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OCR for page 39
Effect of a Child's Death on Birth Spacing:
A Cross-National Analysis
Laurence M. Grummer-Strawn, Paul W. Stupp, and
Zuguo Mei
INTRODUCTION
Our purpose in this chapter is to examine the potential pathways through
which infant and child mortality can affect the interval between births. Specifi-
cally, we examine the extent to which the death of a child tends to shorten the
time until the birth of the next child. To the extent that the shortening of the birth
interval is attributable to direct physiological mechanisms (i.e., premature return
of menses due to breastfeeding cessation), the linkage between mortality decline
and fertility decline can be considered directly causal. On the other hand, if the
linkage between child death and birth interval is related more to maternal behav-
ior, such as cessation of contraceptive use, then the linkage depends on the
widespread use of contraceptives. By teasing out the mechanisms for the linkage,
we will better understand the future course of fertility and mortality in developing
countries.
The analysis uses data from phases 1 and 2 of the Demographic and Health
Surveys (DHS), which collect an extensive set of fertility-related variables for
women of reproductive age and for live births that occurred in the 5 years before
the date of interview. We first present evidence of the magnitude of the effects on
the birth interval for a variety of countries and then address the mechanisms
through which these effects may operate.
PATHWAYS OF INFLUENCE
The effects of infant and child mortality on the intervals between births
39
OCR for page 40
40
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
probably act through both physiological mechanisms and volitional behaviors as
noted by Preston (1978~. Most prominent among the physiological mechanisms
is lactational amenorrhea, by which breastfeeding inhibits the return of ovulation
after a birth. If a child dies, the mother will stop breastfeeding (except perhaps in
the case of a multiple birth), and she will become susceptible to pregnancy more
quickly than if the child had survived. This effect is expected to be most pro-
nounced in societies such as those in sub-Saharan Africa in which breastfeeding
is nearly universal and is performed for an extended time. Studies in the early
1980s showed that the anovulatory effect of breastfeeding is related to its fre-
quency, intensity, and duration (Konner and Worthman, 1980; McNeilly et al.,
1980; Howie et al., 1982~. For this analysis, duration of breastfeeding was the
only variable for which data were collected for all births during a 5-year period.
To explore the role of lactational amenorrhea in spacing of births, we used
data from the DHSs to examine the effect of death of a child on the length of time
until return of menses and until the next live birth. We treat time to return of
menses, which is among the data collected in the DHS, as a proxy for time to
return of ovulation, which is not collected in the DHS.
Beyond its effect in delaying the return of ovulation, breastfeeding may also
reduce the probability of conception among ovulating, sexually active women.
We investigated the effect of the premature truncation of breastfeeding on the
interval from return of menses and resumption of sexual relations to a subsequent
pregnancy.
The DHS do not collect information on pregnancies that do not result in a
live birth (i.e., spontaneous abortions and stillbirths). This restriction prevents us
from considering another potential physiological mechanism by which the death
of a child could lengthen the interval to the subsequent birth. If infant mortality
is correlated with fetal mortality in the same woman it could lengthen the subse-
quent interval due to fetal losses. This effect related to maternal health or genetic
endowment appears realistic, especially in the case of neonatal mortality, and
could be substantial for a subset of women. Unfortunately, we cannot use DHS
data to measure this effect.
The primary behavioral pathways through which death of a child can affect
the interval before a subsequent pregnancy involve sexual activity and contracep-
tive use. If a society observes a norm of traditional sexual abstinence for a period
after a birth, then the death of a child may reduce that period. The DHS contain
data on duration of postpartum abstinence and thus make it possible to investigate
directly whether the death of a child affects this component of the birth interval.
Once a woman has resumed sexual relations and is ovulating after a birth, the
behavioral factors affecting the time to conception are frequency of sexual inter-
course and contraceptive use. We do not have good measures of either of these
variables during the pertinent periods for all of the surveys used in this analysis.
We presume, however, that the shortening of this portion of the interval, once we
control for breastfeeding status, must be related to one or both of these factors.
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LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
41
We predict that the effect of the death of a child during this period of susceptibil-
ity to pregnancy will be more pronounced in countries with more frequent contra-
ceptive use. Unfortunately, the DHS do not generally provide appropriate data
on changes in contraceptive use throughout the intervals of interest, so it was not
possible to examine the effect of contraceptive use. Use of a contraceptive
calendar would be appropriate for this purpose, but only in high prevalence
countries DHS included a contraceptive use calendar.
Frequency of sexual relations is another factor that may be related to the
effect of death of an infant on the length of the susceptible period. A couple
could respond to a child's death by increasing the frequency of sexual activity in
an effort to replace a dead child. Unfortunately, this hypothesis could not be
tested because DHS do not provide data on coital frequency during each period of
the interval.
