P.1 

OAR Box 1126 

Prepped by Ollie Stewart 

Document Number: 

4)11-1-125 

Docket Number: 

A-83-22 

Printed 4/18/2008 9:58:21 AM Sheet 4 of 182 
P.2 

United States 

Environmental Protection 

Agency 

f*~*3-» 

Office of Research and 

Development 

Washington, DC 20460 

\IA 

EPA 600/8-89 049F 

August 1990 

xr-jr- /jtfr 

AEPA Air Quality Criteria for 

Lead: 

Supplement to the 1986 

Addendum 
P.3 

EPA 600/8-89/049F 

August 1990 

MT Quality Criteria ior Leads 

^pplemen t to the 1OT@ Addendum 

Environmental Criteria and Assessment Office 

Office of Health and Environmental Assessment 

Office of Research and Development 

U.S. Environmental Protection Agency 

Research Triangle Park, NC 27711 

Bm 
P.4 

DISCLAIMER 

4. 

This document has been reviewed in accordance with U.S. Environmental Protection 

Agency policy and approved for publication. Mention of trade names or commercial products 

does not constitute endorsement or recommendation for use. 

11 

J 
CONTENTS 

P.5 

Page 

TABLES 

FIGURES 

PREFACE 

ABSTRACT 

AUTHORS 

I. INTRODUCTION 

H. RELATIONSHIP OF BLOOD PRESSURE TO LEAD EXPOSURE 

ffl. NEUROBEHAVIORAL AND GROWTH EFFECTS IN INFANTS 

AND CHILDREN . 

IV. REFERENCES 

iv 

vi 

vii 

viii 

ix 

1 

2 

24 

65 

in 
P.6 

IV 

TABLES 

Number 

1 Coefficients for the natural log of blood lead concentration 

(logPbB) vs. blood pressure (BP) in men with and without 

adjustment for site variables 8 

2 Infants' mental development index scores according to cord 

blood lead group 25 

3 Change in McCarthy general cognitive index (GCI) and 

subscale scores associated with each natural log unitincrease 

in blood lead level 27 

4 Results of multiple regression analyses examining the 

relationship between blood lead level and performance 

on the Bayley Mental Index at 3-24 months of age . 29 

5A The relationship of fetal (maternal and cord) blood lead 

(PbB) levels with later developmental outcomes 34 

5B The relationship of prior and current postnatal blood lead 

(PbB) levels with later developmental outcomes 34 

6 The relationship of prenatal and postnatal log blood lead 

with IQ at age four years, ten months 35 

7 Estimated coefficients of log blood lead concentration from 

simple and multiple regression analyses of McCarthy scores 

at the age of four years 38 

8 Regression of developmental indices on maternal and cord 

blood lead levels 40 

9 Regression of developmental indices at 48 months on 

current and prior blood levels 42 

10 Unexplained correlations between log dentine lead levels 

taking into account test reliability, confounding covariates, 

sample selection factors, and reverse causality via 

pica 46 

J 
P.7 

Number 

11 

TABLES (continued) 

Product moment correlations between maternal or teacher 

behavior ratings and log dentine lead values 

12 Strongest relationships between blood lead measures at 

specified times and later neurobehavioral outcomes as 

detected by prospective studies 

Page 

48 

60 
FIGURES 

P.8 

Number 

Comparison of study results from four larger-scale epidemiology 

studies of lead-blood pressure relationships in adult men ... . 14 

Normal distributions with means of 96 and 100 and standard 

deviations of 16 64 

vi 
P.9 

PREFACE 

This document updates the Addendum to the 1986 EPA Air Quatity Criteria for Lead. 

A draft of this Supplement was prepared in March 1989 and reviewed by the Clean Air 

Scientific Advisory Committee (CASAC) at a public meeting on April 27, 1989. Comments 

on the draft Supplement were received from the public until June 12, 1989. The CASAC 

informed the EPA Administrator in a letter dated January 13, 1990 that the 1986 Lead 

Criteria Document, Addendum, and Supplement afforded "a scientifically balanced and 

defensible summary of our current knowledge of the effects of this pollutant, providing an 

adequate scientific basis for EPA to retain or revise primary and secondary NAAQS [National 

Ambient Air Quality Standards] for airborne lead." This draft final Supplement differs from 

the 1989 draft only in minor corrections and additions. 

vn 
P.10 

ABSTRACT 

The 1986 U.S. EPA document Air Quality Criteria for Lead (EPA-600/8-83/028 

aF-dF) evaluated in detail the most pertinent scientific information concerning sources, routes, 

and levels of lead exposure and associated health effects and potential risks. An Addendum, 

issued as part of Volume 1, pages A1-A67, of the 1986 Criteria Document, focused on 

additional emerging studies concerning the effects of lead on cardiovascular function and on 

early physical and neurobehavioral development. This Supplement to the above materials 

evaluates still newer information concerning (1) lead effects on blood pressure and other 

cardiovascular endpoints, and (2) the effects of lead exposure prenatally and/or postnatally on 

birth outcomes and early physical and neurobehavioral development of children. The 

evaluations contained in this Supplement and the 1986 Criteria Document and Addendum are 

to serve as scientific inputs to decision-making with regard to the review and revision, as 

appropriate, of the National Ambient Air Quality Standards (NAAQS) for lead. 

