- Email: [email protected]
Anemia and spontaneous preterm birth
Anemia and spontaneous preterm birth Mark A. Klebanoff, MD, MPH,. Patricia H. Shiono, PhD,.·b Joseph V. Selby, MD, MPH: Alan I. Trachtenberg, MD, MPH: and Barry I. Graubard, MAe Bethesda, Maryland, and Oakland, California The association between anemia during pregnancy and spontaneous preterm birth was studied with a two-stage case-control design in a large, multiethnic cohort. Results of all hematologic measurements were abstracted from the prenatal and delivery records of 1706 of the 26,901 women in the cohort. Among women delivered of infants at term, mean hematocrit value was low during the early phase of the second trimester, stable until near term, then reached a maximum at 40 weeks' gestation. The mean hematocrit value of black women was consistently lower than that of Asian, Mexican, and white women. Anemia (hematocrit value less than the tenth percentile for ethnic group and duration of pregnancy) at any time during the second trimester was positively associated with subsequent spontaneous preterm birth (odds ratio, 1.9; 95% confidence interval, 1.3 to 2.8). Compared with white women, the odds ratios for preterm birth were 2.0 (95% confidence interval, 1.6 to 2.4) for black, 1.2 (95% confidence interval, 0.9 to 1.6) for Asian, and 1.2 (95% confidence interval, 1.0 to 1.5) for Mexican women. Adjustment for second-trimester anemia had minimal influence on the odds ratios. We conclude that anemia during the second trimester was associated with preterm birth. However, it does not account for the large ethnic differences in preterm birth. (AM J OasTET GVNECOL 1991 ;164:59-63.)
Key words: Anemia, hematocrit, pregnancy, premature labor
Anemia, as measured at the time of delivery, has been associated with spontaneous preterm birth.I.2 The study reported that a hematocrit value of 34 or less was strongly associated with preterm birth and that there was a dose-response relationship between hematocrit value and preterm birth. Anemia was found to account for 60% of the observed differences between black and white women in the occurrence of spontaneous preterm birth. I These findings were later questioned in a subsequent study' that showed that because the hematocrit value rises during the third trimester of a normal pregnancy, comparison of delivery hematocrit values in women giving birth to term and preterm infants produces a spurious association between anemia and preterm birth. The purpose of the present study is to explore further the association between hematocrit measured at various gestational ages before delivery and duration of pregnancy. To accomplish this, a caseFrom the Division ofPrevention Research, National Institute ofChild Health and Human Development," the Center for the Future ofChildren. David and Lucile Packard Foundation,' the Permanente Medical Group, Division of Research,' Bay Area Addiction Research and Treatment/ and the Biometry Branch, Division ofCancer Prevention and Control, National Cancer Institute.' The Kaiser Permanente Birth Defects Study was supported by contract number NOI-HD-286I from the Center for Population Research. National Institute of Child Health and Human Development. Received for publication June 7. 1990; revised August 7, 1990; accepted August 15. 1990. Reprint requests: Mark A. Klebanoff, MD, Division of Prevention Research, National Institute of Child Health and Human Development, National Institutes ofHealth, EPN 640, Bethesda, MD 20892.
6/1/24741
control study" 5 was carried out within a large, multiethnic pregnancy cohort. Methods
The data for this study are from the northern California Kaiser Permanente Birth Defects Study. Detailed methods of this prospective cohort study are published elsewhere. 6 Women who registered for prenatal care at the 13 facilities that served northern California during 1974 to 1977 completed a self-administered questionnaire in English or Spanish that obtained information on a variety of demographic, reproductive, and behavioral variables. Gestational age was determined on the basis of the menstrual history and physical examination at the first prenatal visit. Preterm delivery was defined as a live birth of 24 to 36 completed weeks' gestation. There were 36,504 women initially recruited; exclusion of women found not to be pregnant, women who had induced abortions, women who wanted the infant to be adopted, women not delivered of infants at a Kaiser hospital, and women lost to follow-up resulted in outcome data for 34,660 babies. Women delivered of a live-born singleton infant of at least 24 weeks' gestation and 500 gm were included in these analyses (n = 31,682). Repeat study pregnancies by the same women were excluded (n = 1086), as were women who started prenatal care after 24 weeks' gestation (n = 2266) and women of unknown race (n = 1429). The final sampling frame contained 26,901 women. In the original study, data on hematocrit values and whether delivery was induced were not collected. So