Induced abortion is another factor we could not address. This is probably an
important factor only in very low fertility settings where child mortality is more
rare. Such countries have generally been excluded from the DHS program.
INTERVENING VARIABLES
It is well documented that the length of the birth interval affects child mortal-
ity and that shorter previous intervals are associated with higher infant and child
mortality rates (Sullivan et al., 1994~. Some of the association between a child's
death and a short subsequent birth interval could therefore be confounded by the
association between short previous intervals and a subsequent child death. If
some women are more prone to shorter intervals than average, because of differ-
entials in fecundity, sexual activity, or contraceptive use, then child mortality
may be a spurious factor. It is therefore important that we control for the length
of the previous interval in studying the association between child mortality and
the length of subsequent intervals.
Similarly, child mortality may be higher if there is a shorter subsequent
interval. For a short subsequent interval to affect a child's death, the death should
take place after the conception of the child who is born subsequently. Similarly,
for a death to affect the length of the subsequent interval it should logically occur
before the conception leading to the birth that closes the interval. To avoid the
possibility of reverse causality, we took great care in the modeling to limit the
effect of the death to the period before the conception leading to the subsequent
live birth.
Other factors that may simultaneously affect both child survival and birth
interval are maternal age and education, birth order, sex of the child, whether the
child was wanted, socioeconomic status of the household, and whether the woman
resides in an urban or rural community. Among these factors, sex of the child,
whether the child was wanted, and maternal age tend to mask the association
between death and short birth intervals. If the child is female, she is less likely to
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42
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
die and the birth interval is likely to be shorter because of the desire to have
another birth following the birth of a girl (assuming preference for a son). Simi-
larly, infants who are unwanted experience higher mortality, and subsequent
birth intervals are likely to be longer. Children born to older mothers are at
greater risk of death, but their birth is often followed by longer birth intervals
because of the mother's reduced fecundity. Other factors may tend to generate
the expected association between death of a child and length of the birth interval.
If maternal education or socioeconomic status is higher, if birth order is lower, or
place of residence is urban, infant mortality is likely to be lower and birth interval
longer. Data on these variables are available in the DHS and are thus included in
the multivariate analyses.
This list of intervening factors is by no means comprehensive. Other charac-
teristics of the mother or child might be correlated with both interval length and
the survival of the child. For example, mothers with a strong desire to have
healthy children may opt for longer birth intervals (knowing that they are safer
for the child) but also invest more resources in nutrition and medical care for the
child. The continued presence of a male partner in the household certainly affects
the probability of closing the birth interval, but also influences socioeconomic
well-being of the household and allocation of resources within the household.
Other characteristics that affect birth intervals but that are not necessarily corre-
lated with child mortality, such as the woman's underlying fecundability, can
bias estimation of the parameter estimates as well (Heckman and Singer, 1982,
1984~. This problem of unobserved heterogeneity has been addressed by a vari-
ety of complex statistical procedures (Trussell and Richards, 1985; Trussell and
Rodriguez, 1990; Guo and Rodriguez, 1992~. Unfortunately, we were unable to
find a suitable method that accounts for left- and right-censoring, time-varying
covariates and time-varying effects in a manner that could be applied easily to 45
surveys. Our hope is that the variables controlled in the analysis are reasonable
proxies for the unobserved characteristics, such that the degree of bias is mini-
mized.
DATA AND METHODS
For this analysis, we used data from the DHS phases 1 and 2 (Lapham and
Westoff, 1986~. All data sets available at the time this study was started were
included. Recoded data sets for 46 surveys were provided by Macro Interna-
tional, Inc. (Calverton, Maryland). Conducted in developing countries in the late
1980s and early 1990s, DHS were based on household interviews of women of
reproductive age (age 15-49~. Standard core questionnaires were used, encom-
passing topics on family planning, fertility, child mortality, and maternal and
child health. Additional questions and slight modifications were implemented in
each country. Results are comparable across countries because the DHS use
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LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
43
standard questionnaire modules and standardized sampling procedures, field-
work, and data processing.
The core questionnaire included data on the timing of a woman's live births,
the survival status for these births, and the interval to death of a child, where
applicable. For births occurring in the 5 years before the survey, data were
obtained on the duration of breastfeeding, amenorrhea, and sexual abstinence.]
Our analysis is restricted to births in the previous 5 years because a major aim is
to examine the mechanisms through which the death of a child affects the subse-
quent birth interval, specifically the premature truncation of breastfeeding lead-
ing to earlier return of menses and resumption of sexual activity. Countries
included in the analysis are listed in Table 2-1 and the number of births for each
country is shown.
