are more diverse, as they identify household responses to land reform in
post-natal discrimination in addition to sex selective behaviour. Our results are also the first to
show that the increased parental bias against girls arises from institutional channels independent
of wealth effects as property rights over land improve.
Our evidence comes from two separate but complementary empirical exercises. We first combine
the 1992-93 and 1998-99 waves of National Family Health Survey (NFHS) data with district
sharecropper registration rates from Banerjee et al. (2002), and use a quadruple difference-indifferences strategy to ascertain the impact of district-level land reform on infant mortality risk
by child gender, and the gender of the firstborn child in the household. We find that Hindu boys
experience a significant decline in infant mortality risk of 4.0-5.9 percentage points once tenancy
registration rates in their district of birth exceed 50%. Hindu girls with firstborn older brothers
face no parental discrimination, and also experience an infant mortality decline of 5.1-8.6
percentage points following high district registration. Hindu girls with firstborn older sisters
however see no improvement in infant survival after high district registration, with estimates
even suggesting a post-reform increase in mortality of 1.1-2.3 percentage points. Therefore,
Hindu girls without firstborn older brothers appear to face parental discrimination after the
reform, and the child sex ratio amongst Hindu families becomes more male-biased at age 1. In
contrast, amongst Non-Hindu families, we find all children experienced similar declines in
mortality risk of 7.2-7.8 percentage points as a result of the reform regardless of their gender, or
that of the firstborn child. Firstborn children’s gender in India is plausibly exogenous2, as sex
selection by son-preferring households is typically manifested only at second or later births if the
first child is female (Bhalotra & Cochrane, 2010). This allows us to identify programme impacts
without confounding bias from differential mortality trends by child gender across districts that
may correlate with registration rates. The fact that the fall in relative survival chances of girls
were confined to specific combinations of religion, birth order and gender of the first child
We explicitly test whether this assumption holds in our dataset, and show in our results that land reform has no
statistically significant impact on infant survival, sex ratio at birth, and fertility-stopping at birth order 1.
2
3
render unlikely the possibility that they were driven by other health policy reforms or unobserved
shocks correlated with the land reforms Our results are also robust to controls for a districtspecific linear time trend and district-level public programmes such as the number of health
institutions and kilometres of surfaced road per capita which would similarly affect Hindu and
Non-Hindu households.
We additionally find that Hindu families increase sex selection in response to land reforms, with
the probability of births being male at birth order 2 or higher increasing by 4.8 percentage points
when the district registration rate exceeds 50%. This probability rises to 5.5 percentage points if
the firstborn child in the family is a girl, which ostensibly increases the parental desire for
following births to be male as land security improves. Hence, the sex ratio at birth in Hindu
families is more male-biased following land reforms, and this male bias is further accentuated by
increased infant mortality of girls without firstborn older brothers as described above.
Consistent with the previous literature, and with our infant mortality results, we find no evidence
of such sex selective behaviour in Non-Hindu families in response to land reforms. We do
however see that Hindu as well as Non-Hindu families practice increased son-biased fertility
stopping in response to land reform. When their first child is male, Hindu families are 10.4-10.8
percentage points likelier to cease childbearing after a second child once district registration
exceeds 25%. However, land reform does not induce fertility stopping amongst these families
when their firstborn child is female. Therefore, the similar decline in post-reform mortality for
Hindu girls with firstborn older brothers and Hindu boys is partially explained by the fact that
these girls have fewer siblings and face less competition for resources. In Non-Hindu families,
district registration above 50% reduces the probability of a third birth by 9.6 percentage points,
and this probability declines more sharply by a further 18.2 percentage points when the first
child in the household is male and registration exceeds 25%. This behaviour is again consistent
with the previous literature, which shows that in India, Muslim households (the overwhelming
majority of Non-Hindu households in our sample) exhibit son preference by altering fertility but
not via postnatal discrimination (Bhalotra & Cochrane, 2010; Bhalotra & Zamora, 2009).
We subsequently seek to corroborate the preceding results and attempt to understand causal
mechanisms better with more disaggregated data concerning household economic status, land
productivity, and land reform implementation at the village level (Bardhan & Mookherjee, 2010;
2011). This is a panel data set from surveys of 2,400 households across 89 villages in West
Bengal, with unique information on household immigrant status, land holdings, and the village4
level share of land that was registered under Operation Barga. Using household fixed-effect
regressions of birth-cum-survival, which examine survival probabilities net of fertility changes,
we show that a 1% increase in the share of cultivable village land registered in the previous three
years significantly increased the likelihood of a surviving boy3 in Hindu landless households by
2.5 percentage points in any given year. There was no corresponding increase in likelihood of a
surviving girl in Hindu landless households, and in Hindu landed households there was a
significant decline in the probability of a surviving girl by 1.1 percentage points, indicating that
the increase in male surviving offspring in the household is due to discrimination. These results
are robust to the inclusion of controls for several agricultural support programmes implemented
by the government, and also survive a more stringent specification with district-year fixed effects
that control for time-varying unobservables at the district level. There are no similar effects of
land registration on the probability of surviving boys or girls in Non-Hindu households,
consistent with our other results. The effects are concentrated amongst immigrant landless
households, who were most likely to benefit from sharecropper registration, and native landed
households who were the most likely to lose land privileges after reforms.
All our findings are robust to the inclusion of land rice productivity measures as controls, ruling
out rising income from tenant registration and other programmes such as HYV minikit
distribution as a mechanism for increased gender bias. Rather, biases against women arising from
cultural institutions are the likely mechanisms behind our findings. How such institutions shape
parental preferences in different communities therefore fundamentally determine who gains
from improved land rights. This is in contrast to Almond et al. (2013), which also finds increased
sex selection in favour of sons in China in response to land reform, but does not separate rising
post-reform income and institutional factors as causal mechanisms.
This paper contributes both to the literature on land reforms and on gender bias in developing
countries by establishing a clear connection between them. West Bengal is one of the only Indian
states to have achieved notable success in land reform, and the impacts of Operation Barga on
economic outcomes since its initiation have been closely studied (E.g. see Banerjee et al., 2002;
Bardhan & Mookherjee, 2011; Bardhan, Mookherjee, & Kumar, 2012; Bardhan et al., 2013).
However, we are not aware of any study examining human development impacts of Operation
Barga, with the exception of Deininger et al. (2011) who find that the programme increased
years of schooling of children from poor and lower caste households. Furthermore, India has
3
A surviving child is defined as a child observed in the household at the time of the survey.
5
high prevailing son preference. There is substantial research documenting parental discrimination
against girl children in postnatal health investments in India, which in combination with sexselective abortion has significantly skewed the sex ratio (E.g. see Anderson & Ray, 2010; Babu et
al., 1993; Barcellos et al., 2013; Behrman, 1988; Bhalotra, 2010; Bhalotra & Cochrane, 2010;
Borooah, 2004; Chakravarty, 2010; Jayachandran & Kuziemko, 2011; Murthi et al., 1995; Oster,
2009; Rose, 2000). However, there is a paucity of evidence concerning the role of property
inheritance norms in generating gender imbalance, despite widespread discussion of this issue
(Bardhan, 1974; Rahman & Rao, 2004; Jain, 2014; Carranza, 2014).
The rest of the paper is organised as follows. Section 2 provides a background discussion of
land reform in India, Operation Barga in West Bengal, and prevailing son preference norms.
Section 3 presents the proposed mechanisms of impact that motivate the empirical analysis. In
Section 4 we discuss the data that we use in our estimations. Section 5 outlines our empirical
methodology, and in Section 6 we present our results. Section 7 concludes.
2. Background
2.1 Historical Context
Upon national independence in 1947, the Indian central government began three main types
of land reforms to address large historical inequalities in land distribution. These were abolition
of intermediaries, new tenancy laws to protect against eviction and extraction of excessive rental
crop shares by landlords, and land ceilings to limit the amount of land held by any one
household with the aim of vesting and redistributing surplus land to small farmers.
Implementation of the reforms was left to individual state governments, and barring
intermediary abolition in nearly all states landlords were able to subvert the remaining reform
measures by way of pre-emptive tenant evictions and parcelling land to relatives to avoid state
confiscation of above-ceiling holdings (Appu, 1996). Variation in state-level reform
implementation and legislation over time has been used in previous studies to empirically
estimate land reform impacts on poverty, equity, and human capital (Besley and Burgess, 2000;
Ghatak & Roy, 2007; Ghosh, 2008).
Reforms in the state of West Bengal were spurred by the results of the 1977 state assembly
election. The Left Front coalition won an absolute majority, which it retained until 2011. This
new government promptly created a three-tier system of local governments called panchayats,
6
which for the first time would be democratically elected. These tiers in descending order of size
of jurisdiction were district, block, and finally the gram panchayat that operated at the village level
with a jurisdiction of 10-15 hamlets (mouzas). Many national development programmes as well as
aspects of new state welfare initiatives such as Operation Barga were then decentralised to these
gram panchayats, who were largely responsible for selecting local eligible beneficiaries and lobbying
the upper tiers of the new system for funds (Bardhan & Mookherjee, 2011).
2.2 Operation Barga and the Green Revolution
West Bengal, along with Kerala, is an exceptional state in terms of the effort with which the
state government pursued land reforms. The Left Front implemented Operation Barga
rigorously to consolidate its vote base among small farmers, leading to higher sharecropper
registration in areas where it faced greater competition in newly instituted local elections
(Bardhan & Mookherjee, 2010). Registration gave sharecroppers permanent, hereditary tenancy
rights, and limited the share of the crop payable as rent to landlords to 25 percent.4 By 1981 over
1 million sharecropper tenants were registered, and almost 1.5 million by 1990 (Lieten, 1992).
Estimates of the fraction of sharecroppers registered in the state range from 45% (Bardhan &
Mookherjee, 2011), to 65% (Banerjee et. al., 2002), to as high as 80% (Lieten, 1992).
Besides Operation Barga, the state also aimed to vest land held by households above the
stipulated ceiling of 12.5 acres and redistribute it to the landless and small landowners in small
plots (or pattas). Most vesting of land had already taken place by 1978, so the Left Front
government’s main role was in redistributing this land. Appu (1996) estimates that 6.72 percent
of state operated area was distributed by 1992; several times the national average of 1.34 percent.
However, this land was redistributed in small plots (less than half an acre on average in the
sample of farms in Bardhan & Mookherjee (2011)), and was of low quality for cultivation as
landlords would only part with their lowest quality above-ceiling holdings. Hence unlike tenant
registration, land redistribution had virtually no impact on agricultural productivity.
Importantly, there were other government rural initiatives launched in the state at the same time
that were aimed at boosting agricultural productivity and reducing poverty. Alongside Operation
Barga, the state government also distributed minikits containing high yield variety (HYV) seeds,
4
This share rose to 50 percent if landlords provided all non-labour inputs.
7
fertilisers, and insecticides to farmers throughout the state via gram panchayats.5 Land reform in
combination with minikit distribution led to a substantial increase in agricultural yields in West
Bengal over the 1980s, transforming the state into one of the best agricultural performers in the
country and leading this period to be called West Bengal’s Green Revolution. This period is also
associated with significant declines in poverty and growth in rural employment. Banerjee et al.
(2002) attributed the increase in yields to land reform, citing decreased Marshall-Mill
sharecropping distortions from increased tenancy security. Bardhan & Mookherjee (2011)
however shows that while decreased inefficiencies played a role in increasing yields, it was largely
minikit distribution that was responsible for the agricultural growth in this period. 6 Other
programmes administered in the 1980s with gram panchayats targeting local beneficiaries include
the Integrated Rural Development Programme that provided subsidised credit, and employment
initiatives such as the Food for Work programme, the National Rural Employment Programme,
and the National Rural Employment Guarantee Programme.
2.3 Community Differences in Son Preference
The majority Hindu community in India traditionally exhibits greater son preference than
other religious communities, as evidenced by conditional sex ratios in the population and
empirical evidence on child mortality and education that reflect childhood parental investments
(Bhalotra & Zamora, 2009; Bhalotra & Cochrane, 2010; Bhalotra et al., 2010). The literature in
this regard has focused on Hindu-Muslim differences, as other religious communities make up a
very small part of the population.7
While no definitive explanation has been agreed upon for the differing degrees of son
preference between the Indian Hindu and Muslim communities, existing arguments such as the
Dyson-Moore hypothesis base them in marital institutions and inheritance practices. In North
India including West Bengal, Hindu marriage is exogamous for women, who leave their natal
family village to marry into families in villages much further away to avoid marrying a possible
relative. The distance from natal family after marriage reduces Hindu women’s bargaining power
and also their claim to natal family land, which is seen as bringing no reciprocal benefit and lost
The crops for which seeds were distributed were rice, potatoes, oilseeds, and some other vegetables according to
Bardhan & Mookherjee (2011).
