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Below are the first 10 and last 10 pages of uncorrected machine-read text (when available) of this chapter, followed by the top 30 algorithmically extracted key phrases from the chapter as a whole.
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OCR for page 38
FINAL RESULTS
This section first examines models for the propensity to
breastfeed and for the propensity to contracept. Second,
among women who breastfeed and among women who contracept,
information about other experiences in the birth interval
is incorporated, and models of termination rates for
breastfeeding and for contraception, respectively, are
examined. Finally, factors affecting conception rates
are examined using information about the timing of
breastfeeding, contraception, and child mortality. The
scheme for the analysis ts given by the diagram below.
propens ity to
breastfeed~ --- how long? _
Background ~-
charac terist~
-
other experiences
;~ - ~~ ~:_,~ _
-
^" "~= U`~ll ' ~ fertility
interval ~ ~
- propene ~ ty
"v ,'
contrscept~ + how longly
child mortality
I
II
The details of the variables included in each equation
were given in Table 1.
38
III
OCR for page 39
39
LOGISTIC REGRESSION MODELS FOR .1~ PROPENSITY
TO BREAS'1~;D AND FOR THE: PROPENSITY TO CONTRACE:PT
These two equations use background characteristics to
predict the probability of breastfeeding and the
probability of using contraception in either the last
closed or the open birth interval. Colombia and Costa
Rica have quite different distributions of the two
dependent variables. Nearly all women breastfeed in
Colombia; in Costa Rica, although breastfeeding is very
common, it is not universal. In contrast, many more
women contracept in Costa Rica than in Colombia. Table 4
gives the results of the analysis. Since brea~tfeeding
is nearly universal in Colombia (only 6 percent of women
did not breastfeed in either of the two birth intervals
sampled) , the best prediction is that all women will
breastfeed. Under these circumstances, the model is not
very informative. Even though there is more variability
in the dependent variable in Costa Rica, the model does
not perform much better. Thus, the measured character-
istics indicating social and economic status do not
distinguish well between a woman who will breastfeed and
one who will not.
The analysis of the propensity to use contraception is
somewhat more successful. Although contraceptive usage
is much higher in Costa Rica, the impact of the social
and economic variables is much larger in Colombia. In
Colombia, the probability of use increases with the
education of both the woman and her husband, and is
higher for urban than for rural residents. In Costa
Rica, only the woman's education and urban residence have
an impact, and the effects are much smaller. This
confirms that contraceptive use is both more extensive
and less restricted to particular social and economic
groups in Costa Rica than in Colombia. It is interesting
that, although breastfeeding is less widespread in Costa
Rica than in Colombia, it is not any more restricted to
particular social and economic groups in one country than
in the other. The analysis by period in the appendix
shows that these results are not significantly altered as
births from progressively earlier time periods are
e liminated, suggesting that these results are not solely
attributable to changing practices.
OCR for page 40
40
~ -
TABL`E 4 Logistic Regression for Probability of
Breastfeeding and for Probability of Using Contraception
in Last Closed or Open Birth Interval
Colombia I Costa Rica
Breast feeding Contracept ion ~ Breast feeding Contracept ion
:
constant
work ~ ince
marriages
work may
f ras home
worn ' ~
educat ion
husband 's
educat ion
urban
res idencea
3.626
( .2344)
.0227
(.3011)
-.2051
( .3420)
-.0545
( .0481)
-.0627
( .0402)
-.5516*
( .2676)
-.9816
(.1051)
.1362
( .1552)
-.2262
(.1964)
.2032*
( .0308)
.1222*
( .0279)
.7413*
(.1214)
1.63?
(.1248)
.4303
( .2543)
-.4640
(.2848)
.0241
( .0275)
- .0439
( .0236)
-.0375
(.1626)
.8288
(.1221)
-.0605
( .2050)
.1716
.2519)
· 0900
( .0300)
.0118
t .0262)
.4441*
( .1674)
Mode 1 x2
d . f .
prey it t ive
accuracy
2 not us ing
27.35
.684
6.0
303.5 8.70
5 5
.164
.335
39.7 17.5
1643
47.24
.318
19.2
N
1643
1449
1449
Notes: Indicator variable; layer.