59
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Table I. Derivation of total number of births Preterm Full cohort Sampled Known induction or no labor Labor status unknown Known spontaneous preterm labor and I or more hematocrit values present (cases)* Estimated number of spontaneous preterm births in full cohort* Term Full cohort* Sampled One or more hematocrit values present (controls)*
Asian
Black
Mexican
White
152 152 17 41 94
306 306 55 53 198
243 243 35 49 159
1194 401 57 70 274
129t
239t
199t
Inl 152 140
2228 306 279
2538 243 200
988t 18469 401 362
Total 1895 1102 164 213 725 1555 25006 1102 981
*Values were used to adjust the odds ratios for ethnicity and preterm birth and the adjustments to their variances. tDerived by multiplying the quotient of the number of women with known spontaneous preterm labor with one or more hematocrit values (a), divided by the sum of the number of women with known spontaneous preterm labor with one or more hematocrit values (a) and the number of women with known induction or no labor (b), times the number of women in the full cohort (c): [a/(a + b)J x c. that these data could be gathered by review of the records of a manageable number of preterm and term pregnancies, a two-stage case-control design was used!' 5 All preterm infants born to Asian (n = 152), black (n = 306), and Mexican (n = 243) women were selected, as was a simple random sample of 401 of the ] 194 preterm infants born to white women. Within each ethnic group, one control woman delivered of an infant after 36 weeks' gestation was selected at random for each case. Hematocrit data were abstracted from both the inpatient and outpatient records; data on induction of labor were abstracted from the inpatient records. Of the 2204 women in the sample, charts for 334 were not available, primarily because of off-site storage and personnel shortages at certain facilities. Therefore women with missing charts are unlikely to differ from the remainder of the sample. Inasmuch as spontaneous preterm birth was the outcome of interest, preterm infants born after induced labor or cesarean section in the absence of labor were excluded (n == 164), which reduced the number of records to 1706. The final sample analyzed consisted of women who had spontaneous preterm labor and one or more hematocrit values (n = 94, Asian; 198, black; 159, Mexican; 274, white), or term delivery and one or more hematocrit values (n = 140, Asian; 279, black; 200, Mexican; 362, white). Details of the sampling procedure are shown in Table I. All measures of hematocrit or hemoglobin and the date they were obtained were abstracted from the records. Data on the source of blood (venous or finger stick), the method used to calculate hematocrit (centrifuged or Coulter counter) and the cause of anemia (iron deficiency or other) were not obtained. The ges-
tational age at each measurement was determined by subtracting the date of the last menstrual period or physician's estimate of the last menstrual period from the date the blood was obtained. On average, each woman had 2.6 measures of hematologic status; ]2% had one measure, 31 % had two, 36% had three, 17% had four, and 4% had five or more. In those instances in which only a hemoglobin value was available, it was converted to an equivalent hematocrit value by multiplying by three. Our definition of anemia varied slightly depending on the specific analysis. When the relationship between anemia and subsequent preterm birth was studied, ethnic group-specific definitions of anemia were used to separate the effect of anemia from that of ethnicity. Thus for each ethnic group anemia was defined as a hematocrit value of less than the tenth percentile for women of that ethnic group during each 2-week interval of gestational age on the basis of the measurements of pregnancies that ultimately ended at term. The distribution of hematocrit values was similar for all nonblack women; therefore these groups were combined. The distribution of hematocrit values among black women was shifted toward lower values. Dependent on the specific interval, the tenth percentile of hematocrit was 31 to 33 for nonblack and 29 to 31 for black women. When the role of anemia in explaining the ethnic differences in preterm birth was studied, anemia was defined as the tenth percentile for all term infants in the sample, weighted for the unequal sampling probabilities of each ethnic group. If different tenth percentiles were used for each ethnic group, the prevalence of anemia would not differ between ethnic groups and adjustment for anemia could not effect the relationship between ethnicity and preterm birth.
Anemia and preterm birth
Volume 164 Number 1, Part 1
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40 39 M E
A N H E
~
38
36 35
C
34
I
33
R
T
non-Black
37
M A
T 0
.
