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OCR for page 29
PART I FERTILITY DETERMINANTS AT THE NATIONAL LEVEL
CHAPTER 1
THE PROXIMATE DETERMINANTS OF E ERTILITY
The objective of this chapter is to analyze Brazil's
accelerated fertility decline in demographic terms.
Demographic theory indicates that two sets of variables
are important: one is population composition, particu-
larly age structure and marriage patterns, both of which
mediate the relation between individual reproductive
behavior and birth rates observed in a population; the
other is comprised of the Davis-Blake (1956) ~ intermedi-
ate variables, ~ such as frequency of intercourse,
fertility control, breastfeeding, and abortion, which
d irectly affect reproductive outcome. This chapter uses
a standardization approach to identify compositional
effects, and Bongaarts ' (1980) method for decomposing
natural fertility into its proximate determinants to
identify intermediate variables.
This chapter is based on the available data for Brazil,
which are fragmentary in both regional and time coverage.
The approach is therefore essential detective work, piec-
ing together clues from a variety of sources in an attempt
to draw a picture at the national level. Changes In mar-
r iage patterns and age structure are always prime suspects
in declining birth rates; therefore, the discussion begins
by assessing changes in the distribution of women by mar-
ital status and in mean age at marriage, and then applies
standardization techniques to check whether these changes
and those in age structure played a major role in Brazil's
fertility decline. The available evidence suggests that
they did not. This suggests in turn that the primary
factor in the decline was one of the intermediate vari-
ables affecting marital fertility. The Bongaarts frame-
work is then used to explore this possibility. There are
three potential factors responsible--breastfeeding,
contraception, and abortion. Among these, the limited
29
OCR for page 30
3a
Degree to which the f irst is practiced in Brazil is
sufficient to eliminate it as a primary influence. The
evidence implicating the second is stronger, though
admittedly fragmentary. Finally, although the evidence
is clearly circumstantial, i t is strong enough to
implicate abortion as an important, though necessar fly
indeterminate, influence on Brazil' s fertility decline.
MARITAL STATUS AND MEAN AGE AT MARRIAGE
As noted above, the reporting of marital status in
Brazilian data is problematic. The reported percentage
distribution of Brazilian women aged 15-49 by marital
status in the 1950, 1960, and 1970 censuses and in the
1976 PNAD survey are presented in Table 5; preliminary
results of the 1980 census are also reported. Four
marital status categories are shown--married, divorced
and separated, widowed, and single. Each of these
categories presents its own set of problems.
First, in the married category, there is a problem in
the reporting of women in consensual unions. According
to Henriques (1980), women in consensual unions account
for an important share of Brazilian births, though Brazil
has a lower proportion of such births than a number of
other Latin American countries However, although
Brazilian data include women in consensual unions as a
subcategory of married women, there is strong evidence
that a number of those who report themselves as single
may in fact be in "=nsensual unions {Si}va, 1979:14).
This is suggested by data in Table 6, which shows the
percent of women in the single category who reported
having had a birth from age 15-19 to 40O49 in the
censuses and P=D survey.
Brazilian census authorities have attempted to improve
on the reporting of consensual unions by broadening the
number of categories of mar ital status to include the
type of union. In 1950, when there was no subcategory
for consensual unions, nearly four out of ten single
women reported a birth by the end of their reproductive
years, suggesting that most of these unions were grouped
in the single category (Al~'nann and Wong, 1981a:3561.
This contrasts with 1960, when the consensual union
category was introduced and the proportion of single
women reporting a birth dropped Deco 11 percent. It should
be recalled that tabulation of the 1960 census was
delayed until the late 1970s because of administrative
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31
TABLE 5 Reported Percent Distr ibution of Women by
Marital Status and Age, 1950-80: Brazil
Divorced,
Age Marr ieda Separated
Widowed S ingle Total
1950
15-19 14 ~ 8 0 0 ~ 0 0 1 85 ~ 1 100
20-24 51.9 ~ ~ O 0 0 7 410 4 100013
25-29 70~4 001 lc8 27~7 100~0
30-39 7tje3 002 501 18~4 100~0
40~49 7102 Oa3 1401 1404 . 1Ot)~0
1960
15-19 14~0 (1~3) 007 0~1 8502 100~0
20-24 53~3 (3~6) 2~6 004 43~7 10000
25-29 7400 (5~1) 304 101 2105 - 10000
30~39 80~9 (504) 3~9 3~1 12~1 10000
40~49 76~2 t409) 408 1001 8~9 100~0
1970
15-19 12~0 (1~4) 0~5 0.1 87.4 100~0
20-24 46~9 (3~9) 2~0 0~3 50~8 10t)~0
25-29 71~3 (5~4) 3~0 009 24~8 100~0
30~39 80 ~ 2 (5 ~ 7) 4 ~ 1 2.8 12.9 1000 ~
40~49 `6~2 (4~9) 5~6 8~9 9~3 100~0
1976
lS-19 11~2 (109) 0.6 001 88.1 10000
20-24 46~2 (504) 2~0 002 51~6 100~0
2S-29 69~8 (6~8) 3~0 0~8 26~. 100~0
30~39 19~5 (7.4) 4O9 2~6 13~0 10000
40-49 77.7 (6.6) 5.9 804 800 100.0
lg80
15-19 16~3 {3~5) 0~5 OoO 83c2 100~0
20-24 53~2 (8~0) 2.a 0~3 44~5 100~0
25-29 72e 6 ( 9. l) 3.2 0 0 7 23 ~ 5 100 ~ 0
30_39 79, 9 ( 9 ~ -2 ) 4 ~ 5 2.3 13 ~ 3 100 ~ 0
40~49 77~0 (7~7) 6~3 7~7 9~0 100~0
aFigures in parentheses represents when in consensual unzoned as a
percent of all women, when reported.
bThe total excludes women who did not report marital status.
