Embed Size (px)
Citation preview
Legal abortion and placenta previa
David A. Grimes, M.D., and Therese Techman, B.A.
Atlanta, Georgia
Legal abortion has been postulated to be a risk factor for placenta previa in subsequent pregnancies. To examine this hypothesis, we analyzed the deliveries of 28,665 women. We identified 68 women who had had placenta previa and compared their obstetric histories with those of 68 controls randomly selected from the same group of deliveries. The crude risk ratio for women with a history of one or more legal abortions was 1.4 (95% confidence interval, 0.5 to 3.6; p > 0.05). Standardizing the crude risk ratio for the effects of age and gravidity reduced the risk ratio to 1.1 (95% confidence interval, 0.4 to 2.8). In this predominantly black population legal abortion does not appear to have a significant association with placenta previa in subsequent pregnancies. (AM. J. OBSTET. GYNECOL. 149:501, 1984.)
Induced abortion has been postulated to be a risk factor for placenta previa in subsequent pregnancies. Retrospective cohort studies in the United States1
• 2 have
noted an association between "prior abortion" (type unspecified) and placenta previa, whereas more sophisticated epidemiologic studies in the United States3
• 4 and
IsraeJ5 have not. A recent retrospective cohort study from the United States6 reported that the risk of placenta previa is six times greater for women with a history of induced first-trimester abortion than it is for women with no such history. The authors suggested that scarring of the endometrium from "vigorous uterine curettage" may lead to an abnormal site of placental implantation in subsequent pregnancies.
Since more than one million legal induced abortions are performed in the United States each year7 and since placenta previa is a serious complication of pregnancy, the public health importance of an association, if there is any, between legal abortion and placenta previa would be substantial. However, in this study we found no significant association between these two events.
Method
This case-control study was performed at Grady Memorial Hospital in Atlanta, Georgia. The hospital is the primary health care provider of both inpatient and outpatient care for the indigent population of metropolitan Atlanta; it also serves as a tertiary care center for high-risk obstetric patients in northwest Georgia.
From the Division of Reproductive Health, Center for Health Promotion and Education, Centers for Disease Control, Department of Health and Human Services, Public Health Service, and Department of Gynecology and Obstetrics, Emory University School of Medicine.
Received for publication September 15, 1983; revised November 23, 1983; accepted December 28, 1983.
Reprints will not be available.
Cases of placenta previa were identified through a computerized obstetric record system. The study included all births of infants who weighed ;:o:SOO gm and who were born at the hospital between january 1, 1975, and December 31, 1979. Cases of placenta previa were defined as those involving (1) complete placenta previa (documented by ultrasonography or double setup examination) that necessitated cesarean delivery of the infant or (2) placenta previa of any degree (documented by ultrasonography) that led to bleeding requiring hospitalization during pregnancy. This definition of placenta previa excluded asymptomatic cases of placenta previa that resolved spontaneously as pregnancy progressed and that were clinically unimportant.8
• 9
Controls were identified through the computerized record system. Potential controls were all women without placenta previa who were delivered of infants weighing ;:o:SOO gm at the hospital during the study interval. Selection of controls was made by computer with the use of a random-digit program.
Exposure data concerning obstetrit: and gynecologic events were collected for cases and controls in a uniform manner. A trained abstractor who was unaware of the purpose of the study reviewed the computerized records and hospital charts of all cases and controls. Information about cases and controls was transcribed onto a data collection form designed for the study.
Data gathered included demographic information, history of gynecologic or abdominal operations, contraceptive history, information about any transfer from outside the hospital's catchment area, and complete obstetric history (including dates, gestational ages, outcomes, facility, associated operations, and presence or absence of fever).