Outcome Measures
Interval Between Births
We examined the relationship between the birth interval and the death of the
child that started the interval. Several problems are apparent. First, many of the
birth intervals (a majority in many countries) are open intervals, that is, last
births. For these births, the time to the next birth is censored; thus, we had to use
life table techniques in the data analyses. Second, because a short subsequent
birth interval can contribute to the premature death of the index child (Hobcraft et
al., 1985), we must account for the possibility of reverse causality by ensuring
that the death occurs before the conception of the child whose birth closes the
birth interval. Thus, death must be accounted for as a time-varying variable.
To handle these problems, we have chosen to model birth intervals by using
proportional hazards models; the baseline hazards are modeled with a piecewise
exponential (Trussell and Guinnane, 1993~. This model is specified as
\(tlX)=
X0 (t) exp (X 3),
1Data on the duration of breastfeeding, postpartum amenorrhea, and postpartum abstinence are
based on the mother's retrospective reports. It is well known that the quality of retrospectively
reported data on duration is rather poor. Durations are frequently reported as multiples of 3 or 6
months and thus do not reflect accurate recollection of the events. To test whether this rounding or
"heaping" problem affected our results, we tested models in which we excluded women who re-
ported amenorrhea or breastfeeding durations of 3, 6, 9, 12, 15, 18, 24, 30, or 36 months. In all
cases, the differences were so small that the results described here are not substantively changed.
Data quality could be related to the main variable of interest here (child death). Interviewers may
be reluctant to ask about dead children because of cultural norms. Mothers may inaccurately report
breastfeeding the dead child out of a feeling of guilt.
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44
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
TABLE 2-1 Sample Sizes and Years of Surveys Included in Analysis
Region Number of Births
and Country Survey Year in Previous 5 Years
Africa
Botswana DHS1 1988 3,086
Burkina Faso DHS2 1992 5,828
Burundi DHS1 1987 3,811
Cameroon DHS2 1991 3,350
Egypt DHS1 1988 8,647
Ghana DHS1 1988 4,136
Kenya DHS1 1988-1989 6,980
Liberia DHS 1 1986 5,299
Madagascar DHS2 1992 5,273
Malawi DHS2 1992 4,495
Mali DHS1 1987 3,358
Morocco DHS1 1987 6,102
Morocco DHS2 1992 5,197
Namibia DHS2 1992 3,916
Niger DHS2 1992 6,899
Nigeria DHS2 1990 7,902
Ondo State, Nigeria DHS1 1986-1987 3,280
Rwanda DHS2 1992 5,510
Senegal DHS 1 1986 4,287
Senegal DHS2 1992- 1993 5,645
Sudan DHS1 1989-1990 6,644
Togo DHS1 1988 3,134
Tunisia DHS1 1988 4,477
Uganda DHS1 1988-1989 4,959
Zambia DHS2 1992 6,299
Zimbabwe DHS1 1988-1989 3,358
Asia
Indonesia DHS1 1987 8,140
Indonesia DHS2 1991 15,708
Pakistan DHS2 1990- 1991 6,428
Sri Lanka DHS1 1987 4,010
Thailand DHS 1 1987 3,627
Latin America
Bolivia DHS1 1989 5,814
Brazil DHS1 1986 3,573
Brazil DHS2 1991 3,159
Colombia DHS 1 1986 2,715
Colombia DHS2 1990 3,751
Dominican Republic DHS1 1986 4,767
Dominican Republic DHS2 1991 4,164
Ecuador DHS 1 1987 3,051
E1 Salvador DHS1 1985 3,339
Guatemala DHS2 1987 4,627
Mexico DHS1 1987 5,327
Paraguay DHS2 1990 4,246
Peru DHS1 1986 3,131
Peru DHS2 1991-1992 9,362
Trinidad and Tobago DHS1 1987 1,946
OCR for page 45
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
45
where ~ is the hazard of the birth interval being closed by a subsequent live birth
at time t, X0 is the baseline hazard, and X is a set of covariates. The baseline
hazard is a step function, with steps defined for age groups 9-17, 18-23, 24-29,
30-35, 36-41, and 42-59 months. The hazard of closing the birth interval is
assumed to be zero in the first 9 months of a child's life, allowing for a 9-month
gestation for a subsequent birth. Therefore, all exposure and events in these first
9 months are ignored. We excluded intervals in which the date of either the index
birth or the subsequent birth was imputed and intervals that began with a multiple
birth.