'
* Black
32 31
15-16 17-18 19-20 21-22 23-2425-2627-2829-3031-3233-3435-3637-3839-40
WEEKS GESTATION Fig. 1. Mean hematocrit value at each 2-week interval of gestation among pregnancies delivered at term,
Statistical methods. As described, all Asian, black, and Mexican preterm infants were studied, as were approximately one third of the white preterm infants. Term infants were selected as controls from each ethnic group according to the number of preterm cases from that group. Because the sampling probabilities were known and differed by outcome and ethnic group, odds ratios for ethnicity could be calculated by adjusting for the different sampling fractions according to the method described by Cain and Breslow.' This method adjusts both the odds ratios and the SEs derived from logistic regression for unequal sampling probabilities. To compare our results with those previously described,' the change in hematocrit value during a normal pregnancy that resulted in a term infant was evaluated. The mean hematocrit value at each 2-week interval during the second and third trimesters for women delivered of infants of >36 weeks' gestational age was studied. Next, we evaluated the association between anemia as determined at each 2-week interval of gestational age and subsequent preterm birth. In addition, the occurrence of preterm birth was compared between women who were anemic at any time during the second trimester and women who had hematocrits measured during the second trimester and were not anemic. This analysis was done for all women and separately for women in each ethnic group. Finally, we evaluated the role of anemia in explaining the ethnic differences in preterm birth. The odds ratios were calculated for preterm birth to Asian, black, and Mexican women compared with white women. The contribution
of differing rates of anemia to the ethnic differences in preterm birth was assessed by comparison of the odds ratios before and after adjustment for the presence of anemia at any time during the second trimester. Logistic regression was used for these analyses. Results
The mean hematocrit value at each 2-week interval of gestation during the second and third trimesters is presented in Fig. 1 for black and nonblack women delivered of infants at term. Hematocrit values appeared to fall during the early second trimester, reach a nadir at approximately 26 to 28 weeks' gestation, begin to increase again at 31 to 33 weeks, then increase sharply at term, reaching a maximum at 40 weeks' gestation. Throughout the second and third trimesters, hematocrit values were approximately 2% lower among black than among nonblack women. Because hematocrit values varied widely by gestational age, the relationship between preterm delivery and anemia was evaluated at 2-week intervals of gestational age (Table 11). Odds ratios for preterm birth and anemia were adjusted for ethnicity, age, education, marital status, smoking, and gestational age at first prenatal visit. The odds ratios for anemia during the period from 13 to 26 weeks' gestation were> 1.0, indicating an increased risk of preterm delivery among anemic women. However, anemia after 26 weeks' gestation generally had an inverse relationship with preterm birth. Having been anemic at any time during the second trimester (weeks 13 to 26) was positively asso-
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January 1991 Am.J Obstet Gynecol
Table II. Odds ratios for preterm delivery and anemia by gestational age Weeks' gestation
No.
Adjusted odds ratio*
13-14 15-16 17-18 19-20 21-22 23-24 25-26 27-28 29-30 31-32 33-34 35-36
284 199 148 163 197 203 163 218 267 408 328 391
1.4 2.1 4.3 1.4 2.0 1.4 3.5 0.8 1.4 0.5 0.7 0.5
95% CI
0.5-3.7 0.6-7.4 1.2-15.5t 0.4-5.4 0.8-4.9 0.5-3.7 1.1-l1.3t 0.3-2.1 0.6-3.3 0.1-1.5 0.3-1.6 0.2-1.3
*Adjusted for ethnicity, maternal age, education, marital status, smoking, gestational age at first prenatal visit. tp < 0.05.
ciated with preterm birth [odds ratio, 1.9; 95% confidence interval (CI) 1.2 to 3.0]. The odds ratios by ethnic group were 2.6 (95% CI, 0.7 to 9.6) for Asian women, 1.8 (95% CI, 0.9 to 3.6) for black women, 2.0 (95% CI, 0.8 to 4.7) for Mexican women, and 1,8 (95% CI, 1.0 to 3.5) for white women. Anemia at any time during the third trimester was not analyzed because women delivered of infants at term had more opportunity than did women delivered of infants before term to have anemia diagnosed. This does not apply to the biweekly analyses in Table II because only women who were still pregnant during that interval could have their hematocrit measured. The role of anemia in the explanation of the ethnic differences in preterm birth is shown in Table III. The unadjusted odds ratios for preterm delivery among Asian, black, and Mexican women, compared with white women were 1.2, 2.0, and 1.2, respectively. Despite the association between second-trimester anemia and preterm birth, adjustment for the presence of anemia at any time during the second trimester had little effect on the ethnic differences in preterm delivery. Anemia after 26 weeks' gestation was negatively associated with preterm birth; therefore it could not possibly account for the observed ethnic differences.
