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Maternal Low-Level Lead Exposure and Fetal

Growth

ARTICLE in ENVIRONMENTAL HEALTH PERSPECTIVES · OCTOBER 2010

Impact Factor: 7.98 · DOI: 10.1289/ehp.0901561 · Source: PubMed

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Kitty H Gelberg

New York State Department of Health

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Charlotte Druschel

New York State Department of Health

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Available from: Charlotte Druschel

Retrieved on: 28 September 2015
Environmental Health Perspectives • volume 118 | number 10 | October 2010 1471

Research | Children’s Health

With the banning of lead-based paint in 1977, 

and the phasing out of lead-based gasoline 

in the 1980s and its ban in 1996, the blood 

lead (PbB) concentration among the general 

U.S. population has been declining steadily 

[Centers for Disease Control and Prevention 

(CDC) 2005]. However, the general population exposure to low lead levels continues 

because of the widespread use of lead and its 

ubiquitous nature (CDC 2005). According to 

the 2003–2004 National Health and Nutrition 

Examination Survey (CDC 2005), the mean 

PbB among women 18–49 years of age was 

1.2 μg/dL, with a 95th percentile of 2.6 μg/dL.

PbBs < 10 μg/dL induce adverse effects 

in humans, including elevated blood pressure, impaired nervous system development, 

delayed sexual maturation, neurobehavioral 

effects, depressed renal glomerular filtration 

rate, and reduced heme synthesis [Agency 

for Toxic Substances and Disease Registry 

(ATSDR) 2007]. Furthermore, a clear threshold for these sensitive effects has not been 

identified (ATSDR 2007). Maternal lead can 

readily cross the placenta and enter fetal blood 

circulation starting around week 12–14 of 

pregnancy, making the fetus susceptible to 

lead poisoning (Lin et al. 1998).

It is biologically plausible that lead can 

induce low birth weight, preterm birth, and 

small for gestational age. Lead can potentially impair normal fetal bone growth by 

competing with calcium for deposition into 

bone because lead and calcium have similar chemical characteristics (Potula 2005). 

Experimental evidence provides support for 

a potential effect of lead on preterm birth. 

Lead impedes collagen synthesis and praline 

hydroxylation in mouse, which may have deleterious effects on chorioamniotic membrane 

structure and induce its premature rupture 

(Torres-Sanchez et al. 1999). Rats exposed 

to lead have reduced bone calcium content, 

reduced trabecular bone volume, altered 

growth plate morphology, and enhanced 

activities of spontaneous uterine contraction 

(Irgens 1998; Torres-Sanchez et al. 1999).

Limited epidemiologic studies have been 

conducted to examine maternal low-level 

lead exposure and fetal growth, especially 

using PbBs (Irgens 1998; Magri et al. 2003;

Rothenberg et al. 2002; Sowers 2002; TorresSanchez et al. 1999). Some studies included 

both low-level and high-level lead exposures, 

restricting the conclusions regarding low-level 

lead exposure alone (Torres-Sanchez et al. 

1999). Other studies are based on convenience 

samples such as prenatal clinic and Medicaid 

participants, limiting their generalizability 

(Sowers 2002).

Our study was designed to help address 

some of these issues, using a large populationbased PbB registry in New York state. The 

objectives were to examine whether maternal 

low-level PbB exposure (< 10 μg/dL) was 

inversely associated with birth weight and 

directly associated with the risk of preterm 

birth, and small for gestational age.

Methods

Study population and data sources. The 

study population comprised upstate New 

York (New York State, excluding New York 

City) mothers 15–49 years of age from 2003 

through 2005 who had a PbB test before or 

at the delivery date, and their singleton live 

births. PbBs were obtained from the New 

York State Heavy Metals Registry (HMR), 

which has maintained a statewide database 

since 1982 and receives reports on exposure 

to heavy metals, including lead, mercury, 

arsenic, and cadmium, from physicians and 

laboratories (New York State Department 

of Health Bureau of Occupational Health 

2008a). In 1992 the reporting requirement 

was changed from 25 μg/dL to include all 

test reports regardless of level (New York 

State Department of Health Bureau of 

Occupational Health 2008a). Information 

on birth outcomes and potential confounders 

was acquired from the birth certificate files, 

which are maintained by the New York State 

Department of Health, Bureau of Biometrics.

Study design and data linkage. A retrospective cohort design was used. The existing 

HMR records were linked with birth certificate files to form the study base. At first, 

women with multiple PbB reports were 

identified through deterministic matching 

techniques and transposed into one record 

containing information on all reporting dates 

and PbBs. To minimize the issues of data 

entry errors or missing values on identifiers, 

10 deterministic identifiers were created using 

Address correspondence to M. Zhu, West Virginia 

University, Department of Community Medicine, 

White Birch Towers, 4th Floor, 1299 Pineview 

Dr., Morgantown, West Virginia 26505 USA. 