6 A companion paper Bardhan, Mookherjee, & Kumar (2012) also shows that tenancy reform crowded in large
private investments in irrigation, the growth-inducing effects of which were far greater than those of reduced
Marshall-Mill distortions.
7 We do the same in this section, as Hindu and Muslim children constitute about 98% of our estimation sample.
5
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to the family when daughters inherit. Sons on the other hand care for parents and natal family
members in their old age by remaining with the natal family and working the family land,
eventually inheriting it upon the death of the family patriarchs. Cultural taboos against Hindu
women sharing public spaces with men and working agricultural land also often prevent them
from claiming and cultivating land (Agarwal, 2003). The bridal dowry practice also often entails
loss or mortgage of family land at the time of a daughter’s marriage. With regard to Operation
Barga specifically, Gupta (2002) finds from interviews of 870 households in two West Bengal
districts that 99% of households reported dowry being a serious concern, and that mortgaging
barga land to meet dowry payments was a common practice. She also finds that dowry was largely
a Hindu practice, but that the custom has penetrated younger generations of Muslims.
Under the Mitakshara Hindu doctrine followed in North India, women in fact have no claim to
joint family property, whereas men are entitled at birth to a share of such family property held by
their fathers, paternal grandfathers, and paternal great-grandfathers. 8 In South India close-kin
marriages are more prevalent for Hindu women, allowing them to inherit a greater share of
ancestral land despite prevailing Mitakshara doctrine as they reside close enough to participate in
cultivation on natal family land after marriage. These marital institutions have been used to
explain more favourable female-male sex ratios in South India compared to North India
(Chakraborty & Kim, 2010). In West Bengal the Dayabhaga Hindu system of inheritance is
followed where the concept of joint family property is absent, and all of a Hindu male’s property
is subject to equal claims by his widow, sons, and daughters upon his intestate death (Lingat,
1973). While this appears more gender-equal than the Mitakshara system in theory, in practice
Hindu women nearly always relinquish their inheritance claims to their brothers and sons so as to
avoid social exclusion, intimidation, and losing the family safety net in times of financial crisis
(Agarwal, 2003). Hindu upper caste women also do not physically work agricultural land due to
prevailing social norms. Lower caste women have higher work-force participation rates in
agriculture as wage labourers, but still female employment rates in agriculture in the state have
been persistently low. 9 Hindu women therefore are very much financially dependent on their
male kin, leading them to give up their rights to family land to avoid losing that support.
Some Indian states have since made reforms to the Hindu Succession Act of 1956 to give women equal inheritance
rights to joint family property, but these reforms still explicitly exclude agricultural land from their purview.
9 In the 1991 Census of India only 11.1 percent of women in West Bengal reported having any form of
employment, and only 54.1% of the employed women were cultivators or agricultural labourers. National Sample
Survey data also reflects decreasing female rural employment and increased casualisation of female agricultural
labour since the late 1980s.
8
9
Muslim communities follow inheritance practices based in the Shariat, which guarantees women
at least half as much inheritance as their closest male counterpart inheritors. Consanguineous
marriage is also practiced to keep all ancestral property within the family, allowing Muslim
women to remain close to their natal families after marriage and inherit more family property in
practice similar to Hindu women in South India.10 Marital dowry is also less prevalent among
Muslims, and abortion, sex selective or otherwise, is strictly forbidden under the Shariat. The
effect of these institutions arguably reduces parental neglect of Muslim female children
compared to Hindu female children in many parts of the country including West Bengal, despite
the fact that the Muslim minority population experience nationally higher levels of poverty than
the Hindu majority and Muslim female labour force participation in West Bengal is even lower
than that of Hindu women (Nasir & Kalla, 2006; Chakraborty & Chakraborty, 2010).
3. Causal Impacts and Identification
3.1 Mechanisms of Interest
The preceding discussion of marriage and inheritance norms suggest Operation Barga could
intensify male-biased fertility and child investment patterns, especially for Hindu families whose
land endowments increased as a result of the reforms. Beneficiaries of the programme and
agricultural tenants in general benefited from increased land security and a greater share of
agricultural output due to the programme, while landlords faced reduced land security and rents.
As outlined previously, the programme also created indirect general equilibrium effects in the
economy that affected infant mortality rates. The programme increased land productivity across
all farm sizes, albeit without increasing farm employment or wages. It also activated the land
market, with large landowners increasing sales of their holdings to small landowners11. Both the
direct effect of changed property rights and the indirect wealth effects of increased productivity
arising from the programme induce their own substitution and income effects that alter
household desired fertility and health investments in their children. We condition on productivity
increases in our empirical analysis, and focus on the net impact of the combined substitution and
income effects on infant mortality arising from changing property rights. While the income effect
will raise parental values of both boys and girls, the substitution effect is likely to be biased in
favour of sons, and particularly so amongst Hindu households who will not perceive girls to be
Bittles (2002) reported that 23% of Muslims in India practiced consanguineous marriages in 1992–1993.
The selling and exchanging of land for marriage expenses and dowry of Hindu daughters formed a significant
part of the land market transactions brought to life by Operation Barga, presumably increasing son preference
(Gupta, 1993; Kodoth, 2005).
10
11
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“close” substitutes for boys. Also in the presence of primogeniture, and from empirical evidence
on the patterns of how son preference manifests in Indian sex ratios, we anticipate that the
substitution effect is most evident amongst Hindu families who do not have a firstborn son.
It will be helpful to clarify the nature of wealth and substitution effects in parental preferences
and how these may vary with household and village characteristics. We can represent the value
placed on child i by parents in household j in a given village as follows,
vi = (K - δ1 gi) exp[(α – δ2 gi)Wj + π(1 – fj )(1- gi) θ(β){lj + βt(lj)}]
(1)
where gi is a dummy variable taking value 1 if child i is a girl. δ1 > 0 is an intrinsic gender bias
against girl children. Wj is household wealth, α > 0 is a common wealth effect on child values,
and δ2 >0 is a gender bias in the wealth effect. Household wealth Wj is constituted as,
Wj = θ(β){lj + βt(lj)} + F
(2)
where θ is village land productivity, which is an increasing function in the fraction β of cultivable
land that is tenanted and registered under Operation Barga. lj is land owned by the household,
and t(lj) is net land leased by the household. F is household non-land wealth. Households owning
more land lease in less, so t is a decreasing function of l with t(0) > 0, and t(l*) = 0 for some l* >
0. Therefore those with land holdings less than l* lease in land from those with land holdings
greater than l*.
12
The parameter π represents the ‘substitution’ effect (which is larger for
Hindus), that is the added value of having a boy to inherit the household tenancy rights valued at
θ(β){lj + βt(lj)}. The substitution effect manifests only if the household has no firstborn son, i.e.
the dummy variable for the first child in household j being male, represented by fj, equals zero.
The general equilibrium effect of Operation Barga in the village comes through increased village
land productivity β, which applies uniformly for all landowners. A partial equilibrium property
rights effect is represented by the dependence of lj + βt(lj) on β, which is increasing in lj for those
owning less than l* and decreasing in lj for those owning more. We assume that landless
households form the bulk of sharecropping tenants that benefit directly from registration. 13
As a further restriction, land cultivated by the household is l + t(l) which must be non-negative and increasing in l,
so the t function must have a slope less than one in absolute value.
13
Bardhan et al. (2014) reports that 80% of surveyed households were landless or marginal landowners.
12
11
For these landless households, Operation Barga unambiguously raises household wealth and
value placed on children through a direct improvement in property rights on rented land and
increased village land productivity β. The curvature of the logistic function in (1) also means that
for a unit increase in wealth, there is a larger increase in child value the smaller the size of the
household’s landholding. However, due to the gender bias parameters δ1 and δ2 the value placed
on boys will increase more than for girls, especially in Hindu families.
For landowners with holdings greater than l*, the sign of the wealth effect is ambiguous as the
general equilibrium effects and property rights effects operate in different directions: we cannot
sign the slope of θ(β){lj + βt(lj)} with respect to β, unless the θ function is assumed to be rising
fast enough. Likewise the substitution effect cannot be signed. In any case, the curvature of the
logistic function implies that a unit increment in wealth will lead to smaller increases in child
value for this category of households than for landless households.
14
Additionally, larger
landowning households are wealthier than the landless, and therefore place a higher value on
children to begin with.
Increasing values placed on children will translate into reductions in infant mortality rates.
Effects on fertility are more complicated, owing to possible quantity-quality trade-offs of the
sort emphasized by Becker and Lewis (1973). Son preference and male-biased inheritance
patterns also may lead to a fertility decline that manifests through son-biased fertility stopping,
which would increase infant survival amongst children who grow up in smaller families.
3.2 Other Programmes and Identification
Bardhan, Mookherjee, and Kumar (2012) find that sharecropper registration crowded in
significant private irrigation investments that triggered large productivity spill-overs across both
tenant and non-tenant farms. Hence Operation Barga potentially had large indirect impacts on
small cultivators and wage labourers who were not involved in sharecropping as landlords or
tenants. The 1980s period of accelerated rice and foodgrains production in West Bengal also saw
disproportionate increases in consumption expenditure by the poor and reduced consumption
inequality (Chattopadhyay, 2005). Reduced groundwater costs from irrigation investments
undoubtedly played a role in this period of high agricultural growth and labour-absorptive
We include marginal landowners with lj less than l* in the same category as large landowners in the empirical
analysis. This allows for clean estimation of the effects of improved property rights on infant mortality among the
landless without confounding wealth effects from landholdings, and serves only to bias estimated coefficients
towards zero among landed households.
14
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expansion. As men were the primary wage labourers and small cultivators in the state, their
increased incomes from registration-induced irrigation investment could have increased son-bias
in child health investments among these households if the substitution effect dominated the
income effect of higher labour returns on these investments.
HYV minikit distribution was however the likely primary driver of yield increases and
agricultural growth in West Bengal in the 1980s as noted earlier. The simultaneity of this
programme with sharecropper registration requires us to account for the effects of increased
land productivity on son preference to empirically identify the impact of registration. We
therefore use an estimation approach that conditions on the effects of regional land productivity,
while estimating the impact of sharecropper registration on infant mortality risk.15 By doing this
we control for the combined income and substitution effects of any yield increases in the
aggregate economy via HYV minikit distribution and other programmes on son-bias in parental
health investments that determine infant mortality. At the same time, conditioning on
productivity effectively conditions on income and substitution effects from increased regional
agricultural output arising from sharecropper registration as well, whether from crowded-in
private irrigation investments or increased returns from improved input use on sharecropped
land. This allows us to isolate the net effect of sharecropper registration on infant mortality
through causal channels which are based in institutional preference structures and property rights
improvements, independent of the effects of increased regional yields.
From the discussion thus far, we can make a number of predictions about what the empirical
analysis should uncover. Averaging across households in the village, if there are enough landless
and marginal landowning households relative to the landed, the average effects will be
qualitatively similar to those for the former category. Specifically, village average infant mortality
amongst boys will decline by more than amongst girls. These results will be robust to controls
for productivity, and will be more pronounced for Hindu households. Disaggregating across
household categories, these effects are expected particularly for the landless, and for Hindu
families without a firstborn son. We can further predict that the sex ratio at birth will become
more male-biased among Hindu families without firstborn sons, and that fertility stopping will
become more son-biased in response to land reform.
Rice is the major crop in West Bengal, accounting for more than 70% of gross cropped area consistently during
1971-1991 according to state government economic reviews.
15
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4. Data and Descriptive Statistics
4.2 NFHS Survey Data
The first dataset we use is the National Family and Health Survey (NFHS). The survey uses
interviews with women aged 15-49 to collect information on their fertility history, their children’s
health and mortality outcomes, household wealth, and their own health and education.
Households are chosen for interview using stratified random sampling, with the probability of
selection weighted by the area population as per the last Indian national census. These data
provide us the advantage of being able to exploit information on children’s district and year of
birth, gender, and birth order to analyse land reform impacts across districts and time by child
gender, and by whether the firstborn child in the household is male or female. We use the 199293 and 1998-99 waves of NFHS data for West Bengal, and match surveyed women’s district of
residence to available data on annual district rates of sharecropper registration from 14 districts.
The registration data is from Banerjee et al. (2002), kindly granted to us by the authors.
Panels A and B of Table 1 outline descriptive characteristics of children in the NFHS sample
born during 1967-93 and their mothers. Panel A shows that the 6,468 mothers in this sample
have 3.42 years of education on average, and have an average of 3.39 births. 75% of mothers are
Hindu, and 20.5% of them belong to a scheduled caste or tribe. 68% of mothers reside in rural
areas, and the average age at which they give birth is 19.03 years. Panel B shows that among the
20,148 children born during these years, the neonatal and infant mortality rates are 6.4% and
9.4% respectively. The probability that a child is male is 51.1%, and the probability the child has
a younger sibling is 71.8%.