We began with model 1 in which the only covariate is whether the index
child was alive 10 months ago. In this way, the risk of a conception occurring
after the index child had died was compared with the risk of conception while the
child was living. Whether the index child is dead was treated as a time-varying
covariate, with a lag of 10 months (assuming a gestation of 9 months). Thus, for
example, death of the index child at the age of 6 months can only affect the birth
of the next child at durations of 16 months and greater.
In model 2 of the birth interval we introduced controls for place of residence
(urban or rural), an index of the socioeconomic status of the household, the
mother's highest level of completed education (none, primary, or secondary), the
mother's age at the time of the index birth, the sex of the child, parity, length of
the previous birth interval, and whether the child was wanted at that time, at a
later time, or not at all. The index of socioeconomic status was a continuous
covariate based on a count of the number of goods and services available in the
household from among the following: automobile, motorcycle, bicycle, refrig-
erator, television, radio, and electricity. The mother's age was modeled with a
linear and quadratic term to account for nonlinearity in the declining fecundity
with age. We considered parity and previous birth interval jointly by using a
categorical variable with categories of first-births, parity of two to five with an
interval less than 24 months, parity of two to five with an interval of at least 24
months, parity of six or more with an interval less than 24 months, and parity of
six or more with an interval of at least 24 months. As noted, some of these factors
tend to mask the association between death and short birth intervals, but others
tend to generate such an association. Therefore, we had no prior expectation as to
whether the effect of the index child's death would be stronger or weaker after
adjustment for these factors.
In model 3, we added the effect of breastfeeding on closure of the birth
interval. Breastfeeding was modeled in the same way as was the index child's
death, that is, as a time-varying covariate with a 10-month lag. Although a
woman can continue breastfeeding during pregnancy, we focused only on breast-
feeding before the time of conception to avoid any possibility of reverse causality
that occurs if a woman stops breastfeeding because she is pregnant. The addition
of a breastfeeding dummy variable to the model along with child death had the
effect of creating three possible statuses of the index child 10 months earlier:
OCR for page 46
46
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
dead, weaned but alive, and still breastfed. The parameter on the variable for
death of the index child thus measured the effect of that event relative to the
effect of survival of the index child who was weaned. Because we expected that
a major mechanism for the association between death of the index child and a
shorter subsequent birth interval was the premature truncation of breastfeeding,
we anticipated a large reduction in the effect of a child's death. In fact, if
breastfeeding were the only mechanism for the prolongation of the birth interval,
as might be the case in countries where contraceptives are seldom used, then we
would expect that the effect of child death might be reduced to zero.
Interval from Birth to Menses
To better understand the mechanisms through which the death of a child
might affect the subsequent birth interval, it is useful to consider the various
components of the birth interval. Two events must take place before the concep-
tion of the subsequent birth. The couple must resume sexual relations and the
woman must begin ovulating again. Either event could occur first. Thus, the
birth interval can be thought of in three distinct parts: insusceptible period
(amenorrheic or abstinent), susceptible period, and gestation. Because data on
gestational age are not available in the DHS, the part of the interval related to
gestation was not examined separately.
Components of the Birth Interval
Return of Menses
1 1 1 1 1
Index Sexual
Birth Relations
Fertile Conception Next
Ovulation Birth
The timing of ovulation is not directly observable; thus, no data are available
on ovulation. However, data on the timing of first menses after each birth were
obtained from the DHS. Menses can occur several months before ovulation, with
the first few cycles being anovulatory, or ovulation can precede menses. Never-
theless, the return of menses is considered to be a good proxy for ovulation. For
this reason, we modeled the effect of a child's death on the resumption of menses.
A series of models was estimated, as were the full models for the birth interval,
by adding control variables and breastfeeding in turn. The piecewise hazards
were modeled with steps placed at 0-2, 3-5, 6-8, 9-11, 12-14, 15-17, 18-23, and
24-35 months. The amount of exposure and the number of events beyond 36
months are so low that it was deemed imprudent to fit the model beyond 36
months.
OCR for page 47
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
47
In many cases a woman reported that her menses never returned between one
birth and the next. In such cases, the return of menses was assumed to have
occurred 9 months before the birth. This treatment is consistent with our assump-
tion that menses is a proxy for ovulation. In other cases a woman reported that
menses returned, but the duration of amenorrhea exceeded the current age of the
child (for open intervals) or exceeded the interval from birth to the next concep-
tion (for closed intervals). We dropped these cases from the analysis of the
duration of amenorrhea.
It was expected that the death of the index child would shorten the time from
birth to return of menses (increase the risks) but that this effect would work
exclusively through the truncation of breastfeeding. Once breastfeeding was
accounted for, the effect was expected to be zero.