Telephone: (304) 293-6682. Fax: (304) 293-0265. 

E-mail: mzhu@hsc.wvu.edu

We express appreciation to E. Franko and her colleagues at the New York State Department of Health, 

Bureau of Occupational Health for their assistance 

in access to blood lead reports; to T. Wang at the 

New York State Department of Health, Bureau of 

Biometrics, for assistance in data linkage with birth 

certificates; and to the anonymous reviewers for their 

constructive comments.

The authors declare they have no actual or potential 

competing financial interests.

Received 10 October 2009; accepted 18 May 2010.

Maternal Low-Level Lead Exposure and Fetal Growth

Motao Zhu,1 Edward F. Fitzgerald,2 Kitty H. Gelberg,3 Shao Lin,4 and Charlotte M. Druschel 4

1Department of Community Medicine, West Virginia University, Morgantown, West Virginia, USA; 2Department of Epidemiology and 

Biostatistics, State University of New York at Albany, Rensselaer, New York, USA; 3Bureau of Occupational Health, New York State 

Department of Health, Troy, New York, USA; 4Bureau of Environmental and Occupational Epidemiology, New York State Department of 

Health, Troy, New York, USA

Background: Limited epidemiologic studies have examined the association between maternal 

low-level lead exposure [blood lead (PbB) < 10 μg/dL] and fetal growth.

Objective: We examined whether maternal low-level lead exposure is associated with decreased 

fetal growth.

Methods: We linked New York State Heavy Metals Registry records of women who had PbB 

measurements with birth certificates to identify 43,288 mother–infant pairs in upstate New York 

in a retrospective cohort study from 2003 through 2005. We used multiple linear regression with 

fractional polynomials and logistic regression to relate birth weight, preterm delivery, and small for 

gestational age to PbB levels, adjusting for potential confounders. We used a closed-test procedure 

to identify the best fractional polynomials for PbB among 44 combinations.

Results: We found a statistically significant association between PbB (square root transformed) and 

birth weight. Relative to 0 μg/dL, PbBs of 5 and 10 μg/dL were associated with an average of 61-g 

and 87-g decrease in birth weight, respectively. The adjusted odds ratio for PbBs between 3.1 and 

9.9 μg/dL (highest quartile) was 1.04 [95% confidence interval (CI), 0.89–1.22] for preterm delivery and 1.07 (95% CI, 0.93–1.23) for small for gestational age, relative to PbBs ≤ 1 μg/dL (lowest 

quartile). No clear dose–response trends were evident when all of the quartiles were assessed.

Conclusions: Low-level PbB was associated with a small risk of decreased birth weight with a 

supralinear dose–response relationship, but was not related to preterm birth or small for gestational 

age. The results have important implications regarding maternal PbB.

Key words: birth weight, blood lead, epidemiology, fetal growth, low-level lead exposure, 

pregnancy, preterm birth, small for gestational age. Environ Health Perspect 118:1471–1475 

(2010). doi:10.1289/ehp.0901561 [Online 21 June 2010]
Zhu et al.

1472 volume 118 | number 10 | October 2010 • Environmental Health Perspectives

components from variables including the case 

number, social security number, date of birth, 

first name, last name, telephone number, ZIP 

code, street address, sex, and street address 

of the provider or physician ordering test. 

During each step, 50 matches were randomly 

selected and reviewed to ensure that the 

matches were accurate, using all the potential 

identifying variables: first name, last name, 

middle name, date of birth, street address, 

ZIP code, city, state, phone number, sex, 

street address, name of the provider or physician ordering test, and reporting laboratory 

identification number. A total of 215,426 

women 15–49 years of age were identified 

from 245,050 PbB tests that reported < 10 

μg/dL from 2003 through 2005.

PbB data were then matched with birth 

certificates to identify women who delivered 

live infants. Twenty deterministic identifiers 

were created using components from variables 

including date of birth, social security number, 

first name, middle name, last name, phone 

number, residential street, and ZIP code of the 

mother, and residential street and ZIP code 

of the father. A total of 44,932 singleton live 

births were identified with at least one PbB

test by delivery and the maximal lead level 

< 10 μg/dL. We then excluded records with 

implausible birth weight–gestational age combinations (Alexander et al. 1996) to reduce the 

sample size to 44,873. Approximately 3.5% of 

mothers had multiple singleton births during 

the 3-year period, and we randomly selected 

one birth to finalize 43,288 mother–infant 

pairs. Approximately 3.0% of women received 

multiple PbB tests, so we similarly selected one 

test result at random.

This study was approved by the New York 

State Department of Health and the State 

University of New York at Albany institutional review boards.

Study variables. Exposure. PbB concentration was obtained from the HMR PbB reports. 