We further use district-level data on yields and area under cultivation of rice in West Bengal from
the ICRISAT Village Dynamics in South Asia (VDSA) database to construct measures of annual
district rice productivity in thousands of tonnes of output per one thousand hectares.16 We also
collect information from the annual Economic Survey reports of the West Bengal government
to control for the effects of other programmes and infrastructure in our 14 districts. We
specifically use information on the annual number of medical institutions per capita, kilometres
of surfaced roads per capita, and hectares of patta land distributed per capita in each district.
Panel C of Table 1 shows descriptive statistics of annual district rice productivity for our 14
We construct and control for similar productivity measures of other cereals, but we do not report these
coefficients as these crops form a very small part of total crop yields in West Bengal as described in Section 3.
16
14
sample districts, as well as annual measures of these other district healthcare coverage and
infrastructure measures for the years 1977-1990.17 On average, 1.47 thousand tonnes of rice were
produced per thousand hectares in each district over this period. Also on average each individual
had access to 0.21 kilometres of surfaced road and 0.06 medical institutions in each district over
the same period. We condition on the logs of district rice productivity and of the district
healthcare and infrastructure measures explicitly in our regressions to ensure that our results are
in fact capturing the impact of land reforms rather than alternative policy measures.
Figure 1 shows the evolution of the mean registration rate across our 14 districts over time.
Sharecropper registration appears to have been occurring most quickly up to 1985, after which
the pace of registration slows considerably and remains slow into the 1990s. We exploit this fact
in our robustness checks, when we verify that sharecropper registration was not targeted to
districts with higher pre-reform trends in the female-male difference in mortality and sex ratio at
birth, or gender-biased fertility stopping. In regressions using this registration data, we restrict
the sample to cohorts born after 1978 as there is no positive registration recorded in the data
before this year, despite registration having taken place under the previous government. The final
cohort in our sample is those children born in 1991, as we do not have information on the other
district-level programmes besides land reform outlined above after this year.
4.3 Village Survey Data
Our second dataset is compiled from primary survey data collected from household heads in
West Bengal, regarding their family history, land ownership, immigration status, and other
household characteristics since 1967.18 The survey questionnaire elicited information from the
head about all members residing in the household in 2004, and the year they were born or joined
the household. The survey reports the births of children in the household who have survived till
2004. Hence our estimates using this data will capture the combination of both household
fertility responses and as well as changes in child survival probability, or birth-cum-survival, in
response to land reform. We do not disaggregate land reform impacts on birth-cum-survival by
the gender of the firstborn child as we do with the NFHS data, as non-surviving children are not
reported in survey, making inferring birth order difficult.
These are the years for which these measures enter the regressions as controls; see Section 5.
Further details on this survey data are in Bardhan & Mookherjee (2010), Bardhan & Mookherjee (2011), and
Bardhan et al. (2014).
17
18
15
The survey data was used to construct a household panel for a sample of 2,400 households from
89 villages. For approximately two-thirds of this sample, a consistent history of household
landholdings and demographics could be constructed (comprising the ‘restricted sample’). For
the rest a consistent history could be constructed under specific assumptions of nature of recall
errors. Accordingly, while we report only results from the restricted sample, we verify
independently that the results do not differ qualitatively in the full sample.19
We further combine the household survey data with data on land reforms implemented in each
of the 89 surveyed villages during 1971-2003 for both the land distribution and tenant
registration arms of the land reform programme, collected directly from Block Land Records
Offices. This data on land reforms is of significantly higher quality than the district-level
registration rates in Banerjee et al. (2002) that we match with the NFHS data, as it is compiled
firsthand from official land records rather than from indirect sources. Further, it directly
measures cultivable village land affected by land reform every year, rather than rate of
registration among individual sharecroppers which may not accurately capture the extent of
sharecropping tenancy and sharecropped land in a given geographical region. As such, this data
is a unique resource on land reforms in West Bengal. As a further advantage, this data allows us
to measure the progress of both the distribution and tenant registration arms of the land reform
programme in the actual village where surveyed households reside, which is a much more direct
and local measure of household exposure to land reform than registration rates reported at the
more aggregate geographic and administrative level of the district.
We measure the extent of reform activity as percent village cultivable land distributed to
households in small plots (or pattas), and percent cultivable village land registered as under
sharecropping (under barga) in the three years prior to a reported surviving birth in a household
residing in that village. The survey data show that approximately 15% of surveyed households
had received patta land by 1998. However, the distributed plots were small and of poor quality,
and were also not eligible to be used as collateral for subsidised credit. Land distribution was
therefore largely ineffective in increasing rice productivity. In contrast, plots registered under
barga were of a much larger size (1.5 acres on average), and could be used as collateral for loans
from state financial institutions, therefore yielding much greater positive impacts on rice
productivity than patta land distribution (Bardhan & Mookherjee, 2011). The survey data show
that about 48% of sharecropper tenants, or 6% of surveyed households, had been registered by
19
The results from the full sample are in Table A.8 in the Appendix.
16
the late 1990s. Figure 2 shows the average share of village cultivable land registered over the
years 1867-1998 for the 89 surveyed villages. As in Figure 1, we see that registration peaks in
1985. The decline in average share of cultivable land thereafter is due to the slowdown in
registration, combined with an increase in village cultivable land on average over the 1980s.
Empirical Strategy
5.1 DHS Data: Quadruple Difference-in-Differences
Using the NFHS surveys and the accompanying data on sharecropper registration, we exploit
the variation in registration intensity across districts along with variation in child gender and the
gender of the firstborn child in the household in a quadruple difference-in-differences strategy
to identify programme effects on infant mortality. We implement the following specification,
yijt = α + β1 SCROP50jt-1 * FIRSTSONi * FEMALEi
+ β2 SCROP25 jt-1* FIRSTSONi * FEMALEi
+ δ1 SCROP50 jt-1 * FEMALEi + δ2 SCROP25 jt-1* FEMALEi
+ η1 SCROP50 jt-1 * FIRSTSONi + η2 SCROP25 jt-1* FIRSTSONi
+ φ1 FIRSTSONi + φ2 FEMALEi + φ3 FIRSTSONi * FEMALEi
+ γ1 SCROP50jt-1 + γ2 SCROP25jt-1 + λ Xit + dt + θj + εit
(3)
where yijt is a dummy variable taking value 1 if child i born in district j in year t dies aged 0-12
months and 0 otherwise when we are considering infant mortality. We define dummy variables
SCROP25jt-1 and SCROP50jt-1 which take value 1 if sharecropper registration rate in the district j
where child i resides reaches at least 25% or 50% respectively in the year preceding his birth year,
and 0 otherwise. These are our measures of low and high intensity sharecropper registration. 20 It
is worth noting that these measures are such that SCROP25jt-1 is always equal to 1 when
SCROP50jt-1 is equal to 1, so that high intensity sharecropper registration captures a cumulative
effect of programme intensity on infant mortality. The omitted category of children constitutes
those born in districts where registration is less than 25% in the year preceding birth. We further
define a gender indicator FEMALEi taking the value 1 if child i is a girl, and the indicator
We chose these levels of registration as cut-offs for our treatment indicators based on estimates from a flexible
specification that tested for significant effects of cumulative sharecropper registration rates in 10 percent
increments. These results are available from the authors upon request.
20
17
FIRSTSONi taking value 1 if the firstborn child born to the mother of child i is male. The
regression is limited to children of birth order 2 or higher.
Table 2 shows the evolution of sharecropper registration rates for the 14 districts in our sample.
The light grey cells indicate the year that 25% registration of all sharecroppers is achieved, and
dark grey cells indicate the year when 50% registration is achieved. There is a lot of variation
across districts in the years in which these thresholds are achieved in programme implementation
that we exploit in specification (3). This also alleviates concerns that these years coincide closely
across districts, which would mean we have little dynamic variation in implementation and our
estimates are largely driven by cross-sectional variation across districts. We also report results
from estimating (3) including children born in border districts in the neighbouring state Bihar as
a control group in the regression sample, as these children are never exposed to land reforms.21
The β parameters on the firstborn son-female child-registration rate interaction terms will
therefore capture the differential impact of sharecropper registration intensity on infant
mortality of female children with firstborn older brothers. These impacts are over and above
registration impacts on infant mortality of girls with firstborn older sisters captured in δ
parameter estimates, and on the infant mortality of boys with firstborn older brothers captured
in η parameter estimates. The programme effects for boys with older sisters are captured in the γ
parameter estimates, and the untreated counterpart estimates are captured in the φ estimates. The
four dimensions across which we are taking differences to achieve identification are therefore
districts, year of birth, child gender, and the gender of the firstborn child in the household. The
impacts are identified independently of child birth year and district fixed effects captured in
dummies dt and θj. We also present estimates from mother fixed effect specifications, and after
including a district-specific linear trend in child birth year to control for district trends in
unobservables that may simultaneously determine sharecropper registration rates and infant
mortality risk. The covariate vector Xit includes indicators for child birth order, indicators for
household religion and caste, whether the household is rural, mother’s educational attainment,
and finally linear and quadratic terms of the mother’s age at which the child is born. εit is an
idiosyncratic error term.
These specifications do not include district health and infrastructure programme regressors, as these are not
available for Bihar.
21
18
To explicitly control for the effects of minikit distribution and similar initiatives on infant
mortality via increased agricultural acreage and yields, we estimate specifications including the log
of district productivity of rice (LN RICE PRjt-1) in the year prior to the child’s birth as a
regressor, along with the corresponding interaction terms with the firstborn son and female child
indicators. To further control for any confounding effects of public health improvements,
infrastructural development, and the other arm of the land reform, we also include controls for
the logs of medical institutions per capita, kilometres of surfaced road per capita, and hectares
of patta land distributed per capita in the district in the year preceding the child’s birth along with
the necessary interaction terms.
Differential fertility-stopping by child gender may have increased due to sharecropper
registration, as discussed in Section 3. Again exploiting the plausible hypothesis that the gender
of the firstborn child is random, we re-estimate (3) with the new outcome variable indicator
taking value 1 if child i has a younger sibling and 0 otherwise. However, in this specification we
are only concerned with how having a firstborn son alters fertility-stopping as sharecropper
registration increases, and not how this varies with the gender of child i. We therefore do not
include the interaction terms of the firstborn son indicator with the female child indicator, the
two-way interaction terms of the female child indicator with the registration intensity indicators,
or the three-way interaction terms of the female child, firstborn son, and registration intensity
indicators. We also estimate these specifications sequentially for separate samples of children by
ascending birth order, so as to isolate the level of total fertility at which households alter
childbearing in response to land reform.22
If households also alter their sex selective behaviour in response to land reforms, it would
endogenously change the sample composition of births we consider in the infant mortality
regressions, and potentially worsen the child sex ratio at birth; an even earlier stage than infancy.
To examine whether this is the case, we use the same specification as for the fertility-stopping
analysis, and estimate it for the outcome variable taking value 1 if child i is male and 0 otherwise.
These estimations will reveal if sex selection and postnatal discrimination act in a mutually
reinforcing fashion in response to land reforms.
The results show no impact of land reform on fertility-stopping after the first birth, and are in Appendix Table
A.1. The first statistically significant effects are on the probability of births following the second birth, so we do not
report results for births at birth order three or higher, as these samples are endogenously selected.
22
19
We estimate (3) and its variations for fertility-stopping and the probability of male births using
OLS on the sample of children of birth order two or higher born during 1978-91.23 We carry out
separate estimations for Hindu and Non-Hindu children to account for the different preference
structures over child gender between communities. As there are only 14 districts, we report
results with wild cluster-bootstrapped standard errors as recommended in Cameron et al. (2008),
using the procedure outlined in Busso et al. (2013).
5.2 Test for Targeting of Sharecropper Registration
Our results are subject to reverse causality bias if sharecroppers were more rapidly
registered in districts where families, particularly Hindu families, historically discriminated more
against daughters when they did not have firstborn sons, committed more sex selection, or
engaged in greater son-biased fertility stopping than in other districts prior to Operation Barga.
