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ORIGINAL RESEARCH |
From the Department of Epidemiology, School of Public Health and Community Medicine, Molecular and Cellular Biology Program, Division of Allergy and Infectious Diseases, School of Medicine, Department of Family and Child Nursing, School of Nursing, and Department of Health Services, School of Public Health and Community Medicine, University of Washington, Seattle, Washington; and Divisions of Human Biology and Clinical Research and Program in Infectious Diseases, Fred Hutchinson Cancer Research Center, Seattle, Washington.
Address reprint requests to: Mona T. Lydon-Rochelle, PhD, MPH, CNM, University of Washington, Box 357262, Seattle, Washington 98195–7762; Phone: (206) 221–6576, Fax: (206) 543–6656, e-mail: minot{at}u.washington.edu.
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ABSTRACT |
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METHODS: We conducted a population-based case-control study using linked Washington State birth certificate–hospital discharge data for singleton live births from 1987 to 1998 to assess the association between maternal, pregnancy, delivery, and infant characteristics and short umbilical cord. Cases (n = 3,565) were infants diagnosed with a short umbilical cord. Controls (n = 14,260) were randomly selected from among births without a diagnosis of short umbilical cord.
RESULTS: Case mothers were less likely to be overweight (body mass index 25 or more, odds ratio [OR] 0.7; 95% confidence interval [CI] 0.6, 0.8) and more likely to be primiparous (OR 1.4; 95% CI 1.3, 1.6). Case infants were more likely to be female (OR 1.3; 95% CI 1.2, 1.4), have a congenital malformation (OR 1.6; 95% CI 1.4, 1.8), and be small for their gestational age (risk ratio [RR] 1.6; 95% CI 1.4, 1.9). A short cord was associated with increased risk for maternal labor and delivery complications, including retained placenta (RR 1.6; 95% CI 1.2, 2.3) and operative vaginal delivery (RR 1.4; 95% CI 1.3, 1.5). Adverse fetal and infant outcomes in cases included fetal distress (RR 1.8; 95% CI 1.6, 2.1) and death within the first year of life among term infants (RR 2.4; 95% CI 1.2, 4.6).
CONCLUSION: Modifiable risk factors associated with the development of a short cord were not identified. Case mothers and infants are more likely to experience labor and delivery complications. Term case infants had a 2-fold increased risk of death, which suggests closer postpartum monitoring of these infants.
LEVEL OF EVIDENCE: II-2
We used maternally linked birth records to determine risk factors for a short umbilical cord, to examine the risk of selected labor and delivery outcomes attributable to short umbilical cords, and to identify the risk of selected fetal and infant outcomes attributable to short umbilical cords among women delivering singleton live births.
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MATERIALS AND METHODS |
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Subjects were selected from among all singleton live-born infants from 1987 through 1998 (n = 17,825). Cases (n = 3,565) consisted of all births with a diagnosis of a short umbilical cord (International Classification of Diseases, Ninth Revision, Clinical Modification [ICD-9-CM] code 663.4).15 Controls (n = 14,260) were randomly selected from among the remaining singleton births that occurred during the same years (frequency was matched on the year of delivery) in a 4:1 ratio. The Human Subjects Protection Review Boards at the University of Washington and the Washington State Department of Health approved this study.
To evaluate potential risk factors that may be associated with the development of a short umbilical cord, we selected characteristics that have been reported in previous studies as well as several new characteristics. These characteristics included body mass index; smoking and alcohol intake during pregnancy; various maternal medical conditions; reproductive history, including parity, prior fetal loss, and previous preterm or small-for-gestational-age (SGA) infant; prenatal procedures; various pregnancy complications; and fetal characteristics.
We identified the presence of risk factors and congenital anomalies using birth certificates and hospital discharge data. A risk factor or congenital malformation was considered present if either or both data sources indicated its presence. The following risk factors were identified from hospital discharge ICD-9-CM codes or birth certificates: anemia (280–285), cardiac disease (429), diabetes mellitus (250), hypertension (401–405), epilepsy (345), genital herpes (054.1), oligohydramnios (761.2), hydramnios (761.3), incompetent cervix (761.0), placenta previa (762.0, 641.0, 641.1), gestational diabetes (648.8), preeclampsia (642.4, 642.5), and eclampsia (642.6). We classified congenital malformations into groups with similar characteristics and used ICD-9-CM codes along with birth certificates to identify the various congenital defects: any malformations (740–759), chromosomal (758), gastrointestinal (750–751), circulatory/respiratory (745–748), integument (757), musculoskeletal (754–756), genitourinary (752–753), central nervous system (740–742), and other malformations (743,744,749,759). All remaining maternal characteristics and risk factors not listed above were taken from birth certificates only.