The study level was restricted to < 10 μg/dL, 

which accounted for 99.2% of reports. Atomic 

spectrometry is the method for routine screening and diagnostic work (Parsons 1993). Its 

accuracy is ± 1 μg/dL and the detection limit 

is 1 μg/dL (Parsons 1993). Any errors in the 

measurement of PbB would be expected to be 

nondifferential according to low birth weight 

and other fetal growth outcomes. Laboratories 

are required to pass three of the quarterly proficiency tests every year by the New York State 

Department of Health, Wadsworth Center 

for Laboratories and Research, to ensure the 

accuracy and comparability (Lin et al. 1998). 

The coefficient of variation was approximately 

7% among all laboratories in 2005 (New 

York State Department of Health Wadsworth 

Center 2006).

Outcomes. Birth outcomes were abstracted 

from the birth certificate files. Only singleton live births were selected. Birth weight was 

examined as a continuous variable. Preterm 

birth was defined as the gestational age < 37 

completed weeks from the date of the last menstrual period (March of Dimes Foundation 

2007). Small for gestational age was defined 

as the birth weight below the 10th percentile of birth weight for gestational age based 

on the distribution of 1996–2000 national 

birth weight by gestational week from week 25 

through week 42 (Boulet et al. 2006). Binary 

low birth weight (< 2,500 g) was not examined 

in multiple variable analysis because continuous 

birth weight provides more statistical power 

to detect subtle effects. In addition, low birth 

weight is a mix of preterm, growth-restricted, 

and constitutionally small births; preterm birth 

and small for gestational age were examined in 

this study. Regarding the accuracy of outcomes 

recorded in New York State birth certificates, 

the dates of last menses reported in the birth 

certificate exactly agreed with those recorded in 

medical records for 87% (Roohan et al. 2003). 

The agreement rate was increased to 93% when 

the tolerance was 1 week (Roohan et al. 2003).

Confounders. In addition to the timing 

of lead test in relation to the date of delivery, 

various potential confounders were abstracted 

from the BC files: maternal race (Caucasian, 

African American, other); maternal ethnicity 

(Hispanic or not); maternal age at the time of 

delivery; maternal education (less than high 

school graduate, high school graduate, some 

college or college degree, graduate education); 

participation in financial assistance programs 

(e.g., Medicaid; Family Health Plus; Women, 

Infants, and Children; other) (yes or no); selfreported maternal smoking during pregnancy 

(yes or no); self-reported maternal alcohol 

consumption during pregnancy (yes or no); 

self-reported illicit drug use during pregnancy 

(yes or no); trimester when prenatal care began 

(first trimester, second trimester, third trimester, or no prenatal care); parity (zero, one, two 

or more previous live births); sex of child; in

wedlock (yes or no); and prepregnancy body 

mass index.

Statistical analysis. For continuous outcomes (birth weight in grams and gestational 

age in days), we fitted multiple linear regression with fractional polynomials (Royston 

et al. 1999). We explored one or two terms of 

fractional polynomials in term of xp for PbB, 

where the power p is from –2, –1, –0.5, 1, 

2, 3, and natural logarithmic transformation. 

The selection of final fractional polynomials 

Table 1. Maternal and infant quantitative characteristics, upstate New York, 2003–2005.

Selected percentiles

Characteristic n Mean Minimum 10th 25th 50th 75th 90th Maximum

PbB (μg/dL) 43,288 2.1 0 1 1 2 3 3 9.9

Days from lead test to date of 

birth (day)

43,288 203 0 110 170 204 223 237 1,082

Maternal age (years) 43,288 27.6 15 20 23 27 32 36 49

Body mass index (kg/m2) 40,797 26.4 12.5 19.9 21.9 24.9 29.4 35.2 66.5

Gestational age (week) 43,288 38.8 20 37 38 39 40 41 44

Birth weight (g) 43,288 3,331 205 2,680 3,030 3,365 3,686 3,997 5,610

Table 2. Maternal and infant qualitative characteristics, upstate New York, 2003–2005 (total n = 43,288).

Characteristic n Percentagea

Race

Caucasian 29,434 68.0

African American 7,113 16.5

Other 6,689 15.5

Missing value 52

Ethnicity

Hispanic 8,447 19.7

Missing value 492

Education

Less than high school graduate 10,054 23.4

High school graduate 11,675 27.2

Some college or bachelor degree 16,857 39.3

Graduate study 4,337 10.1

Missing value 365

Smoking

Yes 8,834 20.5

Missing value 149

Alcohol drinking

Yes 493 1.1

Missing value 196

Drug abuse

Yes 1,216 2.9

Missing value 973

Financial assistance program

Yes 25,803 59.8

Missing value 114

Start of prenatal care visit

First trimester 29,187 72.9

Second trimester 8,811 22.0

Third trimester or no prenatal 

care visit

2,056 5.1

Missing value 3,234

Parity

0 17,376 40.4

1 13,715 32.0

2 or more 11,823 27.6

Missing value 374

In wedlock

Yes 20,378 47.4

Missing value 261

Infant sex

Male 22,154 51.2

Low birth weight

Yes 2,744 6.3

Preterm birth

Yes 3,519 8.1

Small for gestational age

Yes 4,092 9.5

Missing value 112

aThe calculation of percentage excluded missing values. 