To test for whether such targeting of registration took place, we assign districts as “treated” or
“untreated” if they had achieved above or below-median levels of registration in 1985
respectively. This exploits the fact that registration occurred at its fastest pace during 1980-1985,
as we see in Figure 1. We then regress our outcomes amongst children of birth order 2 or higher
born before the programme during 1958-77 on a linear time trend in years, interacted with the
above-median registration treatment indicator in 1985. This reveals if pre-programme trends in
our outcomes of interest in a district correlate with that district becoming a “treated” district in
the future. We further disaggregate this correlation between pre-reform trends in outcomes and
registration intensity in 1985 by child gender and gender of the firstborn child. The specification
we estimate is the following,
yijt = α + β1 TRT * TREND * FIRSTSONi * FEMALEi
+ β2 TRT * TREND * FEMALEi + β3 TRT * TREND * FIRSTSONi
+ λ Xit + dt + θ + εit
(4)
where TRT is the indicator for above-median district registration in 1985, and TREND is the
linear time trend for years 1958-77 during which the children in this pre-reform sample are born.
The covariates included in Xit are the same as in (3), except for the controls for other district
programmes and infrastructure which are not available for the pre-reform years. The covariates
We also check for consistency of estimates by including firstborn children in the sample and coding the firstborn
son indicator as zero for these firstborns, and also by restricting the sample to the first two children only. The
results do not change qualitatively, and are available from the authors upon request.
23
20
also include all the remaining three-way and two-way interactions between TRT, TREND, and
the female child and firstborn son indicators.
5.3 Village Data: Fixed Effect Regressions
The main dependent variable we examine in the household head survey data is the event
that a boy or girl who survived until 2003 was born into a household in a given year. Differences
in effects between boys and girls are likely to reflect gender differences in mortality, as chances
of sex-selective abortion during this period in rural West Bengal were low. Since reported births
are of children who survived till 2003, this outcome measures both birth and survival probability.
The specification we use is,
yijt = α + β1 BARGA_LANDjt-1 + β2 PATTA_LANDjt -1 + λ Xijt + dt + θi + εijt
(5)
where the outcome yit takes value 1 if a surviving boy (or girl) is born in household i in village j in
year t and 0 otherwise. This outcome variable captures birth-cum-survival, as it takes value 1
when the surviving child born in year t is a boy (girl), and value 0 if there is no birth in year t, or
if there is a surviving birth in year t that is a girl (boy). Hence estimating (5) separately for
surviving male and female births will capture increases in gender-specific survival rates in
positive coefficients, and declines in fertility combined with reduced survival rates in negative
coefficients. BARGA_LANDjt-1 and PATTA_LANDjt-1 measure the share of cultivable land in
village j that was registered and distributed respectively in the three years preceding year t. The
coefficients β1 and β2 capture respective impacts of the two arms of the programme. While land
distribution benefited many more farmers than tenancy registration, distributed plots were
generally smaller and of lower quality than those covered by tenant registration, and were also
not eligible to be used as collateral for subsidised credit (Bardhan & Mookherjee, 2011). We
therefore do not report the estimates of β2.24 We also do not disaggregate land reform impacts
by the gender of the firstborn child as we do in specification (3), as we do not observe the birth
order of children present in the household in the village data. The terms dt and θi are year and
household fixed effects respectively, and εijt is an idiosyncratic error term.
The regressors Xijt include lagged land owned by the household, an above-ceiling indicator
(whether it owned more land than permitted by the land ceiling), and immigrant status indicators
24
These estimates are available from the authors upon request.
21
(whether the household immigrated after 1967, and year of immigration) which are available for
the full time period of 1971-2003.25 In alternate specifications we include controls constructed
from the farm-level dataset of Bardhan and Mookherjee (2011), which include logs of annual
village rainfall, village land productivity, price of rice, local government expenditures on roads
and irrigation, and kilometres of surfaced road and area irrigated by canals in the district. These
regressors control for potential confounding effects of local public spending on other
programmes and economic shocks. This information is only available for years 1982-95,
restricting the time coverage of the sample, but including these regressors does not qualitatively
change the results. Further, we also estimate (5) inclusive of district-year fixed effects, which
controls for all time-varying unobservables at the district level. The majority of public
infrastructural and health spending decisions affecting individual villages are made at by the
district level of government, making this a relevant robustness check.26
As with specification (3), we present separate estimates for Hindu and Non-Hindu households.
We further disaggregate results by whether households own land, and by household immigrant
status. This is because landless immigrant households are the most likely beneficiaries of tenancy
registration, while native landed households are the most likely to lose privileges on land. We
define landowning classes using pre-reform reported household landholdings in 1977, to avoid
endogenous sample selection on landholdings that may change due to the reform. We also
estimate (5) on years 1978 and after to match the cohorts used to estimate (3).
6. Results
6.1 Firstborn Sons, District Registration Rates, and Mortality
Table 3 reports the estimates from (3) without rice productivity controls and district linear
trends for infant mortality.27 In the pooled sample of children, we find in column (1) that once
districts achieve 50% registration, boys born the following year experience a 5.0 percentage point
decrease in infant mortality risk. The effect is highly significant at the 1% level. There is no
differential effect on mortality risk for boys by gender of the firstborn child. Girls with firstborn
In some specifications we include the share of cultivable village land transacted in the three years preceding year t
to examine if income effects from these post-reform transactions explain any impacts we find. The results do not
change significantly, and are available upon request.
26
Results from specifications including district-year fixed effects are in Appendix Tables A.7. The results including
the full set of controls are available from the authors upon request.
27
The results including children from untreated border districts in neighbouring state Bihar are in Appendix Table
A.2, and show identical patterns to those reported here. The estimates for rice productivity controls along with
those in Table 3 are reported in Appendix Table A.4.
25
22
older sisters however experience no similar reduction in infant mortality risk, as the 50%
registration-female interaction term shows a 5.1 percentage point higher estimated infant
mortality risk than boys, which is significant at the 1% level. Girls with a firstborn older brother
in contrast have a similar or even marginally larger reduction in infant mortality risk than boys as
indicated by the 50% registration-female-firstborn son interaction term, with the estimate
suggesting a 6.0 percentage point decline in their infant mortality after high registration,
significant at the 10% level.28 In column (2) we introduce the log of district rice productivity and
its corresponding interactions with the female child and firstborn son indicators as controls. The
results are largely unchanged, except for increasing the infant mortality decline for girls with
firstborn older brothers to 7.5 percentage points. In column (3) we additionally control for
district-specific linear time trends. Including the trends increases the estimated mortality decline
for male children to 6.3 percentage points once district registration exceeds 50%, and the
coefficient remains highly significant at the 1% level. We continue to find that girls without
firstborn older brothers have unchanged, or even marginally greater infant mortality risk of up to
1.7 percentage points at high registration, while girls with firstborn older brothers experience an
estimated reduction in infant mortality risk of 9.0 percentage points.
We next examine community-specific results by re-estimating the specifications in columns (1)(3) for Hindu and Non-Hindu children separately. The results are reported in columns (4)-(6)
and (7)-(9) respectively.
Columns (4)-(6) show an identical pattern of results for Hindu
households to those for the full sample in columns (1)-(3). In column (4), boys show an
estimated infant mortality decline of 4.0 percentage points once district registration exceeds
50%, marginally significant at the 10% level. Girls without older firstborn brothers show no
mortality decline in contrast, with the estimate even suggesting an increase in infant mortality of
up to 2.3 percentage points after high registration. Girls with firstborn older brothers however
exhibit a decline in infant mortality of 5.1 percentage points. The introduction of controls for
log of district rice productivity and its corresponding interaction terms with the firstborn son
and female child indicators in column (5) increases the estimated infant mortality decline for
boys to 5.2 percentage points in response to high registration. Girls without firstborn older
brothers show no decline in infant mortality risk, or even a marginal increase of up to 1.7
percentage points. Girls with firstborn older brothers experience a reduction in mortality risk of
8.0 percentage points, larger than in column (4) before productivity controls were introduced.
The estimated effects for girls with firstborn older brothers are arrived at by adding the statistically significant
coefficients from the two-way and three-way interactions terms of the firstborn son, female child, and registration
intensity indicators.
28
23
Including the district-linear time trend in column (6) strengthens the results further, with the
infant mortality reduction for boys rising to 5.9 percentage points, more precisely estimated at
5% level significance. Girls without firstborn older brothers continue to show no such decline,
with even a suggested increase in infant mortality of up to 1.1 percentage points. Girls with
firstborn brothers show a mortality decline of 8.6 percentage points, which remains significant at
the 5% level. Therefore, the estimates suggest that as sharecropper registration increases, Hindu
families invest in the health of their sons, but only do the same for their daughters if they already
have a firstborn son.
For Non-Hindu children, we find in column (7) that district sharecropper registration exceeding
50% reduces infant mortality among all children, regardless of their gender and the gender of
the firstborn child, by a highly significant 7.2 percentage points. This is evident from the fact that
nearly all of the two-way and three-way interaction terms of the high registration indicator with
the female child and firstborn indicators show no statistically significant differential impacts from
the baseline estimated effect of registration on male children. The only such differential impact
is found for male children with firstborn older brothers, who appear to benefit less than male
children with older firstborn sisters, but even this differential effect vanishes in columns (8) and
(9) once rice productivity controls and the district-linear time trend are included in the
specification. The inclusion of the rice productivity controls in column (8) also reduces the
estimated reduction in infant mortality from high registration to a statistically insignificant 4.6
percentage points, indicating that productivity is an important competing mechanism in infant
mortality reduction that correlates with sharecropper registration as we would expect. However
the further introduction of the district-linear time trend in column (9) again increases this
estimated beneficial impact of high registration to 7.8 percentage points, which is significant at
the 5% level. The district-linear time trend therefore performs an important role in capturing
otherwise omitted district-specific differential trends in infant mortality that bias our coefficient
of interest towards zero. Columns (7)-(9) therefore show that Non-Hindu families, unlike Hindu
families, do not discriminate by child gender when distributing the health investment gains from
district sharecropper registration amongst their children.
24
6.2 Gender-Differentiated Fertility Stopping
The estimated impacts of sharecropper registration on the probability a child of birth order 2
has a younger sibling are in Table 4.29 In columns (1)-(3) we find a large, stable, and statistically
significant reduction of 11.4-11.8 percentage points in the probability of a third birth once
district sharecropper registration exceeds 25%, but only when the first child in the family is male.
This indicates that land reforms did induce households to reduce fertility, but the fertility
stopping response was significantly son-biased. There is a suggestion of a smaller fertility decline
in response to land reform among families with firstborn daughters as well in column (2), upon
introducing rice productivity controls to the specification, but this effect does not survive the
further introduction of district-linear time trend controls in column (3). Nevertheless, the
coefficient does increase substantially upon the introduction of the rice productivity controls,
indicating the importance of separating the fertility impacts of increasing rice yields due to
registration and other programmes from those of registration. These results also show that
reduced sibling competition for resources is a potentially relevant causal channel for the
estimated infant mortality reduction we find for boys as well as girls with older firstborn brothers
in response to land reforms.
Columns (4)-(6) of Table 4 show the results for the Hindu sample. The same pattern of effects
is found as in the pooled sample in columns (1)-(3), with the probability of a third birth
declining by a statistically significant 10.4-10.8 percentage points once district registration
exceeds 25% , but only if the first child is male. There are no perceptible effects on fertility
stopping after the second birth if the first child is a daughter, again indicating a substantially sonbased fertility stopping response by Hindu households in response to land reforms. Columns (7)(9) report the corresponding results for Non-Hindu households. Notably, the fertility decline
among these households in response to land reforms is much larger than in Hindu households.
In column (7), there is a marginally significant decline in the probability of a third birth of 7.9
percentage points irrespective of the gender of the first child once district registration exceeds
50%. This effect increases to 8.7 percentage points in column (8) once rice productivity controls
are included, and is also more precisely estimated, achieving significance at the 5% level.
Introducing the district-linear time trend in column (9) increases the estimated effect to 9.6
percentage points, still marginally significant at the 10% level. Over and above this decline in
The results including children from untreated border districts in neighbouring state Bihar are in Appendix Table
A.3, and show identical patterns to those reported here. The estimates for rice productivity controls along with
those in Table 4 are in Appendix Table A.5.
29
25
fertility in response to land reforms, there is an additional large and statistically significant decline
in third births once district registration exceeds 25% in all columns (7)-(9) when the first child is
male. This additional son-biased fertility stopping response to land reforms is nearly twice as
large in Non-Hindu families as in Hindu families, at 17.9-18.2 percentage points. Hence NonHindu families also appear to exhibit son preference, but do so by adjusting their fertility rather
than by increasing post-natal discrimination against girl children.