Comment We found that in pregnancies carried to term, the mean hematocrit value rises dramatically after 37 weeks' gestation. This is in agreement with earlier reports,S and with the standards developed by the Centers for Disease Control for hematocrit values during pregnancy.7 Because the same group of women contributed data over most of pregnancy, our study has advantages over purely cross-sectional data in which entirely different women may have had blood drawn during each interval. However, our data are not fully longitudinal; every woman did not have an hematocrit drawn during
every interval. To the extent that women who had blood drawn during any given interval differed from those who did not, our results will be in error. Our results agree closely with established standards'? and are consistent with the observation that plasma volume is relatively constant after 34 weeks' gestation but red cell mass increases linearly throughout pregnancy.8 This suggests that the lack of regular measurements taken longitudinally on each woman was unlikely to influence the results. Our results confirm those of Klebanoff et al.,' who found that when measured prospectively, thirdtrimester anemia was not a risk factor for preterm birth. Our results contradict those of Lieberman et al.,' who found a very strong dose-response relationship between preterm birth and third-trimester anemia (hematocrit value ~34) and that anemia accounted for much of the difference in preterm delivery rates between black and white women. Lieberman et al. I defined anemia with hematocrit values obtained at delivery. Our data suggest that the use of delivery hematocrit values is inappropriate because hematocrit rises sharply at term. Use of delivery hematocrit values will therefore produce a spurious relationship between anemia and preterm birth. We found a moderate relationship between anemia during the second trimester and preterm birth. Anemia during the third trimester tended to be negatively associated with preterm birth. In addition, we found that statistical adjustment for the presence of anemia did not affect the ethnic differences in preterm delivery previously reported in this cohort. 9 The association between second-trimester anemia and preterm delivery is not inconsistent with previous findings on this subject because prior work did not evaluate anemia during the second trimester. 3 Why should second-trimester anemia be associated with preterm delivery when anemia later in pregnancy is not? There is no clear answer. This association has not been reported previously, and it may be an artifact of multiple comparisons. The relationship was consistent within ethnicity, making simple chance less likely. It is possible that a small group of high-risk, severely anemic women had hematocrits measured on multiple occasions. This detection bias would tend to overstate the relationship between preterm birth and anemia. We found that women who had one hematocrit measured during the second trimester were no more likely to be delivered of preterm infants than were women who had none. However, women who had two or more hematocrit measurements during the second trimester were at elevated risk of preterm delivery. In an attempt to reduce detection bias, we analyzed mean and first hematocrit values during the second trimester. The results were unchanged when anemia was defined with the use of the mean of all hematocrit values obtained during the second trimester and with the first hemat-
Anemia and preterm birth
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Table III. Odds ratios and 95% CI for preterm delivery by ethnic group White
Unadjusted Adjustedt
Asian
Odds ratio
Odds ratio
1.0 1.0
1.2 1.2
I
Mexican
Black
95% CI
Odds ratio
0.9-1.6 0.9-1.6
2.0 1.9
I
95% CI 1.6-2.4* 1.5-2.3*
Odds ratio
1.2 1.2
I
95% CI 1.0-1.5* 1.0-1.6
*P < 0.05. tOdds ratio adjusted for the presence of second-trimester anemia.
ocrit value obtained during the second trimester. Nevertheless, the association between anemia during the second trimester preteI'm birth should be considered tentative and needs to be confirmed in other populations. A precise definition of anemia during pregnancy is confounded by various factors such as altitude, ethnicity, use of iron supplements, and changes in plasma volume during pregnancy. Definitions of anemia have ranged from hematocrit values of 534, I to a hemoglobin of <10 gm per deciliter lO (corresponding to hematocrit value of 30), to the Centers for Disease Control's month-specific standards.' Use of either of the two fixed values (534 or <30) produced similar results in our data. Because there is no standard definition of anemia for 2-week intervals during pregnancy, we chose to use the internal standard of the tenth percentile at each 2-week interval, as determined among pregnancies that ultimately resulted in births at term. When evaluating the role of anemia, separate standards were used for black and nonblack women to eliminate ethnicity as a confounder. When evaluating the role of ethnicity in preterm birth, a single standard for anemia was used so as not to overadjust for ethnicity. Rapid changes in red cell mass and plasma volume during pregnancy make interpretation of isolated hematocrit values difficult. Low hematocrit levels may result from inadequate red cell mass or a large expansion of plasma volume. Inadequate expansion of plasma volume, which would result in high hematocrit values, has been associated with an increased risk of a variety of pregnancy complications."·!· Women with normal red blood cell mass and a large expansion of plasma volume would probably be at low risk of adverse outcome despite an apparent anemia. Therefore further studies with clinical determinations of hematocrit value are unlikely to resolve this issue. Anemia in pregnancy should be defined as a function of red cell mass and plasma volume, neither of which are measured in clinical settings. In addition, measured hematocrit is a result of the interaction of physiologic, nutritional, and other life style factors. Despite any observed associations, low hematocrit values may merely be a marker for social and environmental deprivation, rather than being a cause of preteI'm birth. Therefore treatment of anemia alone might not affect the occurrence of spontaneous preterm birth.
In conclusion, anemia during the second trimester was found to be positively associated with spontaneous preterm birth; anemia later in pregnancy tended to be negatively associated with preteI'm birth. Secondtrimester anemia did not contribute to the observed ethnic differences in the occurrence of preteI'm birth. Future studies of anemia during pregnancy should supplement routine clinical data with detailed measures of red cell mass and plasma volume. These additional measurements may clarify the role of anemia in the pathogenesis of preterm birth as well as explain the observed ethnic differences in hematocrit values.
We thank Dr. Sholom Wacholder of the National Cancer Institute for his assistance with the sampling method and Ms. Marilyn Goldhaber for her assistance in the development of this project. REFERENCES I. Lieberman E, Ryan KJ, Monson RR, Schoenbaum SC.
2. 3. 4. 5. 6.
7. 8. 9. 10. II. 12.