There were no missing values for infant sex, low birth 

weight, and preterm birth.
Maternal low-level lead exposure and fetal growth

Environmental Health Perspectives • volume 118 | number 10 | October 2010 1473

was based on a closed-test procedure, which 

maintains the overall type 1 error (alpha level) 

of 0.05 for tests among 44 different combinations (Royston et al. 1999). For each 

outcome, a subset of biologically plausible 

risk factors in addition to PbB was selected 

to enter the model as potential confounders; 

those that remained with a significance level 

of 0.2 were retained (Dales and Urg 1978; 

Mickey and Greenland 1989; Royston et al. 

1999). Fractional polynomials were assessed 

for continuous confounders including gestational age and maternal age for birth weight 

outcome. Because the limit of detection for 

the routine screening and diagnostic laboratory method is 1 μg/dL (Parsons 1993), we 

conducted sensitivity analysis by a) comparing all records; b) excluding PbBs of 0 μg/dL; 

c) excluding PbBs < 1 μg/dL.

Furthermore, the quartiles of PbBs 

(≤ 1 μg/dL; > 1 μg/dL to 2 μg/dL; > 2 μg/dL 

to 3 μg/dL; > 3 μg/dL to < 10 μg/dL) were 

used for binary outcomes including preterm 

birth and small for gestational age. Adjusted 

odds ratios (aORs) of PbBs were estimated 

from logistic regression with fractional polynomials (Allison 1999; Royston et al. 1999). 

The quartiles of PbBs were forced into the 

model. A closed-test procedure was used to 

identify the 1 of 44 combinations of one 

or two fractional polynomials with the best 

model fit for continuous confounder: maternal age. The criteria for selecting and retaining 

confounders in the logistic regression were 

similar to those for linear regression. Analyses 

were conducted using STATA version 11 

(StataCorp, College Station, TX, USA).

Results

The average PbB concentration was 2.1 μg/dL, 

and the median was 2 μg/dL (Table 1). The 

average number of days from lead test to delivery was 203, and the 90th percentile was 237. 

Most PbB tests were conducted between the 

date of last menses and the date of delivery. 

The average birth weight was 3,331 g, and gestational age was 38.8 weeks. Table 2 presents 

the distribution of selected categorical maternal and infant characteristics. Approximately 

68% of births were to white women, and 

Hispanics accounted for 20%. The rates of 

low birth weight, preterm birth, and small for 

gestational age were 6.3%, 8.1%, and 9.5%, 

respectively.

A model that assumed a linear relation 

between the square root of PbB and birth 

weight fit the data better than models with 

all other combinations of fractional polynomial terms evaluated. Consequently, our final 

model included only a single term for PbB 

(raised to the 0.5 power), with an adjusted 

coefficient of –27.4 [95% confidence interval 

(CI), –37.7 to –17.1]. Estimated changes in 

birth weight with a 1-μg/dL change in PbB 

varied across the PbB distribution in the study 

population, consistent with the supralinear 

shape of the dose–response curve dictated by 

the model, so that a 1-μg/dL change in PbB 

from 0 μg/dL to 1 μg/dL was associated with a 

27.4-g decrease in mean birth weight, whereas 

a 1-μg/dL change in PbB from 9 μg/dL to 

10 μg/dL was associated with a 4.4-g decrease 

in mean birth weight (from a predicted mean 

decrease relative to predicted mean birth 

weight when PbB = 0 of 82.3 g to 86.7 g) 

(Table 3). Therefore, the model predicts the 

strongest estimated effects at the lowest levels 

of exposure, without a lower threshold of PbB 

below which there would be no predicted 

effect on birth weight. Figure 1 displays this 

dose–response relationship.

As for sensitivity analysis, the best-fit 

fractional polynomials were PbB–1 + PbB–1

× logarithmic-transformed PbB, after excluding PbB of 0 μg/dL from analysis (data not 

shown). Compared with PbB of 0.5 μg/dL, 

PbB of 9.5 μg/dL was associated with a 51-g 

decrease in birth weight. Compared with PbB 

of 1 μg/dL, PbB of 10 μg/dL was associated 

with a 32-g decrease in birth weight. When 

PbBs < 1 μg/dL were excluded, untransformed 

PbB fit the data the best, and the linear 

regression coefficient was a 7.0-g decrease in 

birth weight for a 1-μg/dL increase in PbB. 