Interval from Birth to Sexual Relations
We estimated a set of models exactly parallel to those for postpartum amen-
orrhea using the timing of the resumption of sexual relations as the outcome. The
baseline hazard was estimated in the same way, and the same set of models was
estimated. Whereas postpartum amenorrhea is related to breastfeeding through
biological mechanisms, postpartum abstinence is related to breastfeeding only
through social norms and cultural taboos. Thus, the degree to which death of a
child affects the duration of postpartum abstinence and the degree to which this
effect is explained by breastfeeding were expected to vary from country to coun-
try.
Duration of Susceptibility to Pregnancy
A series of models was also estimated on the second part of the birth interval,
that is, after both menses and resumption of sexual relations had taken place. The
models were specified in the same way as the models on the full birth interval, but
all exposure during the periods of postpartum amenorrhea and postpartum absti-
nence was excluded. Births were excluded for which return of menses or re-
sumption of sexual relations was reported in the same month as the conception
because they would contribute no exposure. In these models, exposure and the
occurrence of subsequent births are classified by time since the index birth rather
than by time since the start of susceptibility to pregnancy. We estimated alterna-
tive formulations of the model by using time since menses or time since relations,
but results did not change substantively.
The effect of the death of a child on this portion of the interval could be
explained partially by breastfeeding because breastfeeding reduces the probabil-
ity of conception even for ovulating women. After adjustment for breastfeeding,
the effect was expected to be small in countries where contraceptive use was
OCR for page 48
48
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
infrequent because women have few means to control their fertility in response to
the death of a child.
Results
Interval Between Births
The first hazards model estimated (model 1) considered the time to the next
live birth, including only covariates of age and the survival status of the index
child. The coefficients of this model can be used to calculate a survival curve.
The point at which this curve reaches 50 percent represents the median duration
of the birth interval corresponding to the estimated model. In Figure 2-la, the
estimated median duration of the birth interval for intervals after the birth of a
child who survives 5 years is compared with the estimated median birth interval
after the birth of a child who dies in the neonatal period. This comparison is
shown for each of the countries2 for which data were analyzed. It thus gives a
sense of the amount of reduction in the median birth interval that can be associ-
ated with child mortality, without taking other factors into consideration.
The results have been grouped by three regions: Africa, Asia, and Latin
America. For each region the data have been arranged in order from the country
with the longest median duration to the country with the shortest median dura-
tion. Median birth intervals after the birth of children who are still living appear
to be shorter in Africa than in Asia or Latin America. However, birth intervals
after a child has died are generally similar across the three regions.
Table 2-2 gives the ratio of the estimated median interval for subsequent
births of children who survive to the median interval for births of children who
die. This ratio varies between 1.21 and 3.15, indicating that the increase in birth
interval associated with eliminating an early infant death is 21 to 215 percent.
The average reduction in the median interval is 60 percent for the 46 DHS used
for this analysis.
Figure 2-2 shows a plot of the estimated coefficient of the variable DIED for
all three models of the log of the hazard of closing the birth interval (conception
resulting in live birth). Because explanatory covariates are added to the models
in turn (confounders, then breastfeeding), the coefficients on DIED show the
degree to which the effect of DIED, independent of covariates, continues to be a
predictor of interval length. In this way, we try to assess how much of the effect
of the child's death can be explained away by controlling for other factors that we
also expect to affect birth interval length.
2Several of the countries analyzed here were actually surveyed twice. In the ensing presentation
of results we refer to "countries" as a matter of convenience, although the data points of interest in
fact represent countries at a point in time.
OCR for page 49
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
a. Full Birth Interval
on So
c: 45
~ 40
.= 35
In 30 ~ ~ ~ _ -
25
20
15
an 10 - 1 1 ~-~-~ t-} i- 1- ~ +--l ~ ~ 1 ~ .~.,i.1,.! 1- 1 it---
Afr~ca Asia
.. A: ~
~-t-- ~
in.
~a , ~WN . . ~
t '\~:~^ ~ t! ~ \-~-~^
.. . ... .. .. .. ~. . . y
..~2 .. ....
l- I---t- TV- t-~I~-i ,-l- ~
Latln
America
b. Postpartum Amenorrhea
0-' 16-~-
tu 12
E 10
<: 8
~ 4 :~:;~~~~~~~~~ ~ :`
O 2
- A- -4 I 1- 'Africa ~ ~ ~ ~ ~ t- tam ~~-~~-l i iis''a-'~l I I I I I " ILat~n
America
_
- - - -~- - - ~ -
c. Postpartum Abstinence
~ .
_~ l\~_ __ _ _ _ _ _ _ _ _ _ __ _. .
a' 20
v 18
16
~ 14
n 12
c 10
8
6
0 2 ~-~-. ~--~ '--~---~-.