Risk factors accounting for the racial differences in the rate of premature birth. N EnglJ Med 1987;317:743-8. Lieberman E, Ryan KJ, Monson RR, Schoenbaum SC. Association of maternal hematocrit with premature labor. AMJ OBSTET GYNECOL 1988;159:107-14. Klebanoff MA, Shiono PH, Berendes HW, Rhoads GG. Facts and artifacts about preteI'm delivery. JAMA 1989; 262:511-5. Breslow NE, Cain KC. Logistic regression for two-stage case-control data. Biometrika 1988;75: 11-20. Cain KC, Breslow NE. Logistic regression analysis and efficient design for two stage studies. Am J Epidemiol 1988;126: 1198-206. Shiono PH, Hadap S, Ramcharan S, Berendes HW, Gupta S, Pellegrin F. Use of contraceptives prior to and after conception and exposure to other fetal hazards. Contraception 1978;20: 105-20. Centers for Disease Control. CDC criteria for anemia in children and childbearing-aged women. MMWR 1989;38: 400-4. Hytten F. Blood volume changes in normal pregnancy. Clin H.ematology 1985;14:601-12. Shiono PH, Klebanoff MA. Ethnic differences in preterm and very preterm delivery. Am J Public Health 1986;76: 1317-21. Cunningham FG, McDonald PC, Gant NF. Williams obstetrics. 18th ed. Norwalk, Connecticut: Appleton and Lange, 1989;779-80. Goodlin RC, Dobry CA, Anderson JC, Woods RE, Quaife M. Clinical signs of normal plasma volume expansion during pregnancy. AMJ OBSTET GYNECOL 1983;145:1001-9. Goodlin RC, Quaife MA, SmithJ). Hypovolemic pregnant women and their risk determinants. IntJ Gyn.ecol Obstet 1983;21:45-50.
Key words: Anemia, hematocrit, pregnancy, premature labor
Anemia, as measured at the time of delivery, has been associated with spontaneous preterm birth.I.2 The study reported that a hematocrit value of 34 or less was strongly associated with preterm birth and that there was a dose-response relationship between hematocrit value and preterm birth. Anemia was found to account for 60% of the observed differences between black and white women in the occurrence of spontaneous preterm birth. I These findings were later questioned in a subsequent study' that showed that because the hematocrit value rises during the third trimester of a normal pregnancy, comparison of delivery hematocrit values in women giving birth to term and preterm infants produces a spurious association between anemia and preterm birth. The purpose of the present study is to explore further the association between hematocrit measured at various gestational ages before delivery and duration of pregnancy. To accomplish this, a caseFrom the Division ofPrevention Research, National Institute ofChild Health and Human Development," the Center for the Future ofChildren. David and Lucile Packard Foundation,' the Permanente Medical Group, Division of Research,' Bay Area Addiction Research and Treatment/ and the Biometry Branch, Division ofCancer Prevention and Control, National Cancer Institute.' The Kaiser Permanente Birth Defects Study was supported by contract number NOI-HD-286I from the Center for Population Research. National Institute of Child Health and Human Development. Received for publication June 7. 1990; revised August 7, 1990; accepted August 15. 1990. Reprint requests: Mark A. Klebanoff, MD, Division of Prevention Research, National Institute of Child Health and Human Development, National Institutes ofHealth, EPN 640, Bethesda, MD 20892.
6/1/24741
control study" 5 was carried out within a large, multiethnic pregnancy cohort. Methods
The data for this study are from the northern California Kaiser Permanente Birth Defects Study. Detailed methods of this prospective cohort study are published elsewhere. 6 Women who registered for prenatal care at the 13 facilities that served northern California during 1974 to 1977 completed a self-administered questionnaire in English or Spanish that obtained information on a variety of demographic, reproductive, and behavioral variables. Gestational age was determined on the basis of the menstrual history and physical examination at the first prenatal visit. Preterm delivery was defined as a live birth of 24 to 36 completed weeks' gestation. There were 36,504 women initially recruited; exclusion of women found not to be pregnant, women who had induced abortions, women who wanted the infant to be adopted, women not delivered of infants at a Kaiser hospital, and women lost to follow-up resulted in outcome data for 34,660 babies. Women delivered of a live-born singleton infant of at least 24 weeks' gestation and 500 gm were included in these analyses (n = 31,682). Repeat study pregnancies by the same women were excluded (n = 1086), as were women who started prenatal care after 24 weeks' gestation (n = 2266) and women of unknown race (n = 1429). The final sampling frame contained 26,901 women. In the original study, data on hematocrit values and whether delivery was induced were not collected. So
59
60
Klebanoff et al.