Therefore, PbB of 10 μg/dL was associated 

with a 63-g decrease in birth weight, relative 

to PbB of 1 μg/dL. In contrast, the analysis 

using all PbBs including zeros and < 1 μg/dL 

suggested that PbB of 9.5 μg/dL was associated with an 84-g decrease in birth weight, 

relative to PbB of 0.5 μg/dL, and that PbB of 

10 μg/dL was associated with a 59-g decrease 

in birth weight, relative to PbB of 1 μg/dL. 

The analysis with all PbBs provided robust 

estimated effects of lead on birth weight.

A model that assumed a linear relation 

between untransformed PbB and gestational 

age in days fit the data better than models 

with all other combinations of fractional polynomial terms evaluated. Consequently, our 

final model included only a single linear term 

for PbB, with an adjusted coefficient of –0.09 

(95% CI, –0.24 to 0.05) after adjustment for 

timing of lead test, maternal age, race, smoking, alcohol consumption, participation in 

special financial assistance program, parity, 

and infant sex (data not shown).

Table 4 presents the association between 

the quartile PbBs and dichotomous outcomes: 

preterm birth and small for gestational age. 

There were not clear dose–response trends 

when all quartiles were assessed. The aORs 

for PbBs between 3.1 and 9.9 μg/dL (highest 

quartile) was 1.04 (95% CI, 0.89–1.22) for 

preterm birth and 1.07 (95% CI, 0.93–1.23) 

for small for gestational age, relative to 

≤ 1 μg/dL (lowest quartile).

Table 3. Association between PbB concentration 

and birth weight, upstate New York, 2003–2005.

PbB concentration 

(μg/dL)

Difference in birth weight 

in grams (model based)a

Estimate 95% CI

0 Reference

1 –27.4 –17.1 to –37.8

2 –38.8 –24.1 to –53.4

3 –47.5 –29.6 to –65.4

4 –54.8 –34.2 to –75.5

5 –61.3 –38.2 to –84.4

6 –67.2 –41.8 to –92.5

7 –72.5 –45.2 to –99.9

8 –77.6 –48.3 to –106.8

9 –82.3 –51.2 to –113.3

10 –86.7 –54.0 to –119.4

aThe model was a linear regression with fractional 

polynomials after adjustment for timing of lead test, 

gestational age, maternal age, race, Hispanic ethnicity, education, smoking, alcohol drinking, drug abuse, 

in wedlock, participation in special financial assistant 

program, parity, and infant sex. PbB concentration was 

transformed using a square root. The coefficient was 

–27.4 with an SE of 5.3.

Figure 1. Model-based dose–response relationship.

0 2 4 6 8 10

PbB (μg/dL)

3,320

3,300

3,280

3,260

3,240

3,220

Birth weight (g)

Table 4. Association between maternal PbB level and preterm birth, and small for gestational age, 

upstate New York, 2003–2005.

Maternal PbB level

Preterm birth Small for gestational age

Cases (n) aORa 95% CI Cases (n) aORb 95% CI

≤ 1.0 1,069 1.00 Reference 1,168 1.00 Reference

1.1–2.0 1,036 1.03 0.93–1.13 1,268 1.07 0.98–1.17

2.1–3.0 1,171 1.01 0.92–1.10 1,353 1.06 0.98–1.16

3.1–9.9 243 1.04 0.89–1.22 303 1.07 0.93–1.23

aaORs are estimated from logistic regression with fractional polynomials after adjustment for timing of lead test, 

maternal age at delivery, race, Hispanic ethnicity, smoking, drug abuse, in wedlock, participation in special financial 

assistance program, parity, and infant sex. The quartiles of PbB concentration were untransformed, and fractional 

polynomials were used for maternal age. baORs are estimated from logistic regression with fractional polynomials after 

adjustment for timing of lead test, maternal age at delivery, race, education, smoking, drug abuse, in wedlock, participation in special financial assistance program, parity, and infant sex. The quartiles of PbB concentration were untransformed and fractional polynomials were assessed for maternal age.
Zhu et al.

1474 volume 118 | number 10 | October 2010 • Environmental Health Perspectives

Discussion

Overall, maternal PbBs < 10 μg/dL were associated with a small but statistically significant decrease in birth weight. The decrease in 

birth weight for a 1-μg/dL increase in PbB 

ranged from an estimated means value of 4 g 

(from 9 to 10 μg/dL) to 27 g (from 0 to 1 μg/

dL). This is consistent with the estimate of 

a 6.2-g decrease in birth weight per 1-μg/dL 

increase in PbB from a study of 272 mother–

infant pairs in Mexico (Gonzalez-Cossio et al. 

1997); 3.0 g in a study of 4,354 pregnancies in Boston (Bellinger et al. 1991); 0.8 g 

from a study of 54 term neonates in Turkey 

(Atabek et al. 2007); and 0.3 g in a study of 

55 newborns in Brazil (Zentner et al. 2006), 

despite that fact that their mean lead levels 

were higher than ours.