6.3 Sex Ratio at Birth
In Table 5 we report the estimated impact of sharecropper registration on the probability a
birth is male.30 For the pooled sample of all children, we find in column (1) that registration has
no effect on the sex ratio at birth for firstborns. In column (2), we find that the probability of a
birth at birth order 2 or higher being male increases by 3.9 percentage points after district
registration exceeds 50%. The estimate is significant at the 10% level. This estimate increases to
4.5 percentage points in column (3) if the firstborn child is female. As in Table 1, we re-estimate
the specifications in columns (1)-(3) on the samples of Hindu and Non-Hindu children
separately, and report the results in columns (4)-(6) and (7)-(9) respectively. We find in columns
(4) and (7) that sharecropper registration has no significant impact on the sex ratio at birth
among Hindu and Non-Hindu firstborns respectively, as in the pooled sample in column (1). We
find, as with infant mortality, that the increased probability of births being male at birth order 2
or higher is driven entirely by Hindu households. The estimates in column (5) indicate that this
probability increases by a highly significant 4.8 percentage points in Hindu families after districts
achieve 50% registration. This probability increases further to 5.5 percentage points when the
firstborn child is a female, and the coefficient is significant at the 5% level. Columns (8) and (9)
show no such effects on the sex ratio at birth in Non-Hindu families, consistent with the
previous literature. Hence, land reform appears to increase both sex selection and postnatal
discrimination in favour of male children in Hindu families, while also exacerbating son-biased
fertility stopping in both Hindu and Non-Hindu families.
The sex ratio results including estimates for log-rice productivity are in Appendix Table A.6. For brevity, we only
report estimates from the full specification inclusive of rice productivity controls and district-linear time trends in
Table 5 for children of birth order 1, and children of birth order 2 or higher separately.
30
26
6.4 Was Sharecropper Registration Targeted?
The results from specification (4) estimated for infant mortality are in Table 6. Across
columns (1)-(3), we find no statistically significant correlations in the pooled sample of children
between the pre-reform trend in infant mortality and the intensity of registration in the district
in 1985, by either the gender of the child or the firstborn sibling. The coefficients are all also
nearly identical to zero. There are similarly no significant correlations in the corresponding
regressions for Hindu and Non-Hindu households in columns (4)-(6) and (7)-(9) respectively,
lending credibility to our results. The results from the same test for the probability of male births
and having a younger sibling are in Table 7. In each of the pooled, Hindu, and Non-Hindu
samples of children, we see no correlations between the pre-reform trend in outcomes during
1958-77 and registration intensity in 1985, with the estimated coefficients close to zero. There is
hence little sign of sharecropper registration being targeted to districts based on pre-reform
trends in outcomes.
6.5 Village-Level Tenancy Registration and Surviving Births
Table 8 shows the estimated impacts from (5) of Operation Barga on the likelihood of
surviving children being born into households in different landowning classes from the village
survey data.31 Columns (1)-(6) show the programme effects on the likelihood of a surviving girl,
and columns (7)-(12) show the effects on the likelihood of a surviving boy. Column (1) shows a
significant negative effect of 0.8 percentage points for each percent increase of village land
registered on the likelihood of surviving girls being present in the household. This effect
captures a combination of reduced fertility in response to registration, and also potentially
increased gender discrimination that reduces female child survival. However, as the NFHS data
shows that household fertility declines significantly in response to tenancy registration, and
female infant mortality does not increase except possibly for the subsample of girls with
firstborn older sisters, the fertility decline is the likeliest dominant channel of impact. In column
(2) we include average farm-level rice productivity in the village in the year prior to the child’s
birth as a regressor to control for the effects of increased income and other programmes such as
minikit distribution. This data is only available for years 1982-95, which reduces the sample
substantially, but the impact of registration on the probability of a surviving girl remains negative
The results including controls for district-year fixed effects, the results from the full sample, and the results
inclusive of all village and district controls for years 1982-1995 are all qualitatively identical, and are in Appendix
Tables A.7, A.8, and A.9 respectively.
31
27
at 1.2 percentage points and is still highly significant. Columns (3) and (4) show no programme
effects in landless families on the probability of a surviving girl. In contrast, the reduced
probability of surviving girls as a result of registration occurs entirely in landed families, with the
coefficient showing a highly significant decline in this probability of 1.1 percentage points in
column (5), and an increased magnitude of 1.6 percentage points conditional on farm yields in
column (6). Notably, increased farm productivity also reduces the probability of a surviving girl
in landed households.
Turning to the likelihood of surviving male children being present in the household, column (7)
shows that tenancy registration significantly increases this probability by 1.1 percentage points.
As the fertility decline we find in the DHS data in response to registration would exert a negative
effect on this coefficient, this positive programme effect is unambiguously due to increased
survival of male children relative to female children consistent with our proposed hypotheses.
The effect continues to be positive and statistically significant at 0.6 percentage points after
conditioning on farm yields in column (8), despite the reduction in sample size. This increase in
male child survival arising from registration is much larger for landless than landed households,
at 2.6 percentage points in column (9) and 5.4 percentage points in column (10) conditional on
farm yields, compared to 0.7 percentage points in column (11), which is again consistent with
our predictions. Increased farm yields also increase the probability of surviving male births in
landless households as we would expect, as the property rights and wealth effects of registration
on fertility are both positive among the landless.
Tables 9 and 10 explore the heterogeneity of the effect on male and female surviving births
respectively by dividing the sample between Hindu and non-Hindu households (classified using
the name of the household head).32 The results show that the findings in Table 8 are driven
entirely by Hindu households, as we find in the NFHS data. In Table 9 we observe a positive
and significant effect of percent village land registered on male surviving births of 2.5 percentage
points and 0.7 percentage points in landless and landed Hindu households in columns (2) and (3)
respectively. The landless show a stronger response consistent with our predictions, and the
coefficient magnitudes are nearly identical to those for the pooled sample in Table 8, indicating
that the positive discrimination in favour of male children is concentrated in Hindu households.
There are no significant programme effects found for non-Hindu households. The
corresponding results for girls in Table 10 show a negative and significant effect of land
For brevity we do not report the results from specifications including average farm rice productivity in further
results from the village data. These specifications show qualitatively identical results, and are available upon request.
32
28
registration on the likelihood of surviving female births in Hindu households of 1.1 percentage
points, coming from landed households in column (3). Hence landed Hindu households increase
health investments in their sons in response to tenancy reform to improve their survival chances
as we see in Table 9, but they do not do the same for their daughters and the fertility decline
effect dominates in Table 10. This is consistent with the evidence from the NFHS data, which
also shows a widening difference in the survival advantage of sons due to tenant registration. It is
worth noting that in non-Hindu households, the coefficient estimates indicate large (but
imprecisely) estimated increases in survival probability for both sons and daughters in Tables 9
and 10, which is again consistent with what we find in the NFHS data.
Tables 11 and 12 examine Hindu male and female child survival probabilities respectively by
household immigrant status. Table 11 shows that the positive effect of registration on Hindu
male surviving births is concentrated in native landed households and immigrant landless
households in columns (3) and (5). Additionally, the increase in survival probability for male
children is greater among the landless than the landed households, consistent with our
predictions. For surviving female births in Table 12, there is a significant effect of the fertility
decline in response to registration among native landed households in column (3), but this is not
offset by a large increase in survival probability, as it is for boys in Table 11. Hence we again
capture greater relative survival gains to sons due to tenancy registration as per our hypotheses.
7. Conclusions
Our findings show a strong exacerbating effect of increased property rights security on
gender discrimination in Hindu families, with parents showing greater son-bias in sex selective
behaviour, fertility-stopping, and postnatal health investments. In Non-Hindu families on the
other hand, there is a decline in infant mortality for both boys and girls in response to tenancy
registration, and son preference only manifests in fertility stopping behaviour. Institutional
differences in son preference therefore play a significant role in determining the gender
distribution of benefits from improving property rights. In contrast, there is evidence that land
reform alters existing gender-unequal institutions in favour of women in alternative settings.
Tenure regularisation has been argued to significantly improve women’s tenurial and inheritance
claims to land in Rwanda (Ali et al., 2014), and joint spousal titling seemingly increased women’s
intra-household bargaining power in Peru (Wiig, 2013). The rigidity of gender institutions
29
governing land ownership is hence also a likely critical factor in the effectiveness of property
rights formalisation in reducing gender discrimination.
The distribution of land rights within the household underpins family welfare in developing
countries such as India where the majority of the population is engaged in agricultural
occupations. As such, while our results illustrate the importance of ensuring women are directly
benefited by property rights formalisation, the potential for resulting intra-household conflict
also requires acknowledgement. Indeed, though land reforms in India have seemingly reduced
overall suicide rates, land reforms that benefited women relatively more than men have been
argued to increase both female and male suicides due to increased familial tensions (Anderson &
Genicot, 2015). Welfare considerations have also become more urgent as rapid industrialisation
has led to increased government acquisition of agricultural land in India and other fast-growing
developing countries, which can reverse welfare gains from tenancy and land ownership reforms
without adequate compensation to farmers who lose their land as a result (Ghatak &
Mookherjee, 2014). Property rights formalisation therefore requires consideration of
distributional considerations both within households and between entire sections of the
population to prevent unintended welfare losses in the economic development process.
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Figure 1: Average Sharecropper Registration Rate by Year
Notes: The figure shows the average rate of completed sharecropper registration across
the 14 West Bengal districts in the Banerjee at al. (2002) data during 1975-1991.
Percent Cultivable Land Registered
Figure 2: Average Cultivable Village Land Registered by Year
Year
Notes: The figure shows the average percent of village cultivable land registered across
the 89 villages from the household panel data survey during the years 1967-1998. The
percent of cultivable land registered declines after 1985 as registration slowed during this
period, while the amount of cultivable land increased on average.
35
Table 1: NFHS and District Summary Statistics
Freq.
(1)
Panel A
Years of Education
Age at Birth
Total Births
Hindu
Scheduled Caste/Tribe
Rural
Panel B
Infant Death
Neonatal Death
Male Child
Has Younger Sibling
Panel C
Rice Productivity
Patta Area Per Capita
Surfaced Roads Per Capita
Medical Institutions Per Capita
Mean
S.D.
Min.
Max.
(2)
(3)
(4)
(5)
Mother Characteristics: 1967-1993
6,443
3.416
4.297
0
18
6,468 19.034 3.532
5
40
6,468
3.386
2.010
1
11
6,468
0.750
0
1
6,468
0.205
0
1
6,468
0.680
0
1
Child Outcomes: 1967-1993
20,148 0.094
0
1
20,148 0.064
0
1
20,148 0.511
0
1
20,148 0.718
0
1
District Productivity and Programmes: 1977-1990
196
1.473
0.434
0.720
2.595
196
6.518
4.937
0.321 17.986
196
0.208
0.067
0.115
0.392
196
0.056
0.016
0.033
0.115
Notes: Panel A shows mother characteristics, and Panel B shows child outcomes for cohorts
born during 1967-1993. Panel C shows productivity and programme statistics in the 14
districts with sharecropper registration data for years 1977-1990, which are the years for
which they enter as controls in the regressions. Neonatal death takes value 1 if the child dies
aged 0-1 months, and infant death takes value 1 if the child dies aged 0-12 months.
Table 2 – District Sharecropper Registration Rates by Year
Kochibihar
Jalpaiguri
Darjeeling
West Dinajpur
Maldah
Murshidabad
Nadia
24-Parganas
Howrah
Hooghly
Midnapur
Bankura
Burdwan
Birbhum
Mean
1980
1982
1984
1986
1988
1990
1992
19.96
18.44
14.49
34.88
44.15
15.82
22.99
15.22
31.05
25.50
16.79
29.69
11.00
25.01
23.21
39.42
30.88
23.40
66.41
60.93
43.01
37.20
38.54
44.56
40.94
44.89
68.67
35.64
59.26
45.27
42.34
32.30
24.52
70.77
69.75
44.69
42.79
42.84
48.67
46.26
55.70
74.48
39.45
72.49
50.50
50.54
34.95
28.22
73.86
74.40
53.09
49.52
47.67
54.63
55.11
61.09
84.46
45.74
91.37
57.48
54.65
40.25
28.84
75.63
76.57
56.80
52.74
49.51
56.12
58.60
62.84
89.66
49.60
95.67
60.53
57.46
40.25
28.84
76.74
78.32
58.89
53.35
49.84
57.32
61.62
63.64
92.56
51.42
98.30
62.04
58.31
40.25
28.84
76.77
79.24
60.79
55.71
54.05
58.22
63.45
63.82
93.83
53.31
100.00
63.33
Notes: The table shows district sharecropper registration rates by year as reported in Banerjee et. al. (2002).