Sources: 1950-80 from population censuses' 1976 from PLED survey.
breakdowns, and that procedures used in reconstructing it
may have biased the results. The repot ted trend from
1960 to 1976 is puzzling, since it suggests that births
to single women increased. It is possible that PEAS
interviewers, who were more highly trained than census
interviewers, were more careful in identify sng single
mothers.
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32
TABLE 6 Percent of Single Women Who Report Having
Had a Child, by Age, 1950-76: Brazil
Age Group 1950 1960 1970 1976
15-19 1~7 0~4 0~7 1~5
20 24 10~0 2~4 3~8 6~0
2529 22 0 3 5 e 5 8. 7 10 . 2
3~)~39 34O3 8~8 14~6 17~9
40 49 363 0 7 10 ~ 9 16 0 3 23 ~ 5
Source: Published tabulations of census and survey
data.
The impact of misreporting of women in consensual
unions appears to be greatest among younger women. For
women aged 20-24, the proportion reported as married
increases from 51.9 percent in 1950, when consensual
unions were not included as a subcategory, to 53. 3
percent in 1960 ~ when the subcategory was introduced. To
illustrate the possible influence of consensual unions on
reporting, the percentage of women in the consensual
union subcategory is indicated in parentheses in Table 5
next to the percent married after 1960 (the basis of this
percentage is all women in the age category). The
proportion of women aged 20-24 reported as married
decreases from 1960 to 1976 0 then increases sharply to
S3O 2 percent (about equal to 1960) in the 1980 results.
At the same time, the percent of women in consensual
unions increases, particularly ire 1980, when the f igure
is 8 percent of all women ( IS percent of marr fed women) .
Published tabulations of the preliminary 1980 results do
not supply enough information to determine whether, in
editing, single women with births may have been reclassi-
fied as being in consensual unions. The evidence suggests
either that consensual unions have been increas ing or
that reporting of such unions has improved. In all
likelihood, results for earlier periods understate the
proportion of younger women reported as married. Such
underreporting would bias the age-specific marital
fertility rate for these women upward if that proportion
were used in the calculation of the rate; it would also
add a further bias to the extent that consensual unions
were recognized when a woman gave birth to a child.
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33
Although the impact of the underreporting of consens~al
unions is reduced by grouping all single women with births
in the married category, it is not eliminated. One is
still left with the problem of women in informal unions
who have no children, but should be included in the
denominator when fertility rates are calculated. Thi
omission could bias calculation of rates for younger
women, particularly when the beginning of an informal
union is not clearly demarcated and the birth of a child
leads to recognition (or admission) of the union.
Including in the married category only those Musingly
women who had births will lead to overstatement of
marital fertility rates. One way to compensate for this
is to assume that the proportion of single women sat
risk. Is equal to that of married women; that is, if SO
percent of married women aged 20-24 report a birth, then
assume that the single women aged 20024 reporting a birth
represent 50 percent of the single women at risk. This
at-risk group can then be added to the married category.
s
_. . . . . .
This procedure will be employed later in calculating the
denominators for marital fertility rates. The issue of
type of union and its relationship to fertility at the
local level is explored in detail in Part II of this
report.
A second problem of consistency in the Brazilian data
on marital status relates to the reporting of separated
and divorced women. Because Brazil legalized divorce
only in 1977, its effects cannot be observed directly
even in the 1976 survey. However, reporting of foreign
divorces, legal separations, and a special Brazilian
legal substitute for pre-1977 divorces (desquites) was
increasing in the years prior to actual legalization.
The contrast is sharpest between 1950, when practically
no divorces and separations were reported, and the other
data points. Because divorce was not legally recognized,
it was fairly common practice for women who had entered a
second union after separation to be declared as separated
rather than married, leading to further underreporting of
the proportion of women in unions.
Such problems with the reliability of marital status
data suggest that considerable caution is required in the
calculation and interpretation of measures incorporating
those data. This applies not only to the proportion of
women in unions, but also to measures of the average age
at marriage and marital fertility rates. Table 7 sum-
mar izes the results of calculations of the singulate mean
age at marriage (SMAM), which is based on the proportion
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34
TABLE 7 Singulate Mean Age at Marriage (SMAM), by
Region, 1950-76: Brazil
PNAD Reg ion
1960 1970 1976
1. Rio de Janeiro 22.92 23.17 23.93
2 0 Sao Paula 22.25 23.30 23 0 47
30 Southern States 21.66 22.17 22.80
4. Hinas/13spirito
Santa 22 0 36 23. 36 23 ~ 66
50 Northeast Stabs 22.lB 22 23013
6. Brasilia 20.4S 23.38 23.73
7. 1?~tier States =.~] 22.04 °-
Brazi.1
Marriage
First Birth
22 .1l 22. 91 23.3 3a
22.42 23.28 23.84a
aThe Brazil f igure for 1976 excludes rural areas
of Reg ion 7, which was not included in the PNAD
survey.