We found that a model of birth weight 

as a function of square root–transformed 

PbB provided the best fit to the data. This 

model predicted estimated effects of lead 

that were greater at the lower end of the PbB 

distribution than at higher levels (supralinear dose–response relationship). A similar 

supralinear relationship has been reported 

for PbB < 10 μg/dL with IQ and Mental 

Development Index (Canfield et al. 2003; 

Lanphear et al. 2005; Tellez-Rojo et al. 

2006). A pooled analysis of seven international prospective cohort studies found that 

the decrease in full-scale IQ score per 1-μg/

dL increase in PbB estimated from the linear regression with untransformed PbB was 

greater among children with a maximum PbB 

< 7.5 μg/dL than in those with a maximum 

PbB ≥ 7.5 μg/dL (Lanphear et al. 2005). 

Further analysis suggested a linear relationship between the logarithmic-transformed 

PbB and IQ (coefficient: 6.9) (Lanphear et al. 

2005). An analysis of 294 children found a 

logarithmic-transformed PbB was linearly 

associated with Mental Development Index 

(Tellez-Rojo et al. 2006). The estimated effect 

of lead estimated from the linear regression 

with untransformed PbB was larger at < 5 

μg/dL than between 5 and 10 μg/dL (TellezRojo et al. 2006). Researchers have used 

quadratic term (Canfield et al. 2003) and 

logarithmic transformations (Lanphear et al. 

2005; Tellez-Rojo et al. 2006) to describe 

the supralinear relationship between lead and 

intellectual impairment. We found that the 

square root transformation provided the best 

fit for birth weight, compared with 43 other 

fractional polynomials linear, reciprocal, logarithmic, square foot, quadratic, and cubic 

terms. Further studies are needed to confirm 

whether the supralinear relationship between 

PbB and birth weight is best described with 

a square root transformation. Consistent 

with previous studies of intellectual development (Canfield et al. 2003; Lanphear et al. 

2005; Tellez-Rojo et al. 2006), our analysis 

supports that there is no clear threshold for 

the effects of lead on sensitive outcomes such 

as birth weight.

Bellinger et al. reported that the mean gestational age was 0.3 week longer among those 

with umbilical cord PbBs 5.0–9.9 μg/dL, 

relative to PbBs < 5.0 μg/dL (Bellinger et al. 

1991). In contrast, we found that a 1-μg/dL 

increase in maternal PbB was associated with 

a statistically nonsignificant 0.09-day decrease 

in gestational age. Similarly, Jelliffe-Pawlowski 

et al. reported that among women with PbB 

≥ 10 μg/dL, a 1-μg/dL increase in lead level 

was associated with an average 0.3-day decrease 

in gestational age (Jelliffe-Pawlowski et al. 

2006). In a case–control study of 620 pregnant women in Mexico City, compared with 

umbilical cord PbBs < 5.1 μg/dL, the aOR 

of preterm birth for lead level 5.1–9.0 μg/dL 

was 2.72 (95% CI, 1.03–7.19) among primiparous women, but 0.48 (95% CI, 0.21–1.08) 

among multiparous women (Torres-Sanchez 

et al. 1999).

Bellinger et al. found that lead levels 

between 5 and 9.9 μg/dL were not statistically 

related to increased risk in dichotomous preterm birth and small for gestational age, compared with lead levels < 5 μg/dL (Bellinger 

et al. 1991). A cohort study by Sowers (2002) 

of 705 pregnant women in Camden, New 

Jersey, did not find any statistically significant 

association with dichotomous preterm birth, 

or small for gestational age. Consistent with 

their study, our study did not find statistically 

significant associations.

Jelliffe-Pawlowski et al. reported that 

women with PbBs ≥ 10 μg/dL were approximately three times as likely to experience a 

preterm delivery as women with lead levels 

< 10 μg/dL (aOR = 3.2; 95% CI, 1.2–7.4) and 

that their risk of having a small-for-gestationalage infant was more than four times that of 

women with lead levels < 10 μg/dL (aOR = 

4.2; 95% CI, 1.3–13.9) (Jelliffe-Pawlowski 

et al. 2006). Chen et al. (2006), in a study of 

1,611 mother–infant pairs in Taiwan, China, 

suggested that maternal PbBs of ≥ 10 μg/dL 

were related to a doubling risk in low birth 

weight, preterm birth, and small for gestational age compared with maternal PbBs 

< 10 μg/dL. Highly elevated maternal PbBs 

would be expected to have adverse effects on 

fetal growth.