Standard encore are in parentheses.
* indicates significance at the .05 level .
HAZARD MODELS FOR U:RMINATION OF BREASTFEEDING AND
CONTRACEPTIVE DISCONTINUATION
Figure 6 shows the estimated hazard functions for
termination of breastfeeding (top panel) and
contraceptive discontinuation (bottom panel). These
hazards have been estimated without covariates and
contain the same information as the survivor functions
discussed earlier. The horizontal lines show the result
of assuming constant exponential discontinuation rates;
the jagged lines show the estimated hazards when they are
allowed to vary over the eight subperiods defined earlier;
the solid lines represent estimates for Colombia and the
dashed lines those for Costa Rica. First examine the
estimated hazard functions for termination of breast-
OCR for page 41
41
Breaetfeeding
in(t)
.150
.100
.050
in(t)
.200
)~
C ontraception
C ol ombis
_. Costa Rica
~"~_
3 6 1 2 18 ~4
36
48 t
.050
_
3 6 1 2 18 24 36 48 t
FIGURE 6 Hazard Functions for Termination of
Breastfeeding and for Contraceptive Discontinuation
feeding in the top panel. The horizontal lines show that
termination rates are higher in Costa Rica than in
Colombia. Colombia shows a pattern of moderate discon-
tinuation rates for durations of breastfeeding under 18
months, followed by a marked peak in discontinuation
rates between 18 and 23 months, suggesting that the
socially prescribed weaning time may fall within this
OCR for page 42
42
interval. The pattern for Costa Rica shows much higher
discontinuation rates at short durations, with one group
of women discontinuing before 3 months have elapsed and a
second group discontinuing after 6 to 11 months. Turning
to the estimated hazard functions for contraceptive dis-
continuation in the lower panel, the horizonal lines show
that on average, discontinuation rates are only somewhat
higher in Colombia than in Costa Rica. However, the time
pattern of discontinuation is quite different. Colombia
shows much higher discontinuation rates at shorter durar
Lions, and the two curves tend to converge after durations
of use longer than two years. The curves for Costa Rica
show two peaks of discontinuation, one at short durations,
the other at durations of 18 to 23 months. As discussed
below, these differences are primarily accounted for by
differences in the pattern of discontinuation of coitus-
dependent methods in the two countries, while the pattern
of discontinuation of coitus-independent methods is more
. ·,
similar.
Tables S and 6 give the estimated coefficients, their
standard errors, and the antilogs for the equation with
covariates predicting termination of breastfeeding.
These results are displayed graphically in Figure 7. The
strategy used in this and the other analyses of duration
is first to examine the impact of the background variables
alone (Model 1), then to add the covariates describing
other demographic characteristics (Model 2), and finally
to add the covariates that describe other events or
behaviors that occur in the interval (Model 3). The
log-likelihood for each model is given at the bottom of
the tables so that likelihood-ratio statistics can be
computed to determine the statistical significance of the
added covariates. First consider the effects of the
background covariates alone. In Colombia, as in Costa
Rica, the woman's education, her husband's education, and
urban residence are all statistically significant, and
the estimated coefficients are nearly of the same
magnitude. In both countries, these three variables are
all associated with shorter durations of breastfeeding.
Model 2 adds parity, which is also statistically sig-
nificant and of nearly the same order of magnitude. In
both countries, higher parity is associated with longer
durations of breastfeeding. Some of this may be a cohort
effect since average parity at the start of the last
closed interval increases by approximately one for each
five-year cohort of women between ages 20 and 45.