Maternal labor and delivery outcomes included malposition (breech [652.2], transverse [652.3], other malposition [652.0, 652.4, 652.5, 652.6, 652.7, 652.8, 652.9]), induced labor, stimulated labor, abruptio placenta (641.2), prolonged labor (662), prolonged second stage (662.2), third- and fourth-degree perineal lacerations (664.2, 664.3), retained placenta (667), postpartum hemorrhage (666.0, 666.1), and delivery method (spontaneous vaginal delivery, operative vaginal delivery, and cesarean section). Fetal and infant outcomes included gestational age, SGA,16 birth weight, 5-minute Apgar score, asphyxia (768.5, 768.6, 768.9), hypoxic-ischemic encephalopathy (348.1, 997.01, 767.0, 770.8), birth injury (767), fetal distress (656.3, 768.2, 768.3, 768.4), meconium aspiration (770.1), presence of moderate/ heavy meconium, assisted ventilation, and infant death (within the first year of life).
The associations between selected maternal characteristics, medical conditions, reproductive history, prenatal procedures, pregnancy complications, and fetal factors and the presence of a short umbilical cord were estimated by using Mantel-Haenszel stratified analysis to obtain estimates of the odds ratio (OR) and test-based 95% confidence intervals (CIs) using SAS 8.2 software for Windows (SAS Institute, Cary, NC). To assess the association between short umbilical cord and adverse labor and delivery outcomes and to assess adverse fetal and infant outcome association with short umbilical cord, we used Mantel-Haenszel estimates of the relative risk (RR) and 95% CIs. Delivery and fetal and infant outcomes were restricted to cases and controls without congenital malformations, because we wanted to examine the impact of a short umbilical cord on labor and delivery without including outcomes that may be related to congenital defects.
We chose several variables a priori to be assessed as potential confounders or effect modifiers, including maternal age, payment method, the trimester when prenatal care began, body mass index, parity, previous pregnancy losses, presence of malformations, sex of infant, gestational age in weeks, and birth weight (in grams). The OR or RR was adjusted for a potential confounder(s) if the adjusted RR differed from the crude measure of risk by 10% or more. All risk factors and outcomes were assessed on an individual basis. To evaluate effect modification, we applied the Breslow-Day test for homogeneity17 to test for differences between all strata, using P < .05 to denote statistical significance.
We examined birth outcomes for vaginal deliveries alone to eliminate the possibility that short cords may be differentially reported during a cesarean delivery. We also assessed outcomes in term deliveries by restricting one analysis to those who delivered between 37 and 42 weeks of gestation, because prematurity of the infant may influence the outcomes of interest. For comparability, we adjusted variables for the same confounders that we detected in outcomes under unrestricted conditions. However, if the RR did not differ by more than 10% from the adjusted risk estimate in a subanalysis, we reported only the crude RR. Both subanalyses were also restricted to infants born without congenital malformations.
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RESULTS |
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). However, women with short cord–infants tended to be more educated and less likely to have used Medicaid insurance.
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). Primiparous women were at an elevated risk for having a short cord–infant compared with those with 1 previous delivery (OR 1.4; 95% CI 1.3, 1.6). Among women with a previous pregnancy, mothers of short cord–infants were similar to controls in their history of having a previous fetal loss or previous preterm infant. Although there was no significant association with amniocentesis, women with short cord–infants were 20% less likely to have had ultrasound (OR 0.8; 95% CI 0.7, 0.9). Women with short cord–infants did not differ from mothers of control infants regarding likelihood of pregnancy complications. Infants with short cords were more likely to be female (OR 1.3; 95% CI 1.2, 1.4) and to have a congenital malformation (OR 1.6; 95% CI 1.4, 1.8).
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).
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). Based on the Breslow-Day test of homogeneity, significant effect modification was found for prolonged labor and prolonged second stage of labor, both of which were altered by parity; the presence of postpartum hemorrhage and hypoxic-ischemic encephalopathy both differed by birth weight strata; fetal distress was modified by sex of the infant; and risk of infant death was found to be contingent on weeks of gestation.