O
~.
.
.. .
~- ~
. _ _ _ _ _ _ . ~ ~ ^~- ~-~ - -~--d ~N---~-- ~--
~ j ~ 1.~ ~ 1~L I ~ ~ ~ I ( ~} ~ ~I ~1 ~1 1 1 1 I I I ~ ~ i ~ I I ~ 4_4i ~
Afr~ca As~a Lat~n
America
~-~ Ch'ld living --~ ~ Child died I
' ~6P ql - _
FIGURE 2-1 Predicted length of interval by survival status of the index child.
49
OCR for page 63
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
Parameter Estimate
2.5
1 ~
. ~
~1 ~
~_; ~
~ \
~ \
Relative Risk
8.0
0
1 ~- Em, 1 2.0
0.5 ~
~\' ~
-0 . 5 -I - - - - -- - - - \.
l
-1 i .,,, ~ _ ~,, _ 1
0-1 1 12-23 24-35
Months Since Index Birth
-~.0
- 0.5
~ Kenya ~ Mali ~ Burundi ~ Ghana ~ Zambia
63
FIGURE 2-7 Age-dependent effect of child death on rate of menses returning postpar
tum.
Table 2-2 (column 3) gives the ratio of the estimated median duration of
postpartum abstinence for births of infants who were still alive, to the duration
following births of infants who died as neonates. Except in Rwanda, where the
relationship is oddly reversed, the ratios are 1.05 to 2.65. This variation indicates
that the lengthened abstinence associated with eliminating an early infant death is
5-165 percent. The relative effect of a child's death is not as great as that for
postpartum amenorrhea or for the overall birth interval. On average, across the
46 surveys, child survival increased the birth interval by 60 percent, postpartum
amenorrhea by 178 percent, and postpartum abstinence by 47 percent.
Again, three separate models were estimated for the hazard of resuming
OCR for page 64
64
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
sexual relations. Figure 2-8 shows that the effect of DIED estimated in the crude
model is virtually unchanged by the addition of either the confounding variables
or the variable for breastfeeding status. We conclude that the effect of a child's
death on increasing the risk of resuming sexual relations is direct and does not
appear to operate through breastfeeding.
However, there does appear to be a group of Afncan countries for which an
addition of the breastfeeding status variable actually increases the effect of DIED
over that in the initial crude model. In Table 2-5 (model 3 versus model 1), the
percentage of reduction in the excess risk is negative for most Afncan countnes.
This finding is puzzling and requires further investigation.
Parameter Estimate
4.6
41.4
].2
1
0~8
0.6
0.4
0.2
o
-0.2
0 4 t I | c | | | | ~I I ~I I I I I I |
-
-A ~
_ ~___~__
. _ ~ _
Africa
Relative Risk
_ 4.0
2.0
1.0
Asia Latin
America
a Crude
~ W/ Confounders · W/ Breastfeeding
FIGURE 2-8 Effect of child death on rate of resuming sexual relations.
OCR for page 65
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
Interval from Menses and Sexual Relations to Birth
65
For the first set of models for the entire birth interval, some countries showed
a general excess effect of a child's death on the risk of conception resulting in a
live birth, even in comparison to the effect of a living child who was weaned
(model 3 in Figure 2-2 and Table 2-3~. This finding was more likely in Latin
America than in either Africa or Asia. We decided to estimate an analogous
series of three models of the risk of conception resulting in a live birth. In these
models, all the exposure before resumption of menses and sexual relations has
been excluded. In this way, we eliminated postpartum amenorrhea and abstinence
from being explanatory mechanisms for the effect of DIED.
Figure 2-9 and Table 2-6 show the results of these three models. The results
of the first model show that, with just the crude effect of DIED, for all the
countries there is an excess risk for death of a child when conception of a subse-
quent child results in a live birth. This excess risk exists even when postpartum
abstinence and amenorrhea are removed as possible mechanisms. In Africa, the
crude relative risk associated with a child's death is greater than 1.5 in only 8 of
the 23 surveys analyzed in 21 countries, whereas it is greater than 1.5 in 3 of the
5 surveys analyzed in 4 Asian countries and in 8 of the 14 surveys analyzed in 10
Latin American countries. We can thus see that in Africa much of the effect of
DIED is operating through postpartum amenorrhea and abstinence, as expected.
Even within Africa an effect larger than 1.5 is primarily limited to two regions:
North Africa (Morocco, Egypt, and Tunisia) and East Africa (Burundi, Rwanda,
and Zimbabwe).