The primary source of information about obstetric history in hospital charts was a detailed questionnaire completed by a trained interviewer at the time of the first prenatal visit. Thus information about obstetric history was collected in a uniform fashion for most pa-
501
502 Grimes and Techman
Table I. Distribution of cases of placenta
previa and controls by age, race, and gravidity
Cases Controls
Characteristic No. I % No. I %
Age (yr) :s 19 8 ll.8 23 33.8 20-29 40 58.8 39 57.4 2:30 20 29.4 6 8.8
Total 68 100.0 68 100.0
Race Black and other 50 73.5 55 80.9 White 18 26.5 13 19.1
Total 68 100.0 68 100.0
Gravidity* 0 9 13.2 26 38.2 I 10 14.7 14 20.6 2 12 17.6 10 14.7 2:3 37 54.4 18 26.5
Total 68 100.0 68 100.0
*Prior to index pregnancy.
tients before evidence of placenta previa appeared. These data were supplemented by other medical records from the abortion service, emergency room, and
delivery suite. During the study 28,665 infants weighing ;:>:500 gm
were born at the hospital. The computerized record system identified 97 births (3.4 per 1000) complicated by placenta previa. We reviewed the charts of 83 of the 97 women; 14 charts could not be located after three or more attempts. Fifteen patients of the 83 whose charts were located failed to meet the case definition and were excluded; 68 cases were included in the study. For comparison, we used the first 68 charts retrieved from the list of randomly selected potential controls.
After data collection was completed, we attempted to validate the obstetric histories through an independent data source. From 1967 to 1976, records of all family planning visits made to any of five health agencies in metropolitan Atlanta were entered into a central computerized data system. 10 The five agencies (Emory University Family Planning Program at Grady Memorial Hospital, Planned Parenthood of Atlanta, Fulton County Health Department, DeKalb County Health Department, and Atlanta Southside Comprehensive
Health Center) operated 44 clinics for medically indigent patients at 31 locations in Fulton and DeKalb counties. Grady Memorial Hospital patient numbers were used for individual patient identification in the data system of the family planning services. By linking hospital numbers and codes for race, we were able to compare information on obstetric history available from hospital records and from family planning clinics outside the hospital.
July 1, 1984 Am. J. Obstet. Gynecol.
We calculated standardized odds ratios and ">f values by the method of Mantel and Haenszel with the use of a programmable calculatorY Standardized to eliminate
the distorting effect of potentially confounding factors, these ratios gave an estimate of the risk of placenta previa for a woman with a history of one or more legal abortions relative to the risk for a woman with no such
history. This procedure was used to standardize ratios for the
following potentially distorting effects: age; race; marital status; gravidity; parity; history of spontaneous abortion, cesarean delivery, sharp curettage, gynecologic or abdominal surgery, or fever after pregnancy termination; last-used method of contraception; and transfer from another facility. We calculated a 95% confidence interval for the standardized risk ratios. 11
When the interval does not overlap 1.0, the results are significant at the traditional 0.05 level.
Results
Women with placenta previa were generally older and of higher gravidity than were controls (Table I). In both groups, most women were black or of other minority races. A higher percentage of cases (47%) were married than were controls (28%). Likewise, a higher percentage of cases ( 10%) were transferred from other facilities than were controls ( 1% ). A higher percentage of cases (29%) reported having used a method of contraception prior to the index pregnancy than did controls (18%).
Overall, women with a history of one or more legal
abortions did not have a significantly increased risk of placenta previa. Twelve women with placenta previa
(18% of the cases) and nine controls (13%) reported having had a legal abortion. Compared with women with no history of legal abortion, women who had undergone legal abortions had 1.4 times the risk of placenta previa (95% confidence interval, 0.5 to 3.6; p > 0.05). Among women with a history of legal abortion, the pregnancy immediately before the index pregnancy was terminated by legal abortion for eight cases (67%) and for eight controls (89%); one case (8%) and three controls (33%) had had more than one legal abortion.
Standardizing the crude risk ratio simultaneously for the effects of age and gravidity reduced the risk ratio to 1.1 (95% confidence interval, 0.4 to 2.8) (Table II). (Women who had not been pregnant prior to the index pregnancy were excluded from this analysis, since they
cannot have a history of legal abortion.) Standardizing the risk ratio separately for each of the other variables cited, including transfer from another facility, failed to demonstrate a significant association between legal abortion and placenta previa. When patients who were transferred from other facilities were excluded from
Volume 149 Number 5
analysis, the crude risk ratio was 1.6 (95% confidence
interval, 0.6 to 4.1 ). On the other hand, a history of one or more spon
taneous abortions was associated with a crude risk ratio
of 3.2 (95% confidence interval, 1.4 to 7.2). However,
when standardized for the effect of age or gravidity,
this association was no longer statistically significant.