Source: SMa~ calculated from census and survey
distribu~cions of women by marital status considering
single women who reported having had children as
married. First-birth measure based on same
computation as SMAM, but substituting the proportion
of women reported as childless.
of women reported as single at different ages. In order
to reduce the impact of underreporting of consensual
unions, single women reporting a birth were considered
mart fed . SMAMs were calculated for the PNAD reg ions and
Brazil as a whole ire 1960 Of 1970, and 1976. For Brazil as
a whole, the data indicate a comparatively late average
age at marriage and suggest that this average increased
by more ~chan one year between 1960 and 1976. Most of
this increase came between 1960 and 1970, and resulted
from a more than 6 percentage point rise in the propor-
tzan reported as single between the two dates. It was
Dot Audible tc' include single women with births in the
<:a'=ulati~ of t~ SW~£ for 1980; nevertheless, with the
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35
decreased proportion reported as single in the prelimi-
nary results, the average age at mart iage dropped back to
the 1960 level of 22.1 years. It should also be noted
that the preliminary 1980 results are not fully comparable
to those for earlier dates since it was necessary to use
10-year age groups for women over age 30.
The breakdown of SMAlls by region indicates a fairly
homogeneous pattern, with lower average ages in the
Southern and Frontier states, and higher average ages in
the Southeastern states (Regions 1, 2, and 4). The
largest increase in average age at marriage is found In
Region 6 (Brasilia); it may be observed that the compo-
sition of Brasilia's population changed significantly
from the 1960s, when it was being constructed, to the
1970s, when it began to function fully as the national
capital. For the remaining regions, increases range from
.5 to 1.0 years in 1960-70 and .2 to .6 years in 1970-76.
Since there are questions about the reliability of the
mar ital status data on which the SMAMs are based, another
way of looking at the age at which exposure to the risk
of childbear ing starts is to consider the mean age at
first birth. This can be derived from data on the
proportion of women who remained childless at different
ages us ing the same computational procedure employed in
calculating SMAMs. The resulting index, calculated only
for national-level data, is shown in the second row of
results for Brazil in Table 7. For 1960 and 1970, the
average age at first birth is .3 years higher than the
age at marriage. This increases to .5 years in 1976.
The 1980 average, not shown in the table, is 22.1 years,
just equal JO the average age at marriage.
MARITAL FERTILITY
How much did changes in the proportions married and age
at marriage contribute to the acceleration of Brazil's
fertility decline? Table 8 attempts to link total and
age-specific fertility rates (ASFRs) to total marital
fertility and age-specific marital fertility rates
(ASMFRs) using the 1960, 1970, 1976 ~ and 1980 data. The
f irst column of the table shows ASFRs and total fertility
rates calculated for the Brazil report of the Committee
on Population and Demography, National Academy of
Sciences. The total fertility rates are the same as
those reported in Table 2, showing declines of about 6
percent from 1960 to 1970, 24 percent from 1970 to 1976,
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36
TABLE 8 Age-Specif ~c And Total Mar ital Fertility
Rate Calculations, 1960, 1970, 1976, and 1980: Brazi 1
Age
Group ~sppa
Percent Mar r iedb
~ a ~ ~ b ~ ASMP Rc ASMP~
1960
15-19 e0799 14 ~ 0 15 ~ 3 ~ 354
20-24 ~ 2719 53.3 57 ~ 4; 0 472
25 29 03150 74~) 7908 0395
30~34 02615 8003 87cS .299
35 39 01935 81. `e 90 ~ 8 ~ 213
40 44 . - 08 77 O 9 910 7 ~ 099
45-49 .0239 74.3 92.7 .026
Total 6~18 58~9 65~7 9~29
1970
15-19
20-24
25-29
3 0~34
35~39
40~4
45~49
~otal
1976
15-19
20-24
2So29
30~34
35~39
40~~.
45~49
Tota1
1980
15-19
20-24
25 -29
30~341
35-39
4044
4 5 - 49
Pota1
.0753
2564
2971
0 2466
ol821;
~ 08S6
.0225
5~83
.0733
0 2062
0 2240
0 1814
.1292
0 OS88
oO108
~o45
~ 0850
4 20S.
0 2071
D ~L604
c 1030
.0Je188
~ 0093
4.10
1200
46~9
7103
80~2
8101
77~8
741~3
55~5
11~3
46~3
69~8
7808
8003
79~d
75~7
55~2
.163
S32
.726
799
799
770
.770
S6~5
13 o7
51~6
77~6
87~3
91~0
9200
92~8
62~8
14~1
52~3
760S
8704
9106
93~7
94~2
63~3
~ 168
.555
765
867
~ 867
.910
.910
64!~8
. 373e
.497
.383
~ 282
0 200
.093
0024
9~27
· 295e
.394
0 293
0214
o 141
eG63
~ 011
7~06
0 278
0 370
0 271
ol8S
119
054
oO10
6~.