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After adjustment for birth weight, infants with short cords were similar to controls in gestational age distribution. However, short cord–infants were more likely to be SGA (RR 1.6; 95% CI 1.4, 1.9), to have hypoxic-ischemic encephalopathy (RR 1.4; 95% CI 1.2, 1.8), and to have fetal distress (RR 1.8; 95% CI 1.6, 2.1). The risk of hypoxic-ischemic encephalopathy was especially high among infants weighing 2,500–3,999 g compared with other birth weight categories (RR 1.7; 95% CI 1.3, 2.2). Finally, although the overall risk estimate was not significant, term case infants had higher rates of infant death in the first year of life (RR 2.4; 95% CI 1.2, 4.6).
To exclude the possibility of underreported cases in a cesarean delivery, we repeated our analysis shown in Table 4
for vaginal deliveries only (in addition to examining only infants without any malformations); however, the changes in RRs comparably were not substantial enough to report separately. Similarly, to exclude delivery and infant outcomes that were due to complications from having a preterm infant, we repeated the analysis shown in Table 4 for infants born between 37 and 42 weeks of gestation. The majority of delivery and infant outcomes was similar after this restriction except that case women were significantly more likely to have abruptio placenta (RR 1.6; 95% CI 1.2, 2.3; results not shown in table).
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DISCUSSION |
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Our findings confirm previous observations of the association of several factors with short umbilical cords. Seminal observations reported by Naeye1 and Mills10 used data on more than 35,000 singleton pregnancies from different areas within the United States participating in the Collaborative Perinatal Study of the National Institutes of Neurological and Communicative Disorders and Stroke. These studies demonstrated that female infants have shorter cords than male infants, which is consistent with our findings. Second, a positive correlation between umbilical cord length and parity has been reported.9 By extension, we demonstrated that a diagnosis of short cord was more common among primiparas. Third, socioeconomic status has been associated with cord length,1 and in the current study, we found differences in insurance coverage and maternal education, suggesting minor disparities in socioeconomic status between cases and controls. However, differences in insurance payer do not explain the magnitude of differences in the use of ultrasound between cases and controls. Last, in common with a previous study,1 we found no association of short umbilical cords with maternal smoking or alcohol intake.
Several studies2,3,12 have implicated a relationship between the presence of oligohydramnios during pregnancy and the subsequent development of a short umbilical cord. Miller et al2,3 hypothesized that the umbilical cord grows in response to tensile forces exerted on the cord by fetal movements. Moessinger et al12 showed that in the presence of oligohydramnios, the umbilical cords of rat fetuses were 65% of control length. We did not detect an association between oligohydramnios or hydramnios and a greater or lesser likelihood of being a case, respectively. Likewise, more recent studies4,5 using animal models have argued against the "stretch hypothesis," stating that the umbilical cord continues to grow throughout pregnancy in an almost linear fashion. Instead, these studies have proposed a multifactorial explanation for the occurrence of short cords.
Indeed, in this study, no single factor accounted for the occurrence of all short cords, suggesting a multifactorial etiology. Surprisingly, fewer than 10% of case infants were reported as having congenital anomalies. Our data suggest an association of short cords with particular malformation sequences, particularly those caused by chromosomal anomalies. Other important sequences may be those affecting gastrointestinal and circulatory-respiratory systems. Previous studies of the association of short cords with malformation sequences and fetal problems have defined several groups of such sequences and problems among infants including stillborns and those who died shortly after birth due to multiple severe anomalies. These include ADAM sequence, cyllosomus/pleurosomus, acephalus-acardia, presumed primary defect of the umbilical cord and abdominal wall formation, schisis association, and reduced fetal movement.3,19 Such anomalies are usually multisystem disorders, although they principally involve the central nervous system, limbs, and cardiovascular system or are associated with defects in the formation of the anterior abdominal wall. Of interest, in our study, malformations of the central nervous system were not associated with the occurrence of short cords, and the association of musculoskeletal malformations was modest. Such associations, if present, would have supported the "stretch hypothesis" for the lengthening of the cord and indeed formed the basis of some previous arguments2,3 in favor of the "stretch hypothesis." Miller et al3 stated that extremes of decreased fetal movement occur in cases of amelia, acardia, arthrogryposis, and atrophy of spinal muscles. Moessinger et al12 suppressed fetal movement by curarization, which led to the development of shorter cords among rat fetuses. By extension, severe central nervous system malformations also would reduce fetal activity.