36
Table 3 – Infant Mortality, Firstborn Sons, and Registration
(1)
SCROP50 t-1 * FIRSTSON * FEMALE
SCROP25 t-1 * FIRSTSON * FEMALE
SCROP50 t-1 * FEMALE
SCROP25 t-1 * FEMALE
SCROP50 t-1 * FIRSTSON
SCROP25 t-1 * FIRSTSON
SCROP50 t-1
SCROP25 t-1
FIRSTSON * FEMALE
FEMALE
FIRSTSON
District FE
District Rice Productivity
District-Birth Year Trend
Observations
Pre-Reform Outcome Mean
Cohorts
Districts
All Children
(2)
-0.060*
(0.029)
0.033
(0.033)
0.051***
(0.02)
-0.025
(0.03)
0.028
(0.023)
-0.037
(0.023)
-0.050***
(0.02)
0.019
(0.023)
-0.019
(0.13)
0.192
(0.152)
0.024
(0.114)
x
-0.075*
(0.039)
0.030
(0.038)
0.046**
(0.022)
-0.029
(0.032)
0.040
(0.029)
-0.035
(0.024)
-0.050***
(0.021)
0.012
(0.021)
-0.025
(0.143)
0.163
(0.132)
0.038
(0.132)
x
x
8,367
0.098
1978-91
14
8,367
0.098
1978-91
14
Infant Death
Hindu Children
(5)
(3)
(4)
-0.073*
(0.039)
0.031
(0.038)
0.046**
(0.023)
-0.029
(0.032)
0.039
(0.028)
-0.034
(0.023)
-0.063***
(0.023)
0.001
(0.020)
-0.022
(0.140)
0.164
(0.126)
0.032
(0.134)
x
x
x
8,367
0.098
1978-91
14
-0.074**
(0.032)
0.021
(0.05)
0.063***
(0.023)
-0.024
(0.033)
0.015
(0.028)
-0.041
(0.036)
-0.040*
(0.021)
0.006
(0.03)
-0.062
(0.163)
0.181
(0.147)
0.011
(0.17)
x
-0.097**
(0.046)
0.017
(0.051)
0.069**
(0.03)
-0.025
(0.034)
0.043
(0.031)
-0.034
(0.035)
-0.052*
(0.025)
0.000
(0.029)
-0.104
(0.178)
0.170
(0.134)
0.070
(0.198)
x
x
5,448
0.107
1978-91
14
5,448
0.107
1978-91
14
(6)
-0.097**
(0.048)
0.019
(0.051)
0.070**
(0.029)
-0.027
(0.033)
0.043
(0.032)
-0.035
(0.036)
-0.059**
(0.026)
-0.003
(0.031)
-0.022
(0.140)
0.164
(0.126)
0.032
(0.134)
x
x
x
5,448
0.107
1978-91
14
Non-Hindu Children
(7)
(8)
(9)
-0.033
(0.048)
0.062
(0.041)
0.028
(0.039)
-0.039
(0.045)
0.046*
(0.026)
-0.027
(0.029)
-0.072**
(0.035)
0.051
(0.030)
0.045
(0.222)
0.209
(0.247)
0.016
(0.127)
x
-0.029
(0.054)
0.063
(0.045)
-0.001
(0.046)
-0.046
(0.046)
0.029
(0.031)
-0.031
(0.034)
-0.046
(0.029)
0.035
(0.036)
0.087
(0.236)
0.178
(0.259)
-0.026
(0.148)
x
x
2,919
0.074
1978-91
14
2,919
0.074
1978-91
14
-0.028
(0.055)
0.046
(0.044)
-0.001
(0.046)
-0.036
(0.048)
0.029
(0.031)
-0.016
(0.03)
-0.078**
(0.035)
-0.005
(0.039)
0.079
(0.232)
0.179
(0.256)
-0.024
(0.142)
x
x
x
2,919
0.074
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also include the female child, birth year fixed effects, birth order fixed effects, year of interview fixed
effects, indicators for household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child
is born. Specifications for children of birth order 2 or higher also include the firstborn son indicator and its corresponding interaction terms. The district covariates include logs of patta
land area distributed, number of medical institutions, and kilometres of surfaced road per capita and their corresponding interactions with the female child and the firstborn son
indicators. *** p<0.01, ** p<0.05, * p<0.1
37
Table 4 – Younger Siblings and Sharecropper Registration
(1)
SCROP50 t-1 * FIRSTSON
SCROP25 t-1 * FIRSTSON
SCROP50 t-1
SCROP25 t-1
FIRSTSON
District FE
District Rice Productivity
District-Birth Year Trend
Observations
Pre-Reform Outcome Mean
Cohorts
Districts
All Children
(2)
-0.007
(0.045)
-0.118***
(0.053)
-0.037
(0.029)
0.022
(0.041)
-0.315
(0.209)
x
0.009
(0.044)
-0.114**
(0.053)
-0.048*
(0.028)
-0.001
(0.042)
-0.223
(0.192)
x
x
2,686
0.839
1978-91
14
2,686
0.839
1978-91
14
Child Has a Younger Sibling
Hindu Children
(4)
(5)
(6)
(3)
0.012
(0.044)
-0.116**
(0.055)
-0.048
(0.032)
-0.020
(0.056)
-0.201
(0.193)
x
x
x
2,686
0.839
1978-91
14
-0.011
(0.050)
-0.106**
(0.051)
-0.024
(0.033)
0.016
(0.058)
-0.436
(0.258)
x
0.003
(0.052)
-0.104**
(0.049)
-0.035
(0.034)
-0.008
(0.062)
-0.343
(0.251)
x
x
1,919
0.808
1978-91
14
1,919
0.808
1978-91
14
0.008
(0.051)
-0.108**
(0.052)
-0.038
(0.039)
-0.058
(0.079)
-0.347
(0.251)
x
x
x
1,919
0.808
1978-91
14
Non-Hindu Children
(7)
(8)
(9)
0.011
0.020
(0.053)
(0.049)
-0.179*** -0.179***
(0.078)
(0.082)
-0.079*
-0.087**
(0.04)
(0.037)
0.010
0.000
(0.063)
(0.06)
0.108
0.150
(0.133)
(0.119)
x
x
x
767
0.952
1978-91
14
767
0.952
1978-91
14
0.019
(0.050)
-0.182**
(0.086)
-0.096*
(0.051)
0.036
(0.083)
0.205
(0.138)
x
x
x
767
0.952
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also birth year fixed effects, birth order fixed effects, year of interview fixed effects,
indicators for household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the
child is born. The district covariates include logs of patta land area distributed, number of medical institutions, and kilometres of surfaced road per capita and their
corresponding interactions with the female child and the firstborn son indicators. *** p<0.01, ** p<0.05, * p<0.1
38
Table 5 – Male Births and Sharecropper Registration
B. Ord. 1
(1)
All Children
B. Ord. >1 B. Ord. >1
(2)
(3)
SCROP50 t-1 * FIRSTSON
-
-
SCROP25 t-1 * FIRSTSON
-
-
-0.020
(0.037)
-0.010
(0.034)
0.039*
(0.022)
0.048
(0.047)
-0.011
(0.020)
-0.005
(0.027)
0.045*
(0.025)
0.051
(0.058)
-
-0.007
(0.009)
x
x
x
3,248
0.449
1978-91
14
x
x
x
8,367
0.494
1978-91
14
SCROP50 t-1
SCROP25 t-1
FIRSTSON
District FE
District Covariates
District-Birth Year Trend
Observations
Pre-Reform Outcome Mean
Cohorts
Districts
Child is Male
Hindu Children
B. Ord. 1 B. Ord. >1 B. Ord. >1
(4)
(5)
(6)
-
-
-
-
-0.065
(0.050)
0.063
(0.053)
0.048**
(0.024)
0.031
(0.050)
-0.014
(0.019)
0.005
(0.031)
0.055**
(0.029)
0.030
(0.054)
0.094
(0.096)
-
-0.008
(0.010)
x
x
x
8,367
0.494
1978-91
14
x
x
x
2,323
0.433
1978-91
14
x
x
x
5,448
0.493
1978-91
14
Non-Hindu Children
B. Ord. 1 B. Ord. >1 B. Ord. >1
(7)
(8)
(9)
-
-
-
-
0.091
(0.064)
-0.149
(0.098)
0.042
(0.035)
0.096
(0.077)
0.004
(0.047)
-0.033
(0.062)
0.039
(0.041)
0.111
(0.095)
0.186
(0.113)
-
-0.007
(0.017)
-0.005
(0.262)
x
x
x
5,448
0.493
1978-91
14
x
x
x
925
0.488
1978-91
14
x
x
x
2,919
0.498
1978-91
14
x
x
x
2,919
0.498
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also include birth year fixed effects, birth order fixed effects, year of interview fixed effects,
indicators for household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child
is born. The district covariates include logs of patta land area distributed, number of medical institutions, and kilometres of surfaced road per capita and their corresponding
interactions with the female child and the firstborn son indicators. *** p<0.01, ** p<0.05, * p<0.1
39
Table 6 – Test of Targeted Registration: Infant Mortality
Infant Death
Hindu Children
(5)
(6)
Non-Hindu Children
(7)
(8)
(9)
(1)
All Children
(2)
(3)
(4)
-0.002
(0.005)
-0.005
(0.007)
-0.001
(0.008)
-0.000
(0.004)
-0.004
(0.008)
-0.002
(0.009)
-0.009
(0.009)
-0.011
(0.012)
-0.004
(0.012)
TRT * TREND * FEMALE
-
0.007
(0.008)
0.006
(0.008)
-
0.008
(0.009)
0.008
(0.011)
-
0.006
(0.009)
0.001
(0.009)
TRT * TREND * FIRSTSON * FEMALE
-
-
0.003
(0.006)
-
-
0.001
(0.006)
-
-
0.007
(0.007)
Yes
3,389
1958-77
14
Yes
3,389
1958-77
14
Yes
3,389
1958-77
14
Yes
2,428
1958-77
14
Yes
2,428
1958-77
14
Yes
2,428
1958-77
14
Yes
961
1958-77
14
Yes
961
1958-77
14
Yes
961
1958-77
14
TRT * TREND
District FE
Observations
Cohorts
Districts
Notes: Wild cluster bootstrapped standard errors in parentheses. Samples include children of birth order 2 or higher. All specifications also include the female child and
firstborn son indicators and their two-way interactions with the trend, birth year fixed effects, indicators for household religion and caste, whether the household is rural,
mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child is born. *** p<0.01, ** p<0.05, * p<0.1
Table 7 – Test of Targeted Registration: Male Births and Fertility
All Children
Male Child
Younger Sibling
(1)
(2)
TRT * TREND
TRT * TREND * FIRSTSON
District FE
Observations
Cohorts
Districts
Hindu Children
Male Child
Younger Sibling
(3)
(4)
Non-Hindu Children
Male Child
Younger Sibling
(5)
(6)
0.001
(0.004)
0.001
(0.004)
0.001
(0.005)
0.001
(0.005)
0.002
(0.008)
0.002
(0.004)
-
-0.009
(0.007)
-
-0.012
(0.008)
-
0.003
(0.008)
Yes
3,389
1958-77
14
Yes
1,369
1958-77
14
Yes
2,428
1958-77
14
Yes
1,015
1958-77
14
Yes
961
1958-77
14
Yes
961
1958-77
14
Notes: Wild cluster bootstrapped standard errors in parentheses. Samples for the sex ratio regressions include children of birth order 2 or higher, and of birth order 2
for the fertility regressions. The specifications for the probability of having a younger sibling also include the firstborn son indicator and its interaction with the trend,
birth year fixed effects, indicators for household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of
the mother’s age at which the child is born. *** p<0.01, ** p<0.05, * p<0.1
40
Table 8: Land Reform and Surviving Births
1=surviving girl,
0=no birth or surviving boy
Dep. Variable:
Land category:
All
1=surviving boy,
0=no birth or surviving girl
Landless
Some land
All
Landless
(1)
(2)
(3)
(4)
(5)
(6)
(7)
(8)
(9)
(10)
-0.001
(0.005)
-0.012***
(0.002)
-0.023**
(0.011)
-
0.001
(0.003)
-0.002
(0.016)
-0.001
(0.005)
-0.016***
(0.002)
-0.044***
(0.016)
-0.000
(0.001)
0.011***
(0.002)
-
0.003
(0.004)
0.006***
(0.002)
0.024*
(0.012)
-
0.000
(0.003)
-
-0.001*
(0.001)
-0.011***
(0.001)
-
-
Farm Productivity
-0.001
(0.001)
-0.008***
(0.001)
-
0.026***
(0.009)
-
0.054***
(0.002)
0.029*
(0.016)
Household FE
Observations
Households
Years
x
44,590
2,286
1978-99
x
7,869
1,928
1982-95
x
19,548
1,144
1978-99
x
3,627
892
1982-95
x
25,042
1,142
1978-99
x
4,242
1,036
1982-95
x
44,590
2,286
1978-99
x
7,869
1,928
1982-95
x
19,548
1,144
1978-99
x
3,627
892
1982-95
Agricultural land
% Land registered
Some land
(11)
(12)
-0.000
(0.001)
0.007***
(0.001)
-
0.003
(0.004)
-0.005*
(0.003)
0.023
(0.015)
x
25,042
1,142
1978-99
x
4,242
1,036
1982-95
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land registered is computed as the sum over the previous three years of the share of land
affected by each program over the total cultivable land in each village, using official land records. All regressions also include immigrant status indicators, and year and household fixed