vm 
P.11 

AUTHORS 

Dr. J. Michael Davis, Environmental Criteria and Assessment Office (MD-52), OHEA, ORD, 

U.S. Environmental Protection Agency, Research Triangle Park, NC 27711 

Dr. Lester D. Grant, Environmental Criteria and Assessment Office (MD-52), OHEA, ORD, 

U.S. Environmental Protection Agency, Research Triangle Park, NC 27711 

Dr. Winona W. Victery, Environmental Criteria and Assessment Office (MD-52), OHEA, 

ORD, U.S. Environmental Protection Agency, Research Triangle Park, NC 27711 

REVIEWERS 

Dr. David Bellinger, Harvard Medical School, Children's Hospital, Boston, MA 02115 

Dr. Kim N. Dietrich, Institute of Environmental Health, University of Cincinnati, College of 

Medicine, Cincinnati, OH 45276-0056 

Dr. David J. Svendsgaard, Health Effects Research Laboratory (MD-55), OHR, ORD, U.S. 

Environmental Protection Agency, Research Triangle Park, NC 27711 

ix 
P.12 

SUPPLEMENT TO THE 1986 AIR QUALITY CRITERIA FOR LEAD; ADDENDUM 

I. INTRODUCTION 

In the mid-1980s, the 1977 EPA criteria document, Air Quality Criteria for Lead (U.S. 

Environmental Protection Agency, 1977) was updated and revised pursuant to Sections 108-

109 of the Clean Air Act, as amended, 42 U.S.C. 7408 and 7409. The final version of the 

updated criteria document (U.S. Environmental Protection Agency, 1986a), incorporating 

revisions made in response to public comments and review of earlier drafts by the Clean Air 

Scientific Advisory Committee (CASAC), was completed in 1986 to be used as a basis for 

review and, as appropriate, revision of the National Ambient Air Quality Standard (NAAQS) 

for lead. An Addendum (U.S. Environmental Protection Agency, 1986b) to the revised 

document, Air Quatity Criteria for Lead (U.S. Environmental Protection Agency, 1986a), was 

also completed in 1986 and evaluated newly published information concerning two topics: (1) 

the relationship between blood lead levels and cardiovascular effects, and (2) lead exposure 

effects on early development and stature. 

In the three years since the 1986 Addendum was prepared, information pertaining to the 

health effects of lead has continued to emerge with regard to the topics addressed in that 

Addendum. Although newer findings have been generally consistent with the state of 

understanding that was articulated in the 1986 Addendum, they still need to be evaluated as 

part of an updated assessment of the latest scientific information that characterizes the health 

effects of lead (Pb) with clear relevance for decision-making on potential revision of the 

existing lead NAAQS. 

The present update focuses primarily on key findings that have emerged in the literature 

since 1986 in the areas of lead effects on (1) blood pressure and related cardiovascular 

endpoints and (2) child development. It does not attempt to be comprehensive in reviewing 

this literature, nor does it attempt to address all the topics covered in the 1986 Air Quality 

Criteria Document and Addendum. Rather, its explicit purpose is to describe and interpret 

the critical effects of lead that have greatest significance for impending regulatory decisions 

regarding this environmental pollutant, with discussion focussing on key newer papers 

published since completion of the earlier 1986 Addendum. 

1 
P.13 

II. RELATIONSHIP OF BLOOD PRESSURE TO LEAD EXPOSURE 

The 1986 Addendum (U.S. Environmental Protection Agency, 1986b) addressed the 

issue of lead effects on the cardiovascular system in a review of the findings with regard to 

overtly lead-intoxicated individuals, epidemiologic studies of associations between lead 

exposure and increased blood pressure, toxicologic data providing evidence for lead-induced 

cardiovascular effects in animals, and information on the possible mechanisms of action of 

lead on cardiovascular function. The amount and quality of the evidence available at that 

time were adequate to provide an initial evaluation of the topic, but did not come to a fully 

definitive consensus with regard to the contributory role of lead exposure to hypertension. 

Also not fully resolved was the extent to which lead-induced hypertension or other types of 

lead-induced pathogenic effects contribute to more serious morbidity (heart attack, stroke) and 

mortality of the human population. 

For the purposes of this Supplement, primary emphasis is being placed on updated 

discussions of key human population-based studies which reflect primarily non-occupational 

exposure to lead. The largest study populations have been the second National Health and 

Nutrition Examination Survey (NHANES II) for the U.S. population (performed during the 

years 1976-80) and the British Regional Heart Study (BRHS), an ongoing evaluation of men 

aged 40 to 59 from 24 British towns. These studies were earlier described in detail in the 

published literature (Harlan et al., 1985; Pocock et al., 1984; Pirkle et al., 1985; Shaper 

et al., 1981); and further analyses of the subject data sets have been presented and discussed 

extensively at a 1987 U.S. EPA co-sponsored International Symposium on Lead-Blood 

Pressure Relationships and in other recent publications. In addition, analyses of several other 

data sets have been presented by other investigators at the 1987 Symposium or elsewhere in 

the published literature since the 1986 Addendum was prepared. 