Because the number of observations is small, the model
OCR for page 43
43
TABLE 5 Coefficient Estimates for Termination of
B reastf ceding: Colombia
Model 1 Model 2 Model 3
coeff. antilog coeff.
antilog coeff. antilog
work since -.0557 .9458 -.0562 .9454 -.0616 .9403
marriages (.0629) (.0629) (.0628)
work away .0021 1.002 .0012 1.001 .0084 1.008
from home a (.0765) (.0764) (.0765)
women 'e education .0766* 1.080 .0722* 1.075 .0720* 1.075
(.0110) (.0112) (.0111)
husband's .0111 1.011 .0074 1.007 .0093 1.009
education (.0095) (.0096) (.0096)
urban residences .2343* 1.264 .2339* 1.264 .2327* 1.262
(.0516) (.0517) (.0516)
parity ' -.0234* .9769 -.0245 .9758*
(.0082) (.0082)
child survivals -.4022* .6688
(.0922)
Periot 1 (0-2 months)
constant -2.854* .0576 -2.693* .0676 -2.316* .0987
(.0585) (.0808) (.1176)
Periot 2 (3-5 months)
constant -2.787* .0616 -2.624* .0725 -2.245* .1060
(.0617) (.0835) (.1199)
Period 3 (6-11 months)
constant -2.579* .0758 -2.414* .0895 -2.032* .1316
(.0524) (.0776) (.1162)
Period 4 (12-17 months)
constant -2.661* .0699 -2.491* .0828 -2.102* .1222
(.0739) (.0945) (.1294)
Period 5 (18-23 months)
constant -2.429* .0881 -2.261* .1042 -1.866* .1547
(.0910) (.1080) (.1406)
Periot 6 (24+ months)
constant -3.130* .0437 -2.964* .0516 -2.561* .0772
(.1452) (.1562) (.1815)
loglikelihood -6594. -6590. -6582.
N - 2264
Notes:
a Indicator variable: 1-yes.
Standard error. are in parentheses.
* indicates significance at the .05 level.
OCR for page 44
44
TABLE 6 Coefficient Estimates for Termination of
Breastfeeding: Costa Rica
Model 1 Model 2 Model 3
coeff. antilog
coe f f . ant i log coe f f . ant i log
-
work. since -. 0826 .9208 -.0627 .9392 -.0640 .9379
marr iage a ( .0742 ) ( .0744 ) ( . 0744 )
work away .0942 1 .099 .08102 1 .084 .0847 1 .088
f ram home a ( . 0881 ) ( . 0881 ) ( . 0881 )
woman 's . 0439* 1 .045 . 0371* 1 .038 . 0368* 1 .038
educat ion ( .0095 ) ( .0098 ) ( .0098 )
husband's .0131 1.013 .0099 1.010 .0118 1.012
educat ion ( .0088 ) ( .0088 ) ( .0088 )
urban residences .1426* 1.153 .1342* 1.144 .1412* 1.152
( .0556 ) ( . 0557 ) ( . 0558 )
parity -. 0289* .9715 - . 0301* .9703
( . 0089 ) ( . 0089 )
ch i Id survivals - . 5909* .5538
(. 1296)
Period 1 (0-2 months)
constant -2 . 371* .0934 -2 . 160* . 1153 -1 . 597* .2025
( .0561 ) ( .0846) ( . 1484)
Period 2 (3-5 months)
constant -2.511* .0812 -2.297* .1005 -1. 727* .1778
(.0640) (.0908) (. 1535)
Period 3 (6-11 months)
.
constant -2.285* .1018 -2.071* .1260 -1.496*
(.0552) (.0848) (. 1513)
Period 4 ( 12-17 months)
constant -2.418* .0891 -2. 205* . 1102 -1 . 627* . 1966
( .0876) (. 1086) (. 1664)
Period 5 (18-23 months)
constant -2 .465* .0850 -2. 238* .1066 -1 . 657* . 1908
t. 1226) (. 1403) (. 1893)
Per lad 6 ( 24+ months )
constant -2 . 791* .0613 -2. 562* .0772 -1 . 973* . 1391
( . 1719) ( . 1853) ( . 2258)
10gl ikel ihood -5524 . -5519 . -5510 .
N a 1908
Notes:
a Indicator variable: layer.
Standard errors are in parentheses.
* indicates significance at the .05 level .