This study has multiple strengths. For 

example, we used PbBs to measure the 

absorbed dose circulated in the blood through 

various exposure routes and sources for pregnant women, which is more accurate than 

occupation history and other proxy exposure 

measures. By restricting the lead concentrations to < 10 μg/dL, the associations between 

maternal lead level and fetal growth found 

in this study were not influenced by lead 

concentrations > 10 μg/dL, unlike previous 

studies that included lead concentrations 

below and > 10 μg/dL. Because this study was 

based on a statewide registry and the study 

lead concentration was close to the lead distribution among the general population, findings should be more generalizable than those 

based on occupational settings or convenience 

samples. Furthermore, this study had a large 

sample size to detect subtle effects.

A possible limitation is selection bias. We 

found that the mothers in this study were 

younger and less likely to be Caucasian than 

other mothers in upstate New York. The linkage rate of PbB reports with birth certificates 

was higher for mothers 18–19 years of age, 

African Americans, and with low-weight 

births, consistent with the selective screening 

for pregnant women at risk for adverse pregnancy outcome or lead exposure.

Dietary calcium and multiple vitamin use 

during pregnancy could not be controlled, 

as they were not collected on either the birth 

certificates or the HMR. Low dietary calcium intake may increase the gastrointestinal absorption of lead (Bogden et al. 1995). 

Calcium supplementation may reduce the 

lead mobilization from bone during pregnancy and therefore reduce the potential lead 

toxicity (Bellinger 2005; Han et al. 2000). 

Furthermore, residual confounding may exist 

because of the potential misclassification or 

categorization of confounders. For example, 

maternal smoking was recorded as “yes or no” 

in birth certificates. Its sensitivity was 89% 

and specificity was 99% using medical records 

as a gold standard (Roohan et al. 2003). There 

was no detailed information on the duration 

and frequency of smoking.

The results of this study have important 

implications regarding the recommended 

action level for childhood PbB. Although 

10 μg/dL is the current reference level set by 

the CDC (ATSDR 2007), this study suggests 

that maternal PbBs < 10 μg/dL may affect 

fetal growth. This issue is of public health significance; in 2005, the HMR received about 

84,000 reports on women in New York state 

with PbBs < 10 μg/dL, and most of the reports 

were regarding women of reproductive age. 

Our study supports the continuation of lead 

screening during pregnancy, especially among 

women who are at risk because of current 

high-dose exposure, which is recommended 

by the New York State Department of Health 

(New York State Department of Health 

Bureau of Occupational Health 2008b). 

Conclusion

Among pregnant women whose PbB was 

< 10 μg/dL, PbB (square root transformed) 

was inversely associated with birth weight. 

Such findings suggest that the decrease in 

birth weight per 1-μg/dL increase in PbB was 

greater at lower concentrations than at higher 
Maternal low-level lead exposure and fetal growth

Environmental Health Perspectives • volume 118 | number 10 | October 2010 1475

concentrations without evidence of a lower 

threshold of effect. These results are important, given the high prevalence of low-level 

lead exposure among pregnant women and 

the controversy regarding the recommended 

action level for maternal PbB.

References

Alexander GR, Himes JH, Kaufman RB, Mor J, Kogan M. 1996. 

A United States national reference for fetal growth. Obstet 

Gynecol 87:163–168.

Allison P. 1999. Logistic Regression Using the SAS System: 

Theory and Application. Cary, NC:SAS Institute Inc.

Atabek ME, Kurtoglu S, Pirgon O, Uzum K, Saraymen R. 2007. 

Relation of in utero lead exposure with insulin-like growth 

factor-I levels and neonatal anthropometric parameters. 

Int J Hyg Environ Health 210:91–95.

ATSDR (Agency for Toxic Substances and Disease Registry). 

2007. Toxicological Profile for Lead. Case No. 7439-92-1. 

Atlanta, Georgia:ATSDR.

Bellinger D, Leviton A, Rabinowitz M, Allred E, Needleman H, 

Schoenbaum S. 1991. Weight gain and maturity in fetuses 

exposed to low levels of lead. Environ Res 54:151–158.

Bellinger DC. 2005. Teratogen update: lead and pregnancy. Birth 

Defects Research Part A: Clin Mol Teratol 73:409–420.

Bogden JD, Kemp FW, Han S, Murphy M, Fraiman M, 

Czerniach D, et al. 1995. Dietary calcium and lead interact 

to modify maternal blood pressure, erythropoiesis, and 

fetal and neonatal growth in rats during pregnancy and 

lactation. J Nutr 125:990–1002.

Boulet SL, Alexander GR, Salihu HM, Kirby RS, Carlo WA. 2006. 

Fetal growth risk curves: defining levels of fetal growth 

restriction by neonatal death risk. Am J Obstet Gynecol 

195:1571–1577.

Canfield RL, Henderson CR Jr., Cory-Slechta DA, Cox C, 

Jusko TA, Lanphear BP. 2003. Intellectual impairment in 

children with blood lead concentrations below 10 microg 

per deciliter. N Engl J Med 348:1517–1526.