When the total numbers of cases and controls with a
history of uterine curettage (suction or sharp curettage
related to a pregnancy and diagnostic curettage not
associated with pregnancy) were compared, the crude
risk ratio associated with uterine curettage was 1.9
(95% confidence interval, 0.9 to 3.9). Again, no statisti
cally significant association between curettage and pla
centa previa was evident when the risk ratio was stan
dardized for the effect of age or gravidity.
The attempt to corroborate the obstetric histories
through the data system of the family planning services
linked the hospital and clinic records for 34 cases and
35 controls. For the 34 cases the family planning rec
ords revealed a total of two live births and two preg
nancies of unknown outcome not identified through
the hospital records; however, one spontaneous and
one legal abortion identified through hospital records
were not reported in the family planning records. For
the 35 controls the family planning records revealed
one pregnancy of unknown outcome not identified
through the hospital records, whereas one legal abor
tion documented in the hospital records was not re
ported in the family planning records. If the three
pregnancies with unknown outcomes are each consid
ered to be a legal abortion, the revised risk ratio be
comes 1.5 (95% confidence interval, 0.6 to 3.7).
Comment
The major finding of this study was that women at
this hospital with a history of one or more legal abor
tions did not have an increased risk of symptomatic
placenta previa in subsequent pregnancies. This study
analyzed more than 28,000 deliveries among women
predominantly of minority races and low socioeco
nomic status. To what extent these data can be extrapo
lated to other populations is unclear.
Selection bias is unlikely to have influenced these re
sults. All patients identified as having had placenta
previa during the study interval were reviewed for in
clusion in the study. All obstetric admissions to the
hospital were entered into the computerized record
system, and it was unlikely that bleeding because of
placenta previa or cesarean delivery necessitated by it
would have escaped being coded as placenta previa.
Berkson's bias, a potential bias in most hospital-based
case-control studies, should not have been operative,
since nearly all pr,egnant women in the population
served by this hospital give birth at the hospital.
Legal abortion and placenta previa 503
Table II. Distribution of cases and controls by
history of legal abortion, age, and gravidity*
Stratum
Age :s 19, gravida I Cases Controls
Age :s 19, gravida ~2 Cases Controls
Age 20 to 29, gravida I Cases Controls
Age 20 to 29, gravida ~2 Cases Controls
Age ~30, gravida l Cases Controls
Age ~30, gravida ~2 Cases Controls
History of ~1 legal abortion
Yes
2 I
I 0
2 I
4 7
0 0
3 0
No
2 4
0 2
3 7
26 14
15 5
*Excludes women with no pregnancies prior to index pregnancy; age- and gravidity-standardized risk ratio = I. I (95% confidence interval 0.4 to 2.8).
.Fourteen charts of potential cases could not be lo
cated after repeated attempts; however, our success
rate in locating charts was higher than that in other
studies conducted in this hospital. 12 Moreover, it is un
likely that any systematic bias would have influenced
the ability to retrieve charts. All deliveries of infants
weighing 2:500 gm and not complicated by placenta
previa were considered potential controls, and a com
puter-generated, random-number sequence was used
to select controls from this pool.
Ascertainment bias can be an important limitation of
any case-control study in which exposure information
must be gathered from existing records. In pregnancy
cohort studies, the later in pregnancy a history of pre
vious abortion is determined, the more likely is an ad
verse outcome to affect selective recall. 13 In this study
the obstetric history of most patients was obtained at
the first prenatal visit by using a standard ques
tionnaire; this approach should have limited the prob
lem of selective recall. When patients who were trans
ferred from other facilities were excluded from analysis
in order to control for potential ascertainment bias, the
relative risk changed little. In addition, since the ab
stractor had no knowledge of the hypothesis in question,
it could not have influenced the chart review.
Underreporting of previous abortions is another po
tential problem in studies of this type. To reduce the
incidence of this type of error, we compared obstetric
histories as recorded in two independent record sys
tems. We were able to link the records of about half of
504 Grimes and Techman
the cases and controls, and the overall correlation was
high. Although use of multiple data sources may re
duce the bias of false negative misclassification, 13 it does
not eliminate this problem altogether.