January 1991 Am J Obslet Gynecol
Table I. Derivation of total number of births Preterm Full cohort Sampled Known induction or no labor Labor status unknown Known spontaneous preterm labor and I or more hematocrit values present (cases)* Estimated number of spontaneous preterm births in full cohort* Term Full cohort* Sampled One or more hematocrit values present (controls)*
Asian
Black
Mexican
White
152 152 17 41 94
306 306 55 53 198
243 243 35 49 159
1194 401 57 70 274
129t
239t
199t
Inl 152 140
2228 306 279
2538 243 200
988t 18469 401 362
Total 1895 1102 164 213 725 1555 25006 1102 981
*Values were used to adjust the odds ratios for ethnicity and preterm birth and the adjustments to their variances. tDerived by multiplying the quotient of the number of women with known spontaneous preterm labor with one or more hematocrit values (a), divided by the sum of the number of women with known spontaneous preterm labor with one or more hematocrit values (a) and the number of women with known induction or no labor (b), times the number of women in the full cohort (c): [a/(a + b)J x c. that these data could be gathered by review of the records of a manageable number of preterm and term pregnancies, a two-stage case-control design was used!' 5 All preterm infants born to Asian (n = 152), black (n = 306), and Mexican (n = 243) women were selected, as was a simple random sample of 401 of the ] 194 preterm infants born to white women. Within each ethnic group, one control woman delivered of an infant after 36 weeks' gestation was selected at random for each case. Hematocrit data were abstracted from both the inpatient and outpatient records; data on induction of labor were abstracted from the inpatient records. Of the 2204 women in the sample, charts for 334 were not available, primarily because of off-site storage and personnel shortages at certain facilities. Therefore women with missing charts are unlikely to differ from the remainder of the sample. Inasmuch as spontaneous preterm birth was the outcome of interest, preterm infants born after induced labor or cesarean section in the absence of labor were excluded (n == 164), which reduced the number of records to 1706. The final sample analyzed consisted of women who had spontaneous preterm labor and one or more hematocrit values (n = 94, Asian; 198, black; 159, Mexican; 274, white), or term delivery and one or more hematocrit values (n = 140, Asian; 279, black; 200, Mexican; 362, white). Details of the sampling procedure are shown in Table I. All measures of hematocrit or hemoglobin and the date they were obtained were abstracted from the records. Data on the source of blood (venous or finger stick), the method used to calculate hematocrit (centrifuged or Coulter counter) and the cause of anemia (iron deficiency or other) were not obtained. The ges-
tational age at each measurement was determined by subtracting the date of the last menstrual period or physician's estimate of the last menstrual period from the date the blood was obtained. On average, each woman had 2.6 measures of hematologic status; ]2% had one measure, 31 % had two, 36% had three, 17% had four, and 4% had five or more. In those instances in which only a hemoglobin value was available, it was converted to an equivalent hematocrit value by multiplying by three. Our definition of anemia varied slightly depending on the specific analysis. When the relationship between anemia and subsequent preterm birth was studied, ethnic group-specific definitions of anemia were used to separate the effect of anemia from that of ethnicity. Thus for each ethnic group anemia was defined as a hematocrit value of less than the tenth percentile for women of that ethnic group during each 2-week interval of gestational age on the basis of the measurements of pregnancies that ultimately ended at term. The distribution of hematocrit values was similar for all nonblack women; therefore these groups were combined. The distribution of hematocrit values among black women was shifted toward lower values. Dependent on the specific interval, the tenth percentile of hematocrit was 31 to 33 for nonblack and 29 to 31 for black women. When the role of anemia in explaining the ethnic differences in preterm birth was studied, anemia was defined as the tenth percentile for all term infants in the sample, weighted for the unequal sampling probabilities of each ethnic group. If different tenth percentiles were used for each ethnic group, the prevalence of anemia would not differ between ethnic groups and adjustment for anemia could not effect the relationship between ethnicity and preterm birth.
Anemia and preterm birth
Volume 164 Number 1, Part 1
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40 39 M E
A N H E
~
38
36 35
C
34
I
33
R
T
non-Black
37
M A
T 0
.