effects, and % land distributed calculated the same way as % land registered. The land category of the household is defined based on year 1977 landholdings, and regressions include data
for years 1978–1999. *** p<0.01, ** p<0.05, * p<0.1
41
Table 9: Surviving Male Births by Religion
1=surviving boy,
0=no birth or surviving girl
Dep. Variable:
Religion:
Land category:
Agricultural land
% Land registered
Household FE
Observations
Households
Years
All
(1)
Hindu
Landless
(2)
-
Non-Hindu
Landless
Some land
(5)
(6)
Some land
(3)
All
(4)
-0.006
(0.004)
0.223
(0.145)
0.317
(0.385)
-0.005
(0.004)
0.152
(0.212)
x
9,572
464
1978-99
x
3,999
210
1978-99
x
5,573
254
1978-99
0.001
(0.001)
0.010***
(0.002)
0.025***
(0.007)
0.001
(0.001)
0.007***
(0.001)
x
35,018
1,822
1978-99
x
15,549
934
1978-99
x
19,469
888
1978-99
-
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land registered is computed
as the sum over the previous three years of the share of land affected by each program over the total cultivable land in each
village, using official land records. All regressions also include a land ceiling indicator, immigrant status indicators, and year
and household fixed effects, and % land distributed calculated the same way as % land registered. The land category of the
household is defined based on year 1977 landholdings, and regressions include data for years 1978–1999. Religion of the
household is the religion of the household head. *** p<0.01, ** p<0.05, * p<0.1
Table 10: Surviving Female Births by Religion
1=surviving girl,
0=no birth or surviving boy
Dep. Variable:
Religion:
Land category:
Agricultural land
% Land registered
Household FE
Observations
Households
Years
All
(1)
Hindu
Landless
(2)
-
Non-Hindu
Landless
(5)
Some land
(3)
All
(4)
-0.005***
(0.002)
0.179
(0.187)
0.116
(0.266)
-0.005***
(0.002)
0.222
(0.183)
x
9,572
464
1978-99
x
3,999
210
1978-99
x
5,573
254
1978-99
-0.000
(0.001)
-0.009***
(0.001)
-0.001
(0.003)
-0.000
(0.001)
-0.011***
(0.001)
x
35,018
1,822
1978-99
x
15,549
934
1978-99
x
19,469
888
1978-99
-
Some land
(6)
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land registered is
computed as the sum over the previous three years of the share of land affected by each program over the total cultivable
land in each village, using official land records. All regressions also include a land ceiling indicator, immigrant status
indicators, and year and household fixed effects, and % land distributed calculated the same way as % land registered. The
land category of the household is defined based on year 1977 landholdings, and regressions include data for years 1978–
1999. Religion of the household is the religion of the household head. *** p<0.01, ** p<0.05, * p<0.1
42
Table 11: Surviving Hindu Male Births by Immigrant Status
1=surviving boy,
0=no birth or surviving girl
Dep. Variable:
Native/Immigrant:
Land category:
Agricultural land
% Land registered
Household FE
Observations
Households
Years
All
(1)
Native
Landless
(2)
-
Immigrant
Landless Some land
(5)
(6)
Some land
(3)
All
(4)
-0.012
(0.010)
0.039***
(0.005)
0.038***
(0.003)
-0.005
(0.009)
0.564
(0.481)
x
5,919
482
1978-99
x
5,078
443
1978-99
x
841
39
1978-99
0.001
(0.001)
0.008***
(0.002)
0.011
(0.013)
0.001
(0.001)
0.006***
(0.001)
x
29,099
1,340
1978-99
x
10,471
491
1978-99
x
18,628
849
1978-99
-
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land
registered is computed as the sum over the previous three years of the share of land affected by each
program over the total cultivable land in each village, using official land records. All regressions also include a
land ceiling indicator, year of immigration controls, and year and household fixed effects, and % land
distributed calculated the same way as % land registered. The land category of the household is defined
based on year 1977 landholdings, and regressions include data for years 1978–1999. Religion of the
household is the religion of the household head. *** p<0.01, ** p<0.05, * p<0.1
Table 12: Surviving Hindu Female Births by Immigrant Status
1=surviving girl,
0=no birth or surviving boy
Dep. Variable:
Native/Immigrant:
Land category:
Agricultural land
% Land registered
Household FE
Observations
Households
Years
All
(1)
Native
Landless
(2)
-
Immigrant
Landless Some land
(5)
(6)
Some land
(3)
All
(4)
-0.014
(0.009)
0.001
(0.003)
-0.000
(0.003)
-0.004
(0.008)
-0.103
(0.465)
x
5,919
482
1978-99
x
5,078
443
1978-99
x
841
39
1978-99
0.000
(0.001)
-0.010***
(0.001)
-0.003
(0.006)
-0.000
(0.001)
-0.011***
(0.001)
x
29,099
1,340
1978-99
x
10,471
491
1978-99
x
18,628
849
1978-99
-
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land
registered is computed as the sum over the previous three years of the share of land affected by each
program over the total cultivable land in each village, using official land records. All regressions also include a
land ceiling indicator, year of immigration controls, and year and household fixed effects, and % land
distributed calculated the same way as % land registered. The land category of the household is defined
based on year 1977 landholdings, and regressions include data for years 1978–1999. Religion of the
household is the religion of the household head. *** p<0.01, ** p<0.05, * p<0.1
43
Appendix A: Additional Tables
Table A.1: Infant Death, Younger Siblings, and Sharecropper Registration: Birth Order 1
All Children
Infant Death
Younger Sibling
(1)
(2)
SCROP50 t-1 * FEMALE
SCROP25 t-1 * FEMALE
SCROP50 t-1
SCROP25 t-1
FEMALE
District FE
District Covariates
District-Birth Year Trend
Observations
Cohorts
Districts
Hindu Children
Infant Death
Younger Sibling
(3)
(4)
Non-Hindu Children
Infant Death Younger Sibling
(5)
(6)
0.014
(0.025)
-0.041
(0.031)
0.000
(0.025)
0.016
(0.048)
-0.044
(0.12)
0.007
(0.021)
-0.005
(0.017)
-0.015
(0.030)
0.008
(0.037)
-0.245
(0.181)
0.011
(0.026)
-0.029
(0.037)
0.017
(0.028)
0.053
(0.053)
-0.117
(0.17)
0.026
(0.032)
-0.003
(0.026)
-0.039
(0.034)
0.025
(0.052)
-0.286
(0.268)
0.025
(0.042)
-0.052
(0.062)
-0.033
(0.043)
-0.057
(0.055)
0.169
(0.288)
-0.025
(0.044)
-0.022
(0.039)
0.038
(0.04)
-0.032
(0.027)
-0.093
(0.156)
x
x
x
3,248
1978-91
14
x
x
x
3,248
1978-91
14
x
x
x
2,323
1978-91
14
x
x
x
2,323
1978-91
14
x
x
x
925
1978-91
14
x
x
x
925
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. Specifications also include birth year fixed effects, year of interview fixed effects, indicators for
household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the
child is born. The district covariates include logs of for rice and cereal productivity, patta land area distributed, number of medical institutions, and kilometres of
surfaced road per capita and their corresponding interactions with the female child and the firstborn son indicators. *** p<0.01, ** p<0.05, * p<0.1
44
Table A.2: Infant Mortality, Firstborn Sons, and Registration: Bihar Control Districts
(1)
SCROP50 t-1 * FIRSTSON * FEMALE
SCROP25 t-1 * FIRSTSON * FEMALE
SCROP50 t-1 * FEMALE
SCROP25 t-1 * FEMALE
SCROP50 t-1 * FIRSTSON
SCROP25 t-1 * FIRSTSON
SCROP50 t-1
SCROP25 t-1
FIRSTSON * FEMALE
FEMALE
FIRSTSON
District FE
District Productivity Controls
District-Birth Year Trend
Observations
Cohorts
Districts
All Children
(2)
-0.058*
(0.029)
0.020
(0.030)
0.051**
(0.024)
-0.006
(0.023)
0.027
(0.019)
-0.024
(0.024)
-0.043**
(0.021)
0.015
(0.02)
-0.003
(0.029)
-0.023
(0.025)
0.014
(0.023)
x
-0.067*
(0.033)
0.007
(0.035)
0.051**
(0.025)
-0.009
(0.028)
0.031
(0.023)
-0.019
(0.025)
-0.042**
(0.021)
0.016
(0.021)
-0.004
(0.028)
-0.023
(0.023)
0.014
(0.024)
x
x
9,355
1978-91
19
9,355
1978-91
19
Infant Death
Hindu Children
(5)
(3)
(4)
-0.067*
(0.033)
0.008
(0.036)
0.051**
(0.023)
-0.010
(0.029)
0.031
(0.022)
-0.017
(0.026)
-0.061***
(0.023)
-0.009
(0.022)
-0.003
(0.029)
-0.023
(0.024)
0.012
(0.023)
x
x
x
9,355
1978-91
19
-0.077**
(0.033)
0.010
(0.039)
0.066**
(0.029)
-0.002
(0.026)
0.020
(0.024)
-0.029
(0.034)
-0.035
(0.022)
-0.001
(0.027)
0.019
(0.033)
-0.048*
(0.026)
0.010
(0.03)
x
-0.090**
(0.040)
-0.009
(0.046)
0.071**
(0.031)
0.000
(0.029)
0.030
(0.028)
-0.012
(0.033)
-0.040*
(0.022)
-0.004
(0.025)
0.016
(0.033)
-0.047*
(0.027)
0.013
(0.031)
x
x
6,236
1978-91
19
6,236
1978-91
19
Non-Hindu Children
(8)
(9)
(6)
(7)
-0.089**
(0.038)
-0.007
(0.046)
0.071**
(0.031)
-0.001
(0.029)
0.030
(0.028)
-0.013
(0.033)
-0.054**
(0.025)
-0.016
(0.03)
0.016
(0.033)
-0.043
(0.026)
0.012
(0.031)
x
x
x
6,236
1978-91
19
-0.023
(0.045)
0.044
(0.042)
0.024
(0.032)
-0.030
(0.033)
0.041
(0.025)
-0.005
(0.028)
-0.068**
(0.036)
0.051
(0.033)
-0.052
(0.034)
0.042
(0.038)
0.006
(0.026)
x
-0.027
(0.048)
0.048
(0.05)
0.017
(0.035)
-0.044
(0.046)
0.035
(0.03)
-0.020
(0.034)
-0.055
(0.034)
0.051
(0.034)
-0.049
(0.034)
0.037
(0.035)
0.002
(0.027)
x
x
3,119
1978-91
19
3,119
1978-91
19
-0.030
(0.048)
0.040
(0.05)
0.017
(0.036)
-0.041
(0.047)
0.035
(0.03)
-0.010
(0.032)
-0.082**
(0.038)
-0.003
(0.037)
-0.044
(0.035)
0.030
(0.038)
-0.006
(0.029)
x
x
x
3,119
1978-91
19
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also include the female child, birth year fixed effects, birth order fixed effects, year of interview fixed
effects, indicators for household religion and caste, whether the household is rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child
is born. Specifications for children of birth order 2 or higher also include the firstborn son indicator and its corresponding interaction terms. *** p<0.01, ** p<0.05, * p<0.1
45
Table A.3: Younger Siblings and Registration: Bihar Control Districts
SCROP50 t-1 * FIRSTSON
SCROP25 t-1 * FIRSTSON
SCROP50 t-1
SCROP25 t-1
FIRSTSON
District FE
District Productivity
Controls
District-Birth Year Trend
Observations
Cohorts
Districts
(1)
All Children
(2)
(3)
0.001
(0.044)
-0.094***
(0.036)
-0.036
(0.027)
0.027
(0.032)
-0.008
(0.022)
x
-0.001
(0.044)
-0.096**
(0.044)
-0.034
(0.026)
0.015
(0.038)
-0.008
(0.021)
x
-0.001
(0.043)
-0.099**
(0.045)
-0.040
(0.031)
-0.019
(0.053)