As before, a second set of models was estimated in which the confounding
variables were added to the first model. This generally had little effect on the
coefficients for the effect of DIED in Africa and Asia, but did substantially
reduce the effect in most Latin American countries. In Table 2-6, the addition of
the confounding variables in Latin America generally explained 30-60 percent of
the excess risk of conception resulting in a live birth, which is associated with
death of a child in the first model.
When a control for breastfeeding status was added to the model, the effect of
DIED was essentially eliminated for most countries in Africa and Asia. In
Africa, the effect of DIED is insignificant in all countries except Morocco; in
Asia, only Thailand has a relative risk (2.31) greater than 1.17 (Table 2-6~. In
Latin America, five countries continue to show a significant effect of a child's
death, but there is a substantial reduction relative to the crude model. This effect
of breastfeeding is not acting through lactational amenorrhea because menses has
already returned for these women.
We examined interactions of the effect of DIED with the sex of the child,
birth order, previous birth interval, and whether the child was wanted at that time.
We hypothesized that the effect of death of a child would be greatest if the child
were male, were of a low parity (especially first), or were wanted at that time.
OCR for page 66
66
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OCR for page 67
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OCR for page 68
68
Parameter Estimate
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BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
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FIGURE 2-9 Effect of child death on rate of closing the birth interval after menses and
sexual relations have resumed.
For none of these interactions did we find a consistent pattern of effect across the
countries.
DISCUSSION
In this analysis, we have demonstrated that the death of a child has a substan-
tial effect on the birth interval. We estimated that the median birth interval is 60
percent longer when a child lives than when it dies in early infancy. Preston
(1978) estimated that the death of a child would shorten the average birth interval
by 30 percent. However, the methods we used here are quite different from those
Preston used. First, for our estimate, we use the median birth interval associated
OCR for page 69
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
69
with the death of a child, rather than the overall mean birth interval, as the
denominator. Second, the 1978 estimate does not account for the timing of the
events of death of a child and the end of postpartum sterility and therefore allows
for a potential reverse causality. If menses returns early while the child is still
living, the subsequent interval to the next birth will likely be short, which may
itself be a risk factor for infant death (Hobcraft et al., 1985~. Furthermore, early
weaning could lead to both the death of the child and the early return of menses.
In our analysis, the survival status of the child is considered in the month prior to
the month for which risks are being estimated, so reverse causality is not a
problem. Third, our predicted birth intervals are based on creating a survival
function in which the child's status is dead or living for all durations. Thus, the
60 percent estimate presented here is the reduction associated with elimination of
all child death, not just infant death. Finally, whereas Preston's figure was based
only on the mechanism of shortening the period of postpartum sterility, our
estimate includes postpartum sterility as well as the interval when the woman is
susceptible to pregnancy because of not using contraceptives and coital frequency.
Our analysis has demonstrated that a major portion of the effect of a child's
death on shortening the subsequent birth interval operates through the premature
truncation of breastfeeding. Overall, of the excess risk associated with an early
infant death on closing the birth interval, 64 percent is explained by breastfeeding.
This percentage is nearly identical in all three continents.
As expected, we found that the duration of amenorrhea is longest in the
African countries. Death of a child has a substantial effect on the return of
menses. Although premature weaning explains part of this effect, we were sur-
prised to find a large effect remaining even after adjustment for breastfeeding.
The substantial effect of child death on the duration of postpartum abstinence
was also surprising, as was the finding that the effect does not seem to operate
through breastfeeding. We had expected that a major reason for prolonged peri-
ods of postpartum abstinence was a social taboo against intercourse during
breastfeeding and a belief that intercourse would poison the milk (Aborampah,
1985~. Instead, the norms for sexual activity appear to relate to the presence of
the child, rather than to feeding patterns.
The duration of postpartum abstinence from sexual activity is generally not
an important determinant of the length of birth interval because the period of
abstinence is much shorter than the period of amenorrhea. With the exception of
a few countries in Africa where the duration of abstinence is exceptionally long,
the median duration of amenorrhea is longer than the duration of abstinence in
every country. Particularly long durations of abstinence exist in the countries on
the Gulf of Guinea: Ondo State, Burkina Faso, Togo, Ghana, Cameroon, and
Nigeria. Notably, the effect of a child's death is far greater in Togo and Ondo
State than in any other countries, and a large portion of the effect is explained by
breastfeeding.
The effect of a child's death continues to operate even after menses and
OCR for page 70
70
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OCR for page 71
71
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OCR for page 72
72
BIRTH SPACING: A CROSS-NATIONAL ANALYSIS
sexual relations have resumed. Part of the effect operates through the truncation
of breastfeeding when the child dies. It is not clear whether direct physiological
mechanisms are at work here (such as anovulatory cycles after return of menses
or reduced fecundity) or whether indirect effects of the breastfeeding experience
are more important (such as fatigue and nighttime feedings which reduce coital
frequency). But whatever the reason, breastfeeding does seem to confer a protec-
tive effect against closing the birth interval, even after return of menses and
sexual relations.