The 1986 Addendum (U.S. Environmental Protection Agency, 1986b) noted that several 

then-recently-published studies provided generally consistent evidence for increased blood 

pressure being associated with elevated lead body burdens in adults, especially as indexed by 

blood lead levels in various cohorts of adult men. None of the individual studies, it was 

noted, provided definitive evidence establishing causal relationships between lead exposure 

and increased blood pressure, but they collectively provided considerable qualitative evidence 

indicative of significant associations between blood lead and blood pressure levels. It was 
P.M 

fiujher emphasized that estimates of quantitative relationships between blood lead levels and 

blood pressure increases derived from such study results are subject to much uncertainty, 

given the relatively small sample sizes and limited population groups typically studied. On 

the other hand, the above-noted larger-scale studies (NHANES n and BRHS) of general 

population groups, were singled out as providing reasonably good bases for estimation of 

quantitative blood-lead/blood-pressure relationships. 

As reviewed in the 1986 Addendum (U.S. Environmental Protection Agency, 1986b), in 

the BRHS, Pocock et al. (1984) evaluated relationships between blood lead concentrations, 

hypertension, and renal function indicators in a clinical survey of 7,735 middle-aged men 

(aged 40-49) from 24 British towns. Each man's blood pressure, while seated, was measured 

twice in succession by means of a London School of Hygiene sphygmomanometer. Diastolic 

pressure was recorded at phase V disappearance of sounds. The mean of the two readings of 

blood pressure was adjusted for observed variation within each town to correct for any 

differences among three observers. Results for 7,371 men included in data analyses indicated 

correlation coefficients of r = +0.03 and r = +0.01 for associations between systolic and 

diastolic blood pressure, respectively, and blood lead levels. The systolic blood pressure 

correlation, though small in magnitude, was nevertheless statistically significant at p <0.01. 

However, analyses of covariance using data for men categorized according to blood lead 

concentrations only suggested increases in blood pressure at lower blood lead levels; no 

further significant increments in blood pressure were observed at higher blood lead levels 

either before or after adjustment for factors such as age, town, body mass index, alcohol 

consumption, social class, and observer. Evaluation of prevalence of hypertension, defined as 

systolic blood pressure over 160 mm mercury (Hg), revealed no significant overall trend; but 

of those men with blood lead levels over 37 /ug/dl, a larger proportion (30 percent) had 

hypertension when compared with the proportion (21 percent) for all other men combined 

(p = 0.08). Similar results were obtained, for diastolic hypertension defined as > 100 mm 

Hg, i.e., a greater proportion (15 percent) of men with blood lead levels over 37 jxg/dl had 

diastolic hypertension in comparison with the proportion (9 percent) for all other men (p = 

0.07). 

Pocock et al. (1984) interpreted their findings as being suggestive of increased 

hypertension at blood lead levels over 37 /tg/dl, but not at lower concentrations typically 
P.15 

found in British men. However, further analyses reported by Pocock et al. (1985) for the 

same data indicated sigmfisantly higher statistical associations between both systolic 

(p =0.003) and diastolic (p < 0.001) blood pressure and blood lead levels, when adjustments 

were made for variation due to site (town) in multiple regression analyses. The regression 

coefficients for log blood lead versus systolic and diastolic pressure were +2.089 and 

+1.809, respectively, when adjusted for site as well as body mass, age, alcohol, smoking, 

social class, and observer. Noting the small magnitude of the association observed and the 

difficulty in adjusting for all potentially relevant confounders, Pocock et al. (1985) cautioned 

at that time against prematurely concluding that elevated body lead burden has a causal 

influence on blood pressure. 

The 1986 Addendum (U.S. Environmental Protection Agency, 1986b) also noted that 

relationships between blood lead and blood pressure among American adults had begun to be 

evaluated in another large-scale study, as reported by Harlan et al. (1985), Pirkle et al. 

(1985), and Schwartz (1985a,b; 1986a,b). These analyses were based on evaluation of 

NHANES n data, which provide careful blood lead and blood pressure measurements on a 

large-scale sample representative ofthe U.S. population and considerable information on a 

wide variety of potentially confounding variables as well. As such, these analyses avoided 

the problem of selection bias, the healthy-worker effect, workplace exposures to other toxic 

agents, and problems with appropriate choice of control groups that often confounded or 

complicated earlier, occupational studies of blood-lead/blood-pressure relationships. Three 

blood pressure readings were recorded for each subject: while seated early in the 

examination, supine midway in the examination, and seated near the end. First and fifth 

phase sounds were taken as systolic and diastolic pressures, respectively. The second seated 

blood pressure was used in statistical analyses, but analyses using the first seated pressure or a 

mean of the first and second seated pressure yielded similar