-0.003
(0.02)
x
x
x
2,935
1978-91
14
x
2,935
1978-91
14
2,935
1978-91
14
Child Has a Younger Sibling
Hindu Children
(4)
(5)
(6)
0.001
(0.049)
-0.085**
(0.037)
-0.018
(0.032)
0.032
(0.045)
-0.029
(0.026)
x
2,126
1978-91
14
0.000
(0.052)
-0.083*
(0.043)
-0.017
(0.033)
0.019
(0.052)
-0.029
(0.025)
x
0.004
(0.051)
-0.091**
(0.044)
-0.033
(0.039)
-0.048
(0.076)
-0.023
(0.022)
x
x
x
2,126
1978-91
14
x
2,126
1978-91
14
(7)
Non-Hindu Children
(8)
(9)
0.004
(0.053)
-0.158**
(0.065)
-0.085*
(0.049)
0.011
(0.042)
0.081**
(0.039)
x
-0.010
(0.046)
-0.183**
(0.077)
-0.078*
(0.044)
0.003
(0.037)
0.076*
(0.042)
x
x
-0.015
(0.046)
-0.176**
(0.078)
-0.070
(0.043)
0.044
(0.065)
0.065
(0.047)
x
x
809
1978-91
14
809
1978-91
14
x
809
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. The sample in each specification is children of birth order 2. All specifications also include the female
child, birth year fixed effects, year of interview fixed effects, indicators for household religion and caste, whether the household is rural, mother’s educational attainment,
and linear and quadratic terms of the mother’s age at which the child is born. *** p<0.01, ** p<0.05, * p<0.1
46
Table A.4: Infant Mortality, Firstborn Sons, and Registration – Rice Productivity Estimates
All Children
B. Or. 1
B. Or. >1
(1)
(2)
SCROP50 t-1 * FIRSTSON * FEMALE
-
SCROP25 t-1 * FIRSTSON * FEMALE
-
SCROP50 t-1 * FEMALE
SCROP50 t-1 * FIRSTSON
0.014
(0.025)
-0.041
(0.031)
-
SCROP25 t-1 * FIRSTSON
-
SCROP25 t-1 * FEMALE
SCROP50 t-1
SCROP25 t-1
LN RICEPR t-1 * FIRSTSON * FEMALE
LN RICEPR t-1 * FEMALE
LN RICEPR t-1 * FIRSTSON
LN RICEPR t-1
FIRSTSON * FEMALE
FEMALE
FIRSTSON
District FE
District Covariates
District-Birth Year Trend
Observations
Cohorts
Districts
0.000
(0.025)
0.016
(0.048)
-0.058
(0.042)
0.085*
(0.046)
-0.044
(0.12)
Yes
x
x
3,248
1978-91
14
-0.073*
(0.039)
0.031
(0.038)
0.046**
(0.023)
-0.029
(0.032)
0.039
(0.028)
-0.034
(0.023)
-0.063***
(0.023)
0.001
(0.020)
0.043
(0.049)
0.035
(0.044)
-0.041
(0.048)
-0.102**
(0.046)
-0.025
(0.141)
0.163
(0.131)
0.034
(0.137)
Yes
x
x
8,367
1978-91
14
Infant Death
Hindu Children
B. Or. 1 B. Or. >1
(3)
(4)
0.011
(0.026)
-0.029
(0.037)
0.017
(0.028)
0.053
(0.053)
-0.020
(0.049)
0.062
(0.053)
-0.117
(0.170)
Yes
x
x
2,323
1978-91
14
-0.097**
(0.048)
0.019
(0.051)
0.070**
(0.029)
-0.027
(0.033)
0.043
(0.032)
-0.035
(0.036)
-0.059**
(0.026)
-0.003
(0.031)
0.087
(0.068)
-0.009
(0.053)
-0.109
(0.065)
0.006
(0.048)
-0.096
(0.177)
0.171
(0.131)
0.061
(0.197)
Yes
x
x
5,448
1978-91
14
Non-Hindu Children
B. Or. 1
B. Or. >1
(5)
(6)
0.025
(0.042)
-0.052
(0.062)
-0.033
(0.043)
-0.057
(0.055)
-0.170*
(0.093)
0.206**
(0.09)
0.169
(0.288)
Yes
x
x
925
1978-91
14
-0.028
(0.055)
0.046
(0.044)
-0.001
(0.046)
-0.036
(0.048)
0.029
(0.031)
-0.016
(0.03)
-0.078**
(0.035)
-0.005
(0.039)
0.087
(0.068)
-0.009
(0.053)
-0.109
(0.065)
-0.308***
(0.121)
0.085
(0.219)
0.174
(0.247)
-0.017
(0.150)
Yes
x
x
2,919
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also include birth year fixed effects, birth
order fixed effects, year of interview fixed effects, indicators for household religion and caste, whether the household is
rural, mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child is born. The
district covariates include logs of for rice and cereal productivity, patta land area distributed, number of medical institutions,
and kilometres of surfaced road per capita and their corresponding interactions with the female child and the firstborn son
indicators. *** p<0.01, ** p<0.05, * p<0.1
47
Table A.5: Younger Siblings and Sharecropper Registration – Rice Productivity Estimates
All Children
B. Ord. 1
B. Ord. 2
(1)
(2)
Child has a Younger Sibling
Hindu Children
B. Ord. 1
B. Ord. 2
(4)
(5)
Non-Hindu Children
B. Ord. 1
B. Ord. 2
(7)
(8)
SCROP50 t-1 * FIRSTSON
-
0.012
(0.044)
-
0.008
(0.051)
-
0.019
(0.05)
SCROP25 t-1 * FIRSTSON
-
-0.116**
(0.055)
-
-0.108**
(0.052)
-
-0.182**
(0.086)
SCROP50 t-1 * FEMALE
0.007
(0.021)
-
0.026
(0.032)
-
-0.025
(0.044)
-
SCROP25 t-1 * FEMALE
-0.005
(0.017)
-
-0.003
(0.026)
-
-0.022
(0.039)
-
SCROP50 t-1
-0.015
(0.030)
-0.048
(0.032)
-0.039
(0.034)
-0.038
(0.039)
0.038
(0.04)
-0.096*
(0.051)
SCROP25 t-1
0.008
(0.037)
-0.020
(0.056)
0.025
(0.052)
-0.058
(0.079)
-0.032
(0.027)
0.036
(0.083)
LN RICEPR t-1 * FIRSTSON
-
-0.086
(0.072)
-
-0.083
(0.085)
-
-0.050
(0.052)
LN RICEPR t-1
-
-0.061
(0.075)
-
-0.047
(0.09)
-
-0.093
(0.086)
LN RICEPR t-1 * FEMALE
0.026
(0.038)
-
0.036
(0.045)
-
0.001
(0.067)
-
LN RICEPR t-1
-0.030
(0.066)
-
-0.051
(0.081)
-
0.090
(0.103)
-
FEMALE
-0.245
(0.181)
-
-0.286
(0.268)
-
-0.093
(0.156)
-
FIRSTSON
-
-0.201
(0.193)
-
-0.347
(0.251)
-
0.205
(0.138)
Yes
x
x
3,248
1978-91
14
Yes
x
x
2,686
1978-91
14
Yes
x
x
2,323
1978-91
14
Yes
x
x
1,919
1978-91
14
Yes
x
x
925
1978-91
14
Yes
x
x
767
1978-91
14
District FE
District Covariates
District-Birth Year Trend
Observations
Cohorts
Districts
Notes: Wild cluster bootstrapped standard errors in parentheses. Specifications also include the female child and
firstborn son indicators for children of birth order 1 and 2 respectively, birth year fixed effects, year of interview fixed
effects, indicators for household religion and caste, whether the household is rural, mother’s educational attainment,
and linear and quadratic terms of the mother’s age at which the child is born. The district covariates include logs of
patta land area distributed, number of medical institutions, and kilometres of surfaced road per capita and their
corresponding interactions with the female child and the firstborn son indicators. *** p<0.01, ** p<0.05, * p<0.1
48
Table A.6: Male Births and Registration – Rice Productivity Estimates
All Children
B. Or. 1 B. Or. >1
(1)
(2)
SCROP50 t-1 * FIRSTSON
-
SCROP25 t-1 * FIRSTSON
-
SCROP50 t-1
SCROP25 t-1
LN RICEPR t-1 * FIRSTSON
LN RICEPR t-1
-0.020
(0.037)
-0.010
(0.034)
-
FIRSTSON
-0.064
(0.098)
-
District FE
District Covariates
District-Birth Year Trend
Observations
Cohorts
Districts
x
x
x
3,248
1978-91
14
-0.011
(0.020)
-0.005
(0.027)
0.045*
(0.025)
0.051
(0.058)
-0.014
(0.029)
-0.009
(0.051)
0.094
(0.096)
x
x
x
8,367
1978-91
14
Child is Male
Hindu Children
B. Or. 1
B. Or. >1
(3)
(4)
-0.065
(0.050)
0.063
(0.053)
-0.010
(0.131)
x
x
x
2,323
1978-91
14
-0.014
(0.019)
0.005
(0.031)
0.055**
(0.029)
0.030
(0.054)
-0.034
(0.039)
0.023
(0.050)
0.186
(0.113)
x
x
x
5,448
1978-91
14
Non-Hindu Children
B. Or. 1
B. Or. >1
(5)
(6)
0.091
(0.064)
-0.149
(0.098)
-0.241*
(0.136)
x
x
x
925
1978-91
14
0.004
(0.047)
-0.033
(0.062)
0.039
(0.041)
0.111
(0.095)
-0.010
(0.084)
-0.062
(0.098)
-0.005
(0.262)
x
x
x
2,919
1978-91
14
Notes: Wild cluster bootstrapped standard errors in parentheses. All specifications also include birth year fixed effects, birth
order fixed effects, year of interview fixed effects, indicators for household religion and caste, whether the household is rural,
mother’s educational attainment, and linear and quadratic terms of the mother’s age at which the child is born. Specifications for
children of birth order 2 or higher also include the firstborn son indicator. The district covariates include logs of for rice and
cereal productivity, patta land area distributed, number of medical institutions, and kilometres of surfaced road per capita and
their corresponding interactions with the female child and the firstborn son indicators. *** p<0.01, ** p<0.05, * p<0.1
49
Table A.7: Land Reform and Surviving Births – District-Year Fixed Effects
1=surviving girl,
0=no birth or surviving boy
1=surviving boy,
0=no birth or surviving girl
Land category:
All
(1)
Landless
(2)
Some land
(3)
All
(4)
Landless
(5)
Some land
(6)
Agricultural land
-0.001
(0.001)
-0.007***
(0.001)
-
-0.000
(0.001)
0.009***
(0.003)
-
0.000
(0.003)
-0.001
(0.001)
-0.009***
(0.001)
0.020***
(0.007)
-0.000
(0.001)
0.008***
(0.001)
x
x
42,304
2,268
1978-99
x
x
18,404
1,126
1978-99
x
x
23,900
1,142
1978-99
x
x
42,304
2,268
1978-99
x
x
18,404
1,126
1978-99
x
x
23,900
1,142
1978-99
Dep. Variable:
% Land registered
Household FE
District-Year FE
Observations
Households
Years
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land registered is
computed as the sum over the previous three years of the share of land affected by each program over the total
cultivable land in each village, using official land records. All regressions also include a land ceiling indicator, immigrant
status indicators, and year and household fixed effects, and % land distributed calculated the same way as % land
registered. The land category of the household is defined based on year 1977 landholdings, and regressions include
data for years 1978–1999. *** p<0.01, ** p<0.05, * p<0.1.
Table A.8: Land Reform and Surviving Births – Full Sample
Dep. Variable:
1=surviving girl,
0=no birth or surviving boy
1=surviving boy,
0=no birth or surviving girl
Land category:
All
(1)
Landless
(2)
Some land
(3)
All
(4)
Landless
(5)
Some land
(6)
Agricultural land
-0.001
(0.000)
-0.009***
(0.001)
-
-0.000
(0.001)
0.009***
(0.002)
-
-0.003
(0.006)
-0.001
(0.000)
-0.010***
(0.001)
0.042***
(0.011)
0.000
(0.001)
0.005***
(0.001)
x
56,461
2,286
1978-99
x
23,154
1,230
1978-99
x
33,307
1,403
1978-99
x
56,461
2,286
1978-99
x
23,154
1,230
1978-99
x
33,307
1,403
1978-99
% Land registered
Household FE
Observations
Households
Years
Notes: Robust standard errors in parentheses, adjusted for clustering on villages. The variable % land registered is
computed as the sum over the previous three years of the share of land affected by each program over the total
cultivable land in each village, using official land records. All regressions also include a land ceiling indicator,
immigrant status indicators, and year and household fixed effects, and % land distributed calculated the same way as
% land registered. The land category of the household is defined based on year 1977 landholdings, and regressions
include data for years 1978–1999. *** p<0.01, ** p<0.05, * p<0.1.
50