'
* Black
32 31
15-16 17-18 19-20 21-22 23-2425-2627-2829-3031-3233-3435-3637-3839-40
WEEKS GESTATION Fig. 1. Mean hematocrit value at each 2-week interval of gestation among pregnancies delivered at term,
Statistical methods. As described, all Asian, black, and Mexican preterm infants were studied, as were approximately one third of the white preterm infants. Term infants were selected as controls from each ethnic group according to the number of preterm cases from that group. Because the sampling probabilities were known and differed by outcome and ethnic group, odds ratios for ethnicity could be calculated by adjusting for the different sampling fractions according to the method described by Cain and Breslow.' This method adjusts both the odds ratios and the SEs derived from logistic regression for unequal sampling probabilities. To compare our results with those previously described,' the change in hematocrit value during a normal pregnancy that resulted in a term infant was evaluated. The mean hematocrit value at each 2-week interval during the second and third trimesters for women delivered of infants of >36 weeks' gestational age was studied. Next, we evaluated the association between anemia as determined at each 2-week interval of gestational age and subsequent preterm birth. In addition, the occurrence of preterm birth was compared between women who were anemic at any time during the second trimester and women who had hematocrits measured during the second trimester and were not anemic. This analysis was done for all women and separately for women in each ethnic group. Finally, we evaluated the role of anemia in explaining the ethnic differences in preterm birth. The odds ratios were calculated for preterm birth to Asian, black, and Mexican women compared with white women. The contribution
of differing rates of anemia to the ethnic differences in preterm birth was assessed by comparison of the odds ratios before and after adjustment for the presence of anemia at any time during the second trimester. Logistic regression was used for these analyses. Results
The mean hematocrit value at each 2-week interval of gestation during the second and third trimesters is presented in Fig. 1 for black and nonblack women delivered of infants at term. Hematocrit values appeared to fall during the early second trimester, reach a nadir at approximately 26 to 28 weeks' gestation, begin to increase again at 31 to 33 weeks, then increase sharply at term, reaching a maximum at 40 weeks' gestation. Throughout the second and third trimesters, hematocrit values were approximately 2% lower among black than among nonblack women. Because hematocrit values varied widely by gestational age, the relationship between preterm delivery and anemia was evaluated at 2-week intervals of gestational age (Table 11). Odds ratios for preterm birth and anemia were adjusted for ethnicity, age, education, marital status, smoking, and gestational age at first prenatal visit. The odds ratios for anemia during the period from 13 to 26 weeks' gestation were> 1.0, indicating an increased risk of preterm delivery among anemic women. However, anemia after 26 weeks' gestation generally had an inverse relationship with preterm birth. Having been anemic at any time during the second trimester (weeks 13 to 26) was positively asso-
62
Klabanoff at al.
January 1991 Am.J Obstet Gynecol
Table II. Odds ratios for preterm delivery and anemia by gestational age Weeks' gestation
No.
Adjusted odds ratio*
13-14 15-16 17-18 19-20 21-22 23-24 25-26 27-28 29-30 31-32 33-34 35-36
284 199 148 163 197 203 163 218 267 408 328 391
1.4 2.1 4.3 1.4 2.0 1.4 3.5 0.8 1.4 0.5 0.7 0.5
95% CI
0.5-3.7 0.6-7.4 1.2-15.5t 0.4-5.4 0.8-4.9 0.5-3.7 1.1-l1.3t 0.3-2.1 0.6-3.3 0.1-1.5 0.3-1.6 0.2-1.3
*Adjusted for ethnicity, maternal age, education, marital status, smoking, gestational age at first prenatal visit. tp < 0.05.
ciated with preterm birth [odds ratio, 1.9; 95% confidence interval (CI) 1.2 to 3.0]. The odds ratios by ethnic group were 2.6 (95% CI, 0.7 to 9.6) for Asian women, 1.8 (95% CI, 0.9 to 3.6) for black women, 2.0 (95% CI, 0.8 to 4.7) for Mexican women, and 1,8 (95% CI, 1.0 to 3.5) for white women. Anemia at any time during the third trimester was not analyzed because women delivered of infants at term had more opportunity than did women delivered of infants before term to have anemia diagnosed. This does not apply to the biweekly analyses in Table II because only women who were still pregnant during that interval could have their hematocrit measured. The role of anemia in the explanation of the ethnic differences in preterm birth is shown in Table III. The unadjusted odds ratios for preterm delivery among Asian, black, and Mexican women, compared with white women were 1.2, 2.0, and 1.2, respectively. Despite the association between second-trimester anemia and preterm birth, adjustment for the presence of anemia at any time during the second trimester had little effect on the ethnic differences in preterm delivery. Anemia after 26 weeks' gestation was negatively associated with preterm birth; therefore it could not possibly account for the observed ethnic differences.