Breastfeeding explains most of the susceptible period DIED effect in Af-
rica after controlling for breastfeeding, the effect of DIED is insignificant in
every country in this region except Morocco (DUSK. This finding is not surpris-
ing since contraceptive use is generally low in Africa. The effect of child death is
virtually explained away by breastfeeding because women have few mechanisms
to increase their fertility, such as stopping contraception after the death of a child.
On the other hand, the effect is largely explained by confounders in much of
Latin America. One hypothesis for this finding is that, in Latin America, the
confounders (especially mother's education, residence, socioeconomic status,
and parity) are acting as proxies for the propensity to use contraceptives. As
expected, the effect of a child's death on the length of the susceptible period
appears to operate primarily through breastfeeding and contraception.
We were surprised to find that the effect of DIED did not depend on whether
the child was wanted. We would have expected a much stronger effect of the
child's death if the child had been wanted, in that there would be a desire to
replace the lost child. It is unclear whether this result reflects the difficulties
mothers have in describing a child (even one who has died) as unwanted or
whether the wantedness of a previous child is simply not a good predictor of
subsequent behavior.
ACKNOWLEDGMENT
We thank Bridgette James for help in obtaining the Demographic and Health
Survey data sets and Sandy Jewell and Ellen Borland for their assistance in
managing the original data sets.
REFERENCES
Aborampah, O.M.
1985 Determinants of breast-feeding and postpartum abstinence: Analysis of a sample of
Yoruba women, Western Nigeria. Journal of Biosocial Science 17(4):461-469.
Guo, G., and G. Rodriguez
1992 Estimating a multivariate proportional hazards model for clustered data using the EM
algorithm, with an application to child survival in Guatemala. Journal of the American
Statistical Association 87:969-976.
OCR for page 73
LAURENCE M. GRUMMER-STRAWN, PAUL W. STUPP, AND ZUGUO MEI
73
Heckman, J., and B. Singer
1982 Population heterogeneity in demographic models. Pp. 567-599 in K. Land and A. Rogers,
eds., Multidimensional Mathematical Demography. New York: Academic Press.
1984 A method for minimizing the impact of distributional assumptions in econometric models
for duration data. Econometrica 52(2):271-320.
Hobcraft, J.N., J.W. McDonald, and S.O. Rutstein
1985 Demographic determinants of infant and early child mortality: A comparative analysis.
Population Studies 39(3):363-385.
Howie, P.W., A.S. McNeilly, M.J. Houston, A. Cook, and H. Boyle
1982 Fertility after childbirth: Infant feeding patterns, basal PRL levels and postpartum ovula-
tion. Clinical Endocrinology 17(4):323-332.
Konner, M., and C. Worthman
1980 Nursing frequency, gonadal function, and birth spacing among Mung hunter-gatherers.
Science 207:788-791.
Lapham, R.J., and C.F. Westoff
1986 Demographic and Health Surveys: Population and health information for the late 1980s.
Population Index 52:28-34.
McNeilly, A.S., P.W. Howie, and M.J. Houston
1980 Relationship of feeding patterns, prolactin and resumption of ovulation postpartum. Pp.
102-116 in G.I. Zatuchni, M.H. Labbok, and J.J. Sciarra, eds., Research Frontiers in
Fertility Regulation. New York: Harper & Row.
Preston, S.H., ed.
1978 The Elects of Infant and Child Mortality on Fertility. New York: Academic Press.
Sullivan, J., S. Rutstein, and G. Bicego
1994 Infant and Child Mortality. Demographic and Health Surveys Comparative Studies no.
15. Columbia, Md.: Macro International, Inc.
Trussell J., and T. Guinnane
1993 Techniques of event history analysis. Pp. 181-205 in D. Reher and R. Schofield, eds.,
Old and New Methods in Historical Demography. Oxford, England: Clarendon Press.
Trussell, J., and T. Richards
1985 Correcting for unmeasured heterogeneity in hazard models using the Heckman-Singer
procedure. Pp. 242-276 in N. Tuma, ea., Sociological Methodology. San Francisco,
Calif.: Jossey-Bass.
Trussell, J., and G. Rodriguez
1990 Heterogeneity in demographic research. Pp. 111-132 in J. Adams, D. Lam, A. Hermalin,
P. Smouse, eds., Convergent Issues in Genetics and Demography. Oxford, England:
Oxford University Press.
Representative terms from entire chapter:
child death