352e
~ 469
.384
~ 287
.209
~ 102
.033
.l8
. 284e
~ 379
.292
0222
ol47
a071
.015
.05
Sources s
ASF8s Ag~specific fertility rates Pancl on Brazil. Cc~itte.
on Population and ~raphy, Mational Acaday of Scienece.
Pctcent ~rried: in coluan ta) as reported in published
h^h""tione, in colu~ (b) adjusted to include all everqrried
wa~n' and, except for 1980 when data not available, single wa~n
·st risk- ~ expla~d in text.
ASMYRs Ag repecific earitsl fertility rates co~put.d using ASEa
and percent carried, verdant (b).
AS~FRs Age-specific easite1 fertility rates coo puted fro
specla1 e~bolation of birthe in previous year for ever-aarried
wooen, and and adjust d by P/P ratios for all wo.en tl.32 in 1970
and 1.22 in 1916). Not available for 1960 and 1980.
ASMFR for 15-19 cooputed as .75 ASMFR for woann aged 20-24.
OCR for page 37
37
0.5
0.4
LU
a:
a:
0.3
-
-
CC
up
Cal
-
C: 0.2
In
Al
Or
0.1
o
_ _ Santa Cruz-U rban
Parnaiba-Rural
/N
I ~
/
/
\
\
>_ 1
40 45 50
1 5 20 25 30 35
AGE
FIGURE 5 Age-Specif ic Fertility Rates, 1975: Brazil
and about 8 percent from 1976 to 1980. As can be seen in
Figure 5, the shape of the age-specific fertility rate
profile became somewhat flatter with the decline, which
was substantial for all age groups except the youngest.
The largest absolute decline occurred at ages 25-29, with
larger proportional declines among the older age
categories.
A first attempt was made to translate ASFRs into
age-specific marital fertility rates by using data on
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38
marital status from published reports, as shown in column
(a) under percent married" in Table 8. Because of the
underreporting of women in consensual unions, ASMFRs
appear to be biased upward, particularly for women in
their twenties. In 1970, for example, the ASMF~ for the
20-24 group would be .547; this would mean that the
fertility of Brazilian women exceeded that of the Hutter-
ites' one of the populations on whose experience standard
natural fertility schedules are based. To reduce this
bias' the proportion ~married. was adjusted to include
all everom~rried women, single women who reported a
birth, and a prorated number of single women eat risk"
based on the proportion of married women reporting a
birth, as shown In column (b) under percent married. in
the table. This adjustment raised the percent married by
about 7 percentage points for each observation; although
this did not reduce the decline of about 5 percentage
points in the proportion married among women 20-24 between
1960 and 1970, it did narrow the differences between
1970, 1976, and 1980. The increase in the proportion
marr fed among women in older age categor ies ar ises f rom
the addition of widowed, separated, and divorced women.
Even with these adjustments, the ASMFas and total
marital fertility rates for 1960 and 1970 are quite
hiah - nearlv 9~3 for both dates.
By 1976, total marital
_, ~
fertility falls to 7.06, a decline of 24 percent, with a
further decline of 9 percent to 6O44 in 1980* While
total marital fertility shows little change between 1960
and 1970, the rate for women aged 20-24 increases. This
occurs because the decline in the ASFR for that group is
less than the decline In the proportion married, even
after adjustment. The rate for women aged lSol9 is
similarly affected since it was calculated as .75 of the
rate of those aged 20024. Because these rates are so
high, total marital fertility is very sensitive to such
differences, which are as likely to be the result of
differences in reporting (or editing, as suggested above )
as they are to be real. Thus considerable caution is
required in interpreting the 1960-70 period.
Access to public use sample files for the 1970 and
1976 data made it possible to tabulate the observed
nether of births by marital status in the year prior to
interviews; this provided a check on the rates calculated
from ASFRs. Observed births were adjusted by the IF
ratios used in ad j usting A5FRs for that date. The
resulting ASMF" and total marital fertility rates are
shown in the last column of Table 8. These rates are not
OCR for page 39
39
exactly comparable to those calculated directly f rom
ASFRs s ince the ad j us~ment of observed births used in
deriving ASFRs also included a factor to account for the
one-half-year dif ference between women' s reported age and
their age when births actually occurred. Because it is
inappropriate to apply this factor when the denominator
is limited to ever-married women, the age prof files of
ASHFRs differ with the two approaches. The main d~ffer-
ence is that rates based on the tabulations are lower for
women under 25 and higher for women over 30. Total mar-
ital fertility is slightly lower in 1970 and virtually
the same in 1976 in the tabulated results O compar ison of
the two sets of rates suggests that the decline in mar-
ital fertility probably was close to 24 percent over the
1970-76 period; however, there is a need to be cautious
about both the level and age profile of marital fertility
rates .
The age category most seriously affected is 20-24.