CDC (Centers for Disease Control and Prevention). Third 

National Report on Human Exposure to Environmental 

Chemicals. Pub. No. 05-0570. 2005. Atlanta, GA:CDC, 

National Center for Environmental Health.

Chen PC, Pan IJ, Wang JD. 2006. Parental exposure to lead and 

small for gestational age births. Am J Ind Med 49:417–422.

Dales LG, Urg HK. 1978. An improper use of statistical significance testing in studying covariables. Int J Epidemiol 

7:373–376.

Gonzalez-Cossio T, Peterson KE, Sanin LH, Fishbein E, 

Palazuelos E, Aro A, et al. 1997. Decrease in birth weight 

in relation to maternal bone-lead burden. Pediatrics 

100:856–862.

Han S, Pfizenmaier DH, Garcia E. 2000. Effects of lead exposure 

before pregnancy and dietary calcium during pregnancy 

on fetal development and lead accumulation. Environ 

Health Perspect 108:527–531.

Irgens A. 1998. Reproductive outcome in offspring of parents 

occupationally exposed to lead in Norway. Am J Ind Med 

34:431–437.

Jelliffe-Pawlowski LL, Miles SQ, Courtney JG, Materna B, 

Charlton V. 2006. Effect of magnitude and timing of maternal pregnancy blood lead (Pb) levels on birth outcomes. 

J Perinatol 26:154–162.

Lanphear BP, Hornung R, Khoury J, Yolton K, Baghurst P, 

Bellinger DC, et al. 2005. Low-level environmental lead 

exposure and children’s intellectual function: an international pooled analysis. Environ Health Perspect 113:894–899.

Lin S, Hwang SA, Marshall EG, Marion D. 1998. Does paternal 

occupational lead exposure increase the risks of low birth 

weight or prematurity? Am J Epidemiol 148:173–181.

Magri J, Sammut M, Savona-Ventura C. 2003. Lead and other 

metals in gestational hypertension. Int J Gynecol Obstet 

83:29–36.

March of Dimes Foundation. 2007. Quick References and Fact 

Sheets: Preterm Birth. White Plains, NY:March of Dimes 

Foundation.

Mickey RM, Greenland S. 1989. The impact of confounder 

selection criteria on effect estimation. Am J Epidemiol 

129:125–137.

New York State Department of Health Bureau of Occupational 

Health. 2008a. New York State Heavy Metals Registry 2000 

through 2005. Troy:New York State Department of Health, 

Bureau of Occupational Health. 

New York State Department of Health Bureau of Occupational 

Health. 2008b. Physician’s Handbook on Childhood Lead 

Poisoning Prevention. Troy:New York State Department of 

Health Bureau of Occupational Health. 

New York State Department of Health Wadsworth Center. 

2006. Summary of Proficiency Testing Results in the Blood 

Lead Laboratories. Albany:New York State Department of 

Health, Wadsworth Center.

Parsons PJ, Slavin W. 1993. A rapid Zeeman graphite-furnace atomic-absorption spectrometric method for the 

determination of lead in blood. Spectrochim Acta Part B 

48:925–939.

Potula V. 2005. Report from the CDC: is lead exposure a risk 

factor for bone loss? J Womens Health 14:461–464.

Roohan PJ, Josberger RE, Acar J, Dabir P, Feder HM, 

Gagliano PJ. 2003. Validation of birth certificate data in 

New York state. J Community Health 28:335–346.

Rothenberg SJ, Kondrashov V, Manalo M, Jiang J, Cuellar R, 

Garcia M, et al. 2002. Increases in hypertension and blood 

pressure during pregnancy with increased bone lead levels. Am J Epidemiol 156:1079–1087.

Royston P, Ambler G, Sauerbrei W. 1999. The use of fractional 

polynomials to model continuous risk variables in epidemiology. Int J Epidemiol 28:964–974.

Sowers M. 2002. Blood lead concentrations and pregnancy 

outcomes. Arch Environ Health 57:489–494.

Tellez-Rojo MM, Bellinger DC, Arroyo-Quiroz C, LamadridFigueroa H, Mercado-Garcia A, Schnaas-Arrieta L, et al. 

2006. Longitudinal associations between blood lead 

concentrations lower than 10 μg/dl and neurobehavioral 

development in environmentally exposed children in 

Mexico City. Pediatrics 118:e323–e330.

Torres-Sanchez LE, Berkowitz G, Lopez-Carrillo L, TorresArreola L, Rios C, Lopez-Cervantes M. 1999. Intrauterine 

lead exposure and preterm birth. Environ Res 81:297–301.

Zentner LE, Rondo PHC, Mastroeni SSBS. 2006. Lead contamination and anthropometry of the newborn baby. J Trop 

Pediatr 52:369–371.