Comment We found that in pregnancies carried to term, the mean hematocrit value rises dramatically after 37 weeks' gestation. This is in agreement with earlier reports,S and with the standards developed by the Centers for Disease Control for hematocrit values during pregnancy.7 Because the same group of women contributed data over most of pregnancy, our study has advantages over purely cross-sectional data in which entirely different women may have had blood drawn during each interval. However, our data are not fully longitudinal; every woman did not have an hematocrit drawn during
every interval. To the extent that women who had blood drawn during any given interval differed from those who did not, our results will be in error. Our results agree closely with established standards'? and are consistent with the observation that plasma volume is relatively constant after 34 weeks' gestation but red cell mass increases linearly throughout pregnancy.8 This suggests that the lack of regular measurements taken longitudinally on each woman was unlikely to influence the results. Our results confirm those of Klebanoff et al.,' who found that when measured prospectively, thirdtrimester anemia was not a risk factor for preterm birth. Our results contradict those of Lieberman et al.,' who found a very strong dose-response relationship between preterm birth and third-trimester anemia (hematocrit value ~34) and that anemia accounted for much of the difference in preterm delivery rates between black and white women. Lieberman et al. I defined anemia with hematocrit values obtained at delivery. Our data suggest that the use of delivery hematocrit values is inappropriate because hematocrit rises sharply at term. Use of delivery hematocrit values will therefore produce a spurious relationship between anemia and preterm birth. We found a moderate relationship between anemia during the second trimester and preterm birth. Anemia during the third trimester tended to be negatively associated with preterm birth. In addition, we found that statistical adjustment for the presence of anemia did not affect the ethnic differences in preterm delivery previously reported in this cohort. 9 The association between second-trimester anemia and preterm delivery is not inconsistent with previous findings on this subject because prior work did not evaluate anemia during the second trimester. 3 Why should second-trimester anemia be associated with preterm delivery when anemia later in pregnancy is not? There is no clear answer. This association has not been reported previously, and it may be an artifact of multiple comparisons. The relationship was consistent within ethnicity, making simple chance less likely. It is possible that a small group of high-risk, severely anemic women had hematocrits measured on multiple occasions. This detection bias would tend to overstate the relationship between preterm birth and anemia. We found that women who had one hematocrit measured during the second trimester were no more likely to be delivered of preterm infants than were women who had none. However, women who had two or more hematocrit measurements during the second trimester were at elevated risk of preterm delivery. In an attempt to reduce detection bias, we analyzed mean and first hematocrit values during the second trimester. The results were unchanged when anemia was defined with the use of the mean of all hematocrit values obtained during the second trimester and with the first hemat-
Anemia and preterm birth
Volume 164 Number I. Part 1
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Table III. Odds ratios and 95% CI for preterm delivery by ethnic group White
Unadjusted Adjustedt
Asian
Odds ratio
Odds ratio
1.0 1.0
1.2 1.2
I
Mexican
Black
95% CI
Odds ratio
0.9-1.6 0.9-1.6
2.0 1.9
I
95% CI 1.6-2.4* 1.5-2.3*
Odds ratio
1.2 1.2
I
95% CI 1.0-1.5* 1.0-1.6
*P < 0.05. tOdds ratio adjusted for the presence of second-trimester anemia.
ocrit value obtained during the second trimester. Nevertheless, the association between anemia during the second trimester preteI'm birth should be considered tentative and needs to be confirmed in other populations. A precise definition of anemia during pregnancy is confounded by various factors such as altitude, ethnicity, use of iron supplements, and changes in plasma volume during pregnancy. Definitions of anemia have ranged from hematocrit values of 534, I to a hemoglobin of <10 gm per deciliter lO (corresponding to hematocrit value of 30), to the Centers for Disease Control's month-specific standards.' Use of either of the two fixed values (534 or <30) produced similar results in our data. Because there is no standard definition of anemia for 2-week intervals during pregnancy, we chose to use the internal standard of the tenth percentile at each 2-week interval, as determined among pregnancies that ultimately resulted in births at term. When evaluating the role of anemia, separate standards were used for black and nonblack women to eliminate ethnicity as a confounder. When evaluating the role of ethnicity in preterm birth, a single standard for anemia was used so as not to overadjust for ethnicity. Rapid changes in red cell mass and plasma volume during pregnancy make interpretation of isolated hematocrit values difficult. Low hematocrit levels may result from inadequate red cell mass or a large expansion of plasma volume. Inadequate expansion of plasma volume, which would result in high hematocrit values, has been associated with an increased risk of a variety of pregnancy complications."·!· Women with normal red blood cell mass and a large expansion of plasma volume would probably be at low risk of adverse outcome despite an apparent anemia. Therefore further studies with clinical determinations of hematocrit value are unlikely to resolve this issue. Anemia in pregnancy should be defined as a function of red cell mass and plasma volume, neither of which are measured in clinical settings. In addition, measured hematocrit is a result of the interaction of physiologic, nutritional, and other life style factors. Despite any observed associations, low hematocrit values may merely be a marker for social and environmental deprivation, rather than being a cause of preteI'm birth. Therefore treatment of anemia alone might not affect the occurrence of spontaneous preterm birth.
In conclusion, anemia during the second trimester was found to be positively associated with spontaneous preterm birth; anemia later in pregnancy tended to be negatively associated with preteI'm birth. Secondtrimester anemia did not contribute to the observed ethnic differences in the occurrence of preteI'm birth. Future studies of anemia during pregnancy should supplement routine clinical data with detailed measures of red cell mass and plasma volume. These additional measurements may clarify the role of anemia in the pathogenesis of preterm birth as well as explain the observed ethnic differences in hematocrit values.
We thank Dr. Sholom Wacholder of the National Cancer Institute for his assistance with the sampling method and Ms. Marilyn Goldhaber for her assistance in the development of this project. REFERENCES I. Lieberman E, Ryan KJ, Monson RR, Schoenbaum SC.
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