Two factors could account for an overstatement of the
age-specific marital fertility rate for this group. One
is the underreporting of married women discussed above.
The other is that the adjustment factors derived using
the Brass technique may be too hzgh for this group.
Estimates of ASMFRs using model marital fertility
schedules suggest that this may in fact be the case (see
Berquo and Leite, 1979; Altmann and Wong, 1981a; Leite,
1981~. One way to visualize the extent of possible bias
is to compare observed ASMFRs to model schedules, as
shown in Figure 6. Both of the 1970 ASMFR schedules from
Table 8 are plotted against these model schedules. The
models are based on a total marital fertility rate of
9.25, with the Coale-Trussel index of fertility control
set at three levels: 0.0, the natural fertility level,
and O.3 and 0.S, indicating moderate fertility control.
The model schedule with the index set at 0.5 provides a
very close approximation to both of the observed 1970
ASMER schedules for ages 25-29 to 40-44. The observed
rates for age 20-24 exceed the model schedule, though the
second schedule (estimated from births reported for
~married. women, broadly defined) is closer than the rate
calculated from dividing ASFRs by the proportion married.
The compar ison suggests that the 1970 base for calcu-
lating declines between 1970 and 1976 among the younger
age groups may be too high, resulting in an exaggeration
of declines for women in these age categor ies.
OCR for page 49
49
Abor t ion
Abortion is illegal in Brazil, and the Brazilian penal
code carries a sanction of several years of imprisonment
at hard labor for both abortionists and women who practice
it (Milanesi, 1970:12-13). However, the code is rarely
invoked, except as a legal ploy in such efforts as the
attempt to block the importation and use of the IUD as an
~abortifacient. n A ~ rtion is believed to be widespread;
rates of mortality and hospitalization resulting from
abortion-related infections are high, though not well
documented.
There are no reliable statistical data on abortions at
the national level in Brazil, though there are reported
estimates (based on some rather implausible extrapola-
tions from the experience of hospitals in a few locale
i ties ) . According to these estimates, the annual number
of abortions runs as high as three million per year,
which would mean eight abortions for every ten live
births (Rodrigues et al., lB75; Carvalho et al.' 1981).
The hand evidence that does exist suggests that ~ or,
urban, and noncontracepting women are more likely to
resort to induced abortion, and that only a fraction of
there women identify themselves when the question is
asked in survey interviews. Martine ( 1975) suggests that
another reason for low observed abortion rates is the
high mortality rate among women having abortions who, as
a consequence, would not be accounted for. Questions on
abortion have been asked in a number of fertility surveys
conducted in Brazil since the early 1960s, including the
NIER and CPS; these provide a basis for calculation of
the abortion index.
The way the incidence of abortion is reported varies
by survey. For purposes of calculating Bongsarts' index
of the effect of abortion on fertility, the ideal would
be the total induced abortion rate; thin is analogous to
the total fertility rate and indicates the number of
abortions a woman would have during her reproductive
lifetime if current rates prevailed. Either annual
age-specif~c abortion rates or the total number of
abortions for women who have completed their reproductive
life cycle could be used to calculate such a rate. Bows
ever, samples are rarely large enough to give reliable
estimates by the first method, and in only a few of the
Brazilian studies are rates broken down by age. Varia-
bility in measures affects both the numerators and
denominators of rates. In some cases, numerators refer
OCR for page 50
so
to induced abortion, in others to all abortions (spontane.
ous and induced); some numerators indicate the number of
abortions, others the number of women ever having had an
abortion. Denominators include the number of women of
reproductive age (usually restricted to women in unions)
and/or the total number of pregnancies reported by these
women.
Table 13 summarizes available statistical information
on abortion in Brazil. Hutchinson's data on Rio de
Janeiro in 1963 indicate that 9 percent of married women
of reproductive age had experienced an abortion, while a
1965 REWRAP study of Sao Paulo reveals that 18 percent of
pregnancies ended in abortions, about one-third of which
were induced. Martine's data on poor women in Rio de
Janeiro in 1969 suggest a substantially higher incidence
of abortion amona low-income arouDs.
_
Etges' study of
three municipalities in Rio Grande do Sul in 1973 shows a
higher rate of induced abortions as a percent of all
pregnancies in Porto Alegre than in two smaller municioal°
ities that he sampled. Questions on abortion have been
included in all six of the CPSs for which tabulations are
available. Women were asked how many abortions (spontane-
ous and induced} they had experienced, and whether their
last abortion was spontaneous or induced. The proportion
of induced abortions in the ~978 Sao Paulo survey was
much lower (closer to one ninth) than the one°third figure
of ache 1965 Sao Paulo survey: this suggests (l) that
induced abortion is underreported in the CPS data, and/or
(2) that a higher level of contraceptive use in 1978 Nay
have resulted in a substitution of contraception for
abortion (Nakamura et ale, 1979:17). The proportion of
women reporting having had an induced abortion was hither
in urban areas of Sao Paulo.
In other CPSs, the propor-
tion of women reporting that they had ever had an abortion
was higher in rural areas, but the proportion of preq-
nancies ending in abortion was higher in urban areas.
CPS abortion data have been tabulated by women's age;
this permitted a rough approximation of the total abortion
rate, which was calculated as the proportion of the
difference between the total pregnancy rate and the total
fertility rate that could be attributed to induced
abortion. This difference averaged . S7 per woman in the
Northeast, where the total fertility rate was 6. 3 per
woman. If the reported 10-}5 percent share of induced
abortion were accepted, the total abortion rate would be
a very low .06 to .08 per woman; if all reported termina-
tions of pregnancy were attributed to induced abortion,
OCR for page 51
51
then the total induced abortion rate would be .57 per
woman.
Berquo {1980) has calculated total abortion rates for
the nine localities included in the NIHR survey. These
range from .068 abortions per woman in Santa Cruz-Urban
in the state of Rio Grande do Sul to .735 in Parnaiba-
Rural in the state of Piaui. There is no consistent
rura1-urban pattern, though there is a suggestion that
abortion rates are higher in larger cities and in poorer
reg ions . I ~ is also pass ible to speculate on total
abortion rates using CPS data on the proportion of
pregnancies ending in abortion. In Salvador, Bahia, for
example, assuming that one-third of reported abortions
are induced and the total fer tility rate is about 4, the
total induced abor tion rate would be about 0 2 per woman,
for Bahia-RuraL, with a total fertility rate of 7, the
total abortion rate is about the same. For Recife,
calculation based on a total fer tility rate of 4 yields
an abor tion rate of . 25 per woman, about half of the rate
indicated in the NIHR data; this is again consistent with
the interpretation that tl) the CPSs underreport abortion
more than does the NIMR survey, and/or ( 2 ) increased
contraceptive use is being substituted for abor tion.
The Bongaarts index for measuring the effect of
abortion on- the fertility rate is defined by the
expression
Ten
C_ ~ ~
a ~
TFR ~ 0.4 {Lou) TAR
_,
where "u. is the parameter already estimated when calcu-
lating the contraception rate. Ca becomes zero if all
pregnant yes are abor ted, and is equal to 1 if all at e
successful. In applying this index to recent Brazilian
fertility trends, one can at best speculate on orders of
magnitude suggested by these survey results for the total
induced abor tion rate. There are several qualif ications
to be noted here. First, it is very likely that the
survey data presented above greatly understate the abor-
tion rate. Second, in addition to showing the prevalence
of abortion, war iation in 80ngsarts ' abortion index also
reflects the level of fertility and of contraceptive use
( to the extent that both, in turn, are related to the
potential number of pregnancies to be abor ted) . Thus
abor tzon, contraception, and fertility must be balanced
in any calculations. Finally, uncer tainty about the
level of the abor tion rate also relates to the impact of
OCR for page 52
52
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OCR for page 53
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OCR for page 54
54
nonmarriage. It is one thing to assume that most births
accrue to women in unions; it is quite another to assume
that all pregnancies, particularly those that are aborted,
are within unions. In all likelihood, a significant
fraction of abortions may be experienced by women who are
not in unions. To the extent that these abortions are
understated, the estimate of nonmarriage may be over-
stated.
FOE illustrative purposes, a range for the total
abortion rate of 0.5 to 1.5 per women may be assumed.
With an estimated 1976 fatal fertility rate of 4.4,
yielding about 3.75 million births in Brazil, a range
estimate of between 500, 000 and 1, 2SO, 000 annual induced
abortions is implied. While these figures fall well below
the 3 million annual abortions derived by extrapolation
from newspaper and other reports, they seem more plaus-
ible than the very low abortion rates yielded by survey
data.
Breastfeedinq/Postpart~m Amenorrhea
Statistical data on breastfeeding and postpartum amenor-
rhea in Brazil are also quite deficient. Questions on
breastfeeding were included only in more recent
inquiries ~ the NIBR and CPSO However, the data are more
consistent for the purpose of calculating Bongaarts'
index of the effect of postpartum infecundibility on
fertility than was the case with abortion. Mean or median
months of breastfeeding are provided in reports on the
NIBR and CPS by Anderson et al. (1981) and Berquo (1980},
as shown in Table 14. Calculation of the Bongsarts index
requires a transformation of mean or median months of
breastfeeding to months of postpartum amenorrhea, which
i s the ~ i. in the formula for the index: Ca ~ 20/18 . ~
~ i. Anderson calculated i for Bahia, Pernambuco, Rio
Grande do Nor te, and Paraiba using the Lesthaege-Page
model schedules, and a formula based on this schedule was
used to transform the remaining observations.
While not representative of the entire Brazilian
population, the data in Table 14 do span a broad enough
spectrum of experience, from poorer Northeastern areas to
the industrialized Southeast and the more developed rural
areas of the South, to permit speculation about a plau-
sible national-level index. The index values in the
table, as well as the breastfeeding data on which they
are based, show that the practice of breastfeeding is
OCR for page 55
55
TABLE 14 Reported Breastfeeding, Su~ary of Survey Data:
Brazil
Plonthe of
Breset-
feeding
Monthe of
A~nor-
rhea
Index
S ite of Survey Date Meen ~dian Median Ci
.
Sao Paulo~Urban 1978 < 1 20 0 .97
Sao Paulo~Rural 1978 7O. dO. .87
Pinui, Teresina 1979 3 0 3 2 0 7 0 94
Pieui- - st 1979 901 So3 ·84
Bahia p Salvador 1980 ~O 8 ~ ~ 2 ~ 9 . 93
Bahia-Rest 1980 9. ~ 3 0 B 4.0 .89
Paraibe-Urban 1980 4.7 <1 2. ~ .96
Paraiba-}tural 1980 5.5 10 8 3 0 2 . 92
Pern~-huco, Recife 1980 3O7 <1 2.2 .97
Pernambucs - Rest l9BO ~O ~ < 1 2 01 . 9?
Rio Grande do tiorte-Urban 1980 4.2 <1 205 .95
Rio Grande do Norte-Rural 1980 5.0 <1 2.8 .g.
Cachoeira-Urban (ES) 1975 8.5 5.0 .85
Santa Cruz-Urban (~;S1 1975 4.2 3OO O93
Santa Cruz-Rural (RGS) 1975 S.5 3.6 .9a
Sao Jose-Urban (SP) 1975 6.0 3.8 .90
Sertaosin~Rural (SP) 1975 7.3 4. ~ .87
Recife-Urban (PB) 197S 3.3 20 7 .94
Conceicso A.-Rural (PA) 1975 8.5 5.0 .85
Parnaiba-Urban (P$) 1975 4.8 3.3 .92
Parnaiba-Rural (PI) 1975 603 3O9 .89
Sources: 1975 data froa Berquo (1980sTa:bl. II) S all otb'ar data fro.
Anderson et a1. (1981:Table 5J.
very lin~ited. Indeed, Brazil falls at the very low end
of the spectrum of <:ountries for which breastfeeding data
are available, including Latin American countries, which
have low rates compared to other regions of the world
(Kent, 1981~. The index values range from .84 to .97,
with lower values (higher breastfeeding ~ in rural areas
and higher values in the cities. Values are also higher
in areas in which contraceptive use is higher, suggesting
that the practice of breastfeeding decreases as contracep-
tive use increases. These values are much more co~ar able
to tho~e of o~er industtsalized countries than to those
of developing countries' they are high even in comparison
with other Latin American countries that also have higher
values in comparison with other developing regions.
~For purposes of computing an index of postpartum
infecundabili~cy at the national level, an initial 1970
level of 4 month e was assumed. Since there is evidence
OCR for page 56
~6
to suggest that increased contraceptive use leads to a
decrease in breastfeeding, the assumption was reduced to
3.5 months in 1976 and 3 months in 1980 in the national-
level estimates of proximate determinants that follow.
A Speculative Overview of Trends in the Proximate
Determinants of Total Fertility Rates, 1970-80
Table 15 links estimates of the total fertility rate to
the total fecundity rate (assumed to be 15.3 births per
woman) ~ using the evidence on contraception, abortion,
and postpartum amenorrhea presented in the previous three
sections on Bongsarts' indices for each of these proximate
determinants It also incorporates the information on
marriage rates discussed earlier in this chapter. Two
variants are preseOnted for each of the three observation
points: one assumes the lower value in ache range of
estimates of the total abortion rate, with higher levels
of contraceptive use (lower levels of the index of non-
contraception); the other assumes lower contraceptive
TABLE 15 Estimates of Proximate Determinants of Total
Fertility Rate, 1970-80: Braz il
1970 1976 1980
Measure and
variant
A B A B A B
Assumptione
Percent users (U) 0032 0.25 0.47 0.~l 0052 Oe`46
Effectiveness (E) 0 0 80 0 0 80 0.86 0.86 0.88 0.88
Sterilization factor lelO loll 1~125 1~125 10125 1~125
Months of infecundity 4~0 4.0 3.5 3.S 3~0 3.0
Tom abortion rate 0.5 1.5 0.5 1.5 0.5 1.5
Besultea
Total fecundity rate 15.3 15.3 15.3 15.3 15.3 15.3
Infecundity index ~ .89 0.89 0.91 0.91 0. 93 0 0 93
Natural Bar ital fertility 13 0 6 13. 6 13 0 9 13. 9 14 . 2 14. 2
Abortion index Q. 96 0.88 0.94 0.84 0 O 93 0.82
Contraception fertility 0.72 0.78 0.54 0.60 0.49 0~55
Total ~aarital fertility 9.34 9. 39 7 e 04 7 0 03 6 ~ 42 Be 42
Non~rriage index 0.63 0.63 0.63 0.63 0.64 0.64
Total Fertility Rate 5.89 5.92 4.43 ~ . 43 4.11 4.11
aData in se<:ond panel have been rounded s total fertility rate based on data
before rounding.
OCR for page 57
57
use, and the higher estimate of the total abortion rate.
This is only one of several tradeoffs that a range of
values in the abortion rate implies. Higher abortion
rates could also imply less effective contraception, as
well as ~ lower impact of nonmar r iage to the extent that
abortions are used to terminate the pregnancies of women
outside of unions.
The underlying assumptions are summarized in the first
panel of the table. The contraceptive use and effective
ness rates in van iant #A. are based on Table 12 0 The
reduction in the contraceptive use rate in variant ABE is
proportional to the increase in the abortion index implied
by the higher total abortion rate. Both variants assume
that there is a decline in postpartum infecundity of about
0.5 months from 1970 to 1976 that relates to increased
contraceptive use, with a similar decline for 1976 to
1980. No change is assumed between 1970 and 1976 in the
index of nonmarriage, which is consistent with the
observed trend for that period, while a small increase is
shown for 19800
In variant ~A., the main factor accounting for the
decline in the total fertility rate from its level of
around 5.9 in 1970 to 4.4 in 1976 is the decrease in
Cc; this reflects increased contraceptive use, as well
as increased effectiveness deriving from a higher propor-
tion of more effective methods (the pill and steriliza-
tion) among those used. To the extent that variant ~A.
understates the abortion rate, this conclusion should be
modified by assuming a smaller increase in contraceptive
use, less improvement in contraceptive effectiveness, or
a lower level of nonmarriage. variant ABE illustrates
the first of these possibilities.
Extrapolating to 198G, the table suggests that the
pace of the decline in total fertility could be slower in
the last half of the decade.
This is partly a result of
using levels of Cc in 1976 large enough to account for
the 1970-76 decline, combined with Cc values for 1980
that are consistent with the survey results reported in
Table 11. Variant ABE suggests an even slower decline in
Cc, assuming that abortions played a greater role in
the decline in total fertility than is implied in variant
~A. ~ It is worth noting that special tabulations of the
1977 and 1978 PNAI) survey data in fact suggest a slowing
of the pace of the decline of total fertility (l~eite,
1981). An alternative hypothesis is that flange in the
design of the 1976 survey produced an overstatement of
the decline, a point that will prove difficult to test
OCR for page 58
58
until data for making alternative estimates of fertility
during the decade become available (ye., 1980 census
data, for which the own-children method could be applied).
CONCLUSIONS
.
The 1970s brought a significant acceleration in the
decline of fertility in Brazil: the crude birth rate
fell from 41 per 1~000 in the late 1960s to 34 per 1,000
in the late 1970s, the total fertility rate, a more
refined measure of fertility, declined from around 508
births per l,OOO women around 1970 to around 404 births
by 1976. Of the three demographic factors that could
account for a decline in the birth rate (proportions
married, age structure, and marital fertility), declining
marital fertility was clearly responsible for the change.
Assessment of the effect of nuptiality on fertility is
complicated by the questionable reliability of reported
marital status in the availat~le census and survey data.
A number of factors suggest that the proportion of younger
women (up to age 25) reported as married has been under-
stated. Doubts about the proportions married remain even
after adjustments to include single women reporting births
and separated women who might have entered into a second
union and did not report it because divorce was not
legally recognized. The whole question of pregnancies
to women not in unions and the termination of such preg-
nancies remains a major area of doubt in examining the
prox ~ te determinants of Brazil's fertility decline.
Calculations of the singulate mean age at marriage (SMAM)
from census and survey data indicate a rise of about 0.8
years in the mean age at marriage between 1960 and 1970,
and about half that from 1970 to 1976. An alternative
measure of Me age at which women begin being exposed to
the risk of childbearing, the singulate mean age at first
birth, reveals a similar trend. Even after adjustment of
the marital status data to account for possible undero
reporting of the proportions married among women in their
early twenties, there is little evidence of change between
1970 and 1976; this suggests that changes in marital
status had only a limited effect on fertility decline
during that interval. Because of the demographic echo of
increased births in the l9SOs and 1960s, the age structure
of the Brazilian population had a slightly positive impact
on birth rates.
OCR for page 59
59
Thus, on the basis of an admittedly speculative recon-
s traction of What could have happened. when Brazil' s
accelerated fertility decline is decomposed into demo-
graphic components and into the proximate determinants of
fertility, it can be concluded that a decline in marital
fertility was the primary factor responsible. This
decline in marital fertility can in turn be traced to an
increased use of effective contraception, combined with
an indeterminate abortion component both within and
outside of marriage. Though national-level data on the
proximate variables are lacking, survey data can be used
to construct a nat$onal-level index of noncontraception,
which declined from . 72-0 78 in 1970 to . 54-. 60 in 1976
and ·49.O55 in 1980, suggesting that increased contracep~
tion played a major role in the decline. Survey data also
indicate a very low prevalence of breastfeeding, and
suggest a moderate attenuation of the fertility-reducing
effect of postpartum amenorrhea as contraception
increased. The major unknown variable is abortion. The
more one accepts the f ragments of evidence that abortion
is widespread in Brazil, particularly among low~income
g roups, the more one must ad just the importance attached
to increased contraception and no change in marriage
r ates.
Finally, it should be emphasized that doubts remain
about the reliability of data on marital status, particle
larly since a significant number of pregnancies mung
women not in unions may be terminated by abortion, and
informal unions may be formalized only after a live birth.
Census and survey data on the proportion of women in
unions may therefore reflect after-the-fact social adjust-
ment processes more than ache exact demographic accounting
needed to decompose the proximate determinants of
fertility.
Representative terms from entire chapter:
marital status