OCR for page 45
45
C al ombia
h (
.02
.01
h (t)
.02
.01
survive 1_ yes; urb a n- n o
s urvival-~; urban n o
...... survival no; urban - no
. . ·. .
.. ..
3 6 12 1 8 24
Costs Rica
'e .,..~e "~ -
·. .
t
. .
.
~1 1
3 6 12 18 24 t
FIGURE 7 Hazard Functions with Covariates for
Termination of Breastfeeding
OCR for page 46
46
could not be reestimated for each cohort of women, and
the statistical package used does not permit age to be
treated properly as a time-varying covariate. The issue
therefore remains unresolved.
Model 3 adds a dummy variable indicating whether the
child survived past age two. The coefficients of this
variable are of the same sign and are statistically
significant in both countries, but the estimate for Costa
Rica is nearly 50 percent greater than that for Colombia.
The corresponding hazards are shown in Figure 7. The top
panel in the figure graphs the results for Colombia, and
the bottom those for Costa Rica. The scale has been
adjusted for the mean parity and mean education of the
woman and her husband. The sol id line gives the hazard
for women whose child survives past age two--that is,
well past the usual weaning time--and who do not live in
an urban area. When we compare the two countries, there
is a tendency among women who have not yet stopped breast-
feeding to stop between 18 and 23 months in Colombia and
between 6 and 11 months in Costa Rica. The dashed line
shows the estimated hazard for women who live in urban
areas and whose children survive. The dotted line shows
the hazard for women whose children do not survive.
Clearly, breastfeeding durations are much shorter for
these children in Costa Rica, where durations of breast-
feeding are already relatively short. It was not possible
to obtain more detailed information on the interactions
of child mortality and breantfeeding from these data, for
reasons already cited; however, in both countries, child
mortality is lower among children who are breastfed than
among all children, even in Costa Rica, where mortality
in general is quite low.
Tables 7 and 8 give the estimated coefficients for the
equation predicting contraceptive discontinuation These
results are displayed graphically in Figure 8. Model 1
includes the effects of the background covariates alone.
Husband's education is statistically significant in both
countries, while urban residence is significant only in
Colombia. This is no longer the case when other covari-
ates are added (Models 2 and 3). In the larger models,
h usband ' s education is no longer s ignif icant; urban
residence becomes significant in Colombia, reaches
borderline significance in Costa Rica, and is associated
with longer durations of use. This last effect is larger
in Colombia than in Costa Rica, possibly due to the
greater concentration of family planning efforts in urban
areas in Colombia. Model 2 adds parity, desire for an
OCR for page 47
47
additional child, and contraceptive method. All three
variables are statistically significant in both countries.
Higher parities are associated with longer duration of
use in both countries, and the coefficients are about the
same size. Women who desire an additional child and
those who use a coitus-dependent method tend to use for
shorter durations; these effects are somewhat larger in
Costa Rica than in Colombia. When breastfeeding and child
survival were added, neither var table was statistically
significant in either country.
Model 4 contains terms for the interaction of desire
for an additional child and method of contraception with
subperiod to see how the shape of the hazard is modified
by each of these covariates. These results are presented
graphically in Figure 8. As before, the results for
Colombia are given by the top panel and those for Costa
Rica by the bottom panel. The scale for Costa Rica has
been expanded to show the detail; the level of discon-
tinuation is therefore considerably higher in Colombia.
The shape of the discontinuation curves is quite differ-
ent for the two countries.
The solid line gives the
estimated hazard function for women who use coitus-
independent contraception and who do not desire additional
children. In Colombia, high discontinuation rates for
these women are concentrated in the first 3 to 5 months
of use. In Costa Rica, the curve is not only lower, but
much flatter, although somewhat higher discontinuation
rates can be found in the first year of use. The line of
dashes and dots shows the estimated hazard for women who
use a coitus-dependent method and who do not want an
additional child. In Colombia, discontinuation rates are
relatively higher for durations of use under 18 months,
then drop to join those for the coitus-independent
methods. In Costa Rica, the discontinuation rates are
relatively high at rather short durations of use, moderate
at intermediate durations, and quite high for durations
longer than 18 months. The two dashed lines give the
estimated hazards for each of the two classes of methods
for women who desire an additional child. In Colombia,
the principal effect of this variable is to shift the
curve upward by about the same amount at all durations.
In Costa Rica, for each set of methods, it is only after
durations of use of one year or more that the discontinua-
tion curve for women who desire an additional child
consistently diverges from that of women who do not.
This suggests that in Costa Rica, for durations of use of
one year, women who want an additional child are as likely
OCR for page 48
48
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OCR for page 58
58
h (I)
.02
~'~
Colombia
_ _ _ . C osta Rica
A;
3 ~ 12 1 ~ 24 36 He t
FIGURE 9 Bazard Functions for Live-Birth Conceptions
other variables are added. Model 2 adds contraceptive
method and parity. Use of either coitus~dependent or
cottu~independent contraception lengthens the time to
next conception. The "pact of coitus-independent
contraception is nearly twice that of coitus-dependent
contraception in both countries, but the impact of either
set of methods is greater in Costa Rica than in Colombia.
This result holds when other variables are added to the
model, suggesting that contraceptive efficacy may be
greater in Costa Rica {see Goldman et al., 1982). Bigher
parities are associated with longer intervals in be ~
countries, and the estimated coefficients are close in
value.
Model 3 removes the background covariates and adds the
full set of t~.e-varying intermediate variables. Model 4
retur N the background variables to the acde1. Likelihood
ratio tests show that, although the effect of the back-
ground variables is small, it is statistically signifi-
c~nt. A co peridot of Models 3 And ~ reveals that few
coefficients change Then the background variables are
added. As already noted, along the background variables,
only urban residence is artistically significant. Figure
10 displays to iapact of the ti_'rarying cavariates
frog Hodel d. The results for Col~bla are in the upper
panel Id those for Costa Rice in the later panel. The
solid line gives the estimated hazard for men who do
not breasted and do not contracept' this corresponds to
OCR for page 59
59
the Natural fertility. line in Diagram 2 with gestation
subtracted from all intervals. Although the shapes of
the hazards for the two countries accord moderately well,
that for Costa Rica suggests an unexpected rise in the
interval 3 to 5 months. As expected from previous bio-
metric research reviewed earlier, the impact of breast-
feeding is more marked at durations of less than one
year, after which the two curves converge, while the
largest impact is for durations of under 3 months. The
dotted line, which shows the case where the child does
not survive, is most useful for comparison. For example,
if a child were breastfed for 6 months and then died, the
risk of conception for its mother would be the dashed line
for the first 6 months and the dotted line thereafter.
The estimated effects of this variable are rather larger
than expected, particularly at very short durations. Some
of this effect may be due to the tendency of short birth
intervals and high infant mortality to be mutually rein-
forcing in a manner that is not completely captured by
either breastfeeding or contraceptive behavior. The two
lowest lines in each panel show the conception rates for
women who contracept. The variable indicating contracep-
tive method shifts the hazard to a very low level in both
countries; the time-varying cover iates act principally to
rearrange the shape of the hazard slightly so that the
effect of contraception is not a simple proportional
shift of the solid line. This is particularly noticeable
in the last subperiod (24+ months), where the hazard for
contracepting women levels off from its downward course.
The similarity of these results for the fertility
equation in the two countries is striking, particularly
in light of the behavioral differences shown by the models
of breastfeeding and contraception discussed earlier:
not only are the signs on many of the coefficients the
same but many of the coefficients are close in magnitude.
Although different propensities and durations of breast-
feeding and contraception are used to obtain close to the
same level of fertility, the impact of a particular
behavior is nearly the same. This reinforces the idea
expressed earlier that this portion of the model accesses
fundamental biometric aspects of fertility, whereas the
other portions of the model are more behavioral in nature.
OCR for page 60
60
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OCR for page 61
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FIGURE 10 Hazard Functions with Covariates for
Live~Birth Conceptions
t
Representative terms from entire chapter:
discontinuation rates