Our findings of increased risk of prolonged second-stage1 and operative vaginal delivery13 is consistent with previous studies. However, our findings of increased risk of retained placenta and postpartum hemorrhage among women with short umbilical cords have not been previously reported. Finally, we found increased likelihood of important fetal and infant outcomes, including being SGA, having hypoxic-ischemic encephalopathy, fetal distress, and infant death, among case infants than controls. Our most striking finding was the 2-fold increase in risk of death among term infants born with short cords. Clausson et al20 reported a 3-fold increase in infant deaths among term SGA infants without congenital malformations in a population-based study of 510,029 singleton term and postterm births recorded in the Swedish Birth Registry, but they did not specifically examine cord length. In addition, our findings are consistent with previous studies, which reported low birth weight7 and fetal distress13,18 among infants with short cords.
The major limitation of this study is the potential for misclassification of case status, because we identified cases by ICD-9-CM codes, whereas previous studies adopted an absolute cord length as a case definition.1,7,8,18,19 Underreporting of short umbilical cord cases would be present if neonates with umbilical cords close to the normal length were noted with a lower frequency. In this instance, we would expect the risk estimates to be closer to the null. Nonetheless, differential misclassification can also be present if newborns with short cords were noted as having a short cord more readily in the presence of other adverse outcomes. Our findings did confirm previous reports regarding short umbilical cords, such as low socioeconomic status and fetal distress,13,18 suggesting that differential misclassification was minimal. We restricted all outcome analyses and subanalyses to newborns without malformations under the assumption that, in the presence of any malformation, the short umbilical cord may be missed. In addition, we attempted to assess potential misclassification of the short cord in cesarean deliveries by examining outcomes in vaginal deliveries only but found no differences in the outcomes. Finally, gestational age or the presence of congenital anomalies could potentially lead to differential recognition of cord anomalies, resulting in differential misclassification of case status. We addressed the concern by restricting delivery and fetal and infant outcome analyses to infants without recognized malformations. The potential effect of misclassification by gestational age also is unlikely to have had a significant impact on the associations we report as 94% of cases were term deliveries.
Although reference standards for cord length have been reported,10 variation exists in the definition of short cord.1,7,8,18,19 Naeye1 adopted a cord length of 40 cm, whereas Nnatu et al8 used a measurement of less than 46 cm. Other studies7,18,19 defined short cords as less than or equal to 35 cm in length. Their reported prevalence of short cords ranged from 2%7 to 6%1,18 to 10%,8 contrasting with our incidence of 0.4%. However, most of the previous studies were not population based;7,8,18,19 had restrictive exclusion criteria, including gestation less than 37 weeks, malposition, and cases of umbilical cord traumas;1,7,18 or examined a high-risk population,8 which may account for our lower incidence of short cord. In addition, the prevalence of short umbilical cords was consistent between years 1987 and 1998 in the State of Washington.
Although we were able use both birth certificate and hospital discharge data for the majority of risk factors and outcomes, giving us greater likelihood of identifying these factors, some factors were only provided in the birth certificate data. Underreporting or failure to record risk factors, such as smoking and alcohol intake, is possible, and bias may have been introduced if case mothers reported these risk factors with a differing accuracy compared with control mothers. As such, we would expect our findings to result in a null risk estimate.
Despite these limitations, our study has several strengths. We identified a large number of cases, lending greater precision and power to our analyses than had been possible with previous studies. Selection bias was unlikely because we included all identified cases and a large sample of randomly selected controls. The availability of linked data enabled us to more accurately identify variables than through use of either of the data sources alone. In addition, the definitions and classifications of risk factors and outcomes remained consistent between 1987 and 1998 because the ICD-9 codes and birth certificate standardized definitions had not changed.
In summary, we demonstrated that case infants had greater odds of having certain organ or chromosomal malformations, which suggest that there may be a specific sequence association. Although we did not identify any modifiable risks that would predispose infants to the development of a short cord, several previously unreported complications of labor and delivery in both mothers and infants were identified. Because antenatal and intrapartum screening or diagnoses for short cords do not exist at present, preventive measures are unavailable. Most importantly, we recognized a 2-fold increase in infant death among term cases, which suggests closer monitoring of infants born with a short cord after birth. Moreover, the presence of a short cord at delivery may indicate an important surveillance marker for infants in the first year of life. However, the relationship between short cords and first year of life mortality among term infants should be further addressed in future studies.
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Footnotes |
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The authors thank William O’Brien for data linkage and management support.
doi:10.1097/01.AOG.0000102706.84063.C7
Received May 23, 2003. Received in revised form September 17, 2003. Accepted September 26, 2003.
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2. Miller ME, Jones MC, Smith DW. Tension: the basis of umbilical cord growth. J Pediatr 1982;101:844.[Medline]
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