Although some potentially confounding variables
were considered in this analysis, others could not be
adequately evaluated. Cigarette smoking appears to be
an important risk factor for placenta previa, 14 yet reli
able smoking histories could not be obtained from the
hospital records. Several studies have found that
women with a history of abortion are more likely to
smoke during pregnancy. 13
It is unlikely that our findings are due to chance.
With 97.5% certainty we can rule out risks larger than
2.8 (standardized for age and gravidity). Thus this
study was unlikely to have missed identifying a strong
association between legal abortion and placenta previa.
The lack of association between legal abortion and
placenta previa in this study is consistent with the
findings of other large studies3- 5 that have addressed
this question. In addition, this study does not support
the hypothesis that uterine curettage increases the risk
of placenta previa by damaging the endometrium. Our
study suggests that having had one or more legal abor
tions does not increase the risk of symptomatic placenta
previa in subsequent pregnancies.
William Graves, Ph.D., Kenneth F. Schulz, M.B.A.,
and Carol J. Hogue, Ph.D., provided technical assistance.
REFERENCES
1. Brenner WE, Edelman DA, Hendricks CH. Characteristics of patients with placenta previa and results of "expec-
Bound volumes available to subscribers
July 1, 1984 Am. J. Obstet. Gynecol.
tant management." AM J OssTET GYNECOL 1978;132: 180-9.
2. Cotton DB, ReadJA, Paul RH, Quilligan EJ. The conservative aggressive management of placenta previa. AM J 0BSTET GYNECOL 1980;137:687-95.
3. Schoenbaum SC, Monson RR, Stubblefield PG, Darney PD, Ryan KJ. Outcome of the delivery following an induced or spontaneous abortion. AM J 0BSTET GYNECOL 1980;136:19-24.
4. Madore C, Hawes WE, Many F, Hexter AC. A study on the effects of induced abortion on subsequent pregnancy outcome. AM J 0BSTET GYNECOL 1981;139:516-21.
5. Harlap S, Davies AM. Late sequelae of induced abortion: Complications and outcome of pregnancy and labor. Am J Epidemiol 1975;102:217-24.
6. Barrett JM, Boehm FH, Killam AP. Induced abortion: A risk factor for placenta previa. AM J 0BSTET GYNECOL 1981; 141:769-72.
7. Centers for Disease Control: Abortion surveillance 1978. Atlanta, November, 1980.
8. Rizos N, Doran TA, Miskin M, Benzie RJ, FordJA. Natural history of placenta previa ascertained by diagnostic ultrasound. AMJ 0BSTET GYNECOL 1979;133:287-91.
9. Tremewan RN, Chandra M, Duff GB. The mid-trimester low lying placenta: A prospective study. N Z Med J 1980;92: 151-3.
10. Smith JC, Goldsby JB. A family planning services data system. Fam Plann Perspect 1970;2:41-6.
11. Rothman KJ, Boice JD Jr. Epidemiologic analysis with a programmable calculator. Bethesda, Maryland: National Institutes of Health, 1979; NIH publication No. 79-1649.
12. Grimes DA, Ross WC, Hatcher RA. Rh immunoglobulin utilization after spontaneous and induced abortion. Obstet Gynecol 1977;50:261-3.
13. Hogue CJR, Cates W Jr, Tietze C. The effects of induced abortion on subsequent reproduction. Epidemiol Rev 1982;4:66-94.
14. Naeye RL. Abruptio placentae and placenta previa: Frequency, perinatal mortality, and cigarette smoking. Obstet Gynecol 1980;55:701-4.
Bound volumes of the AMERICAN JOURNAL OF OBSTETRICS AND GYNECOLOGY are avail
able to subscribers (only) for the 1984 issues from the Publisher, at a cost of $70.95
($90.00 international) for Vol. 148 (January-April), Vol. 149 (May-August), and Vol. 150
(September-December). Shipping charges are included. Each bound volume contains a
subject and author index and all advertising is removed. Copies are shipped within 30
days after publication of the last issue in the volume. The binding is durable buckram with
the JouRNAL name, volume number, and year stamped in gold on the spine. Payment must accompany all orders. Contact The C. V. Mosby Company, Circulation Department, 11830
Westline Industrial Drive, St. Louis, Missouri 63146, USA; phone (800) 325-4177, ext. 351.
Subscriptions must be in force to qualify. Bound volumes are not available in place of
a regular JOURNAL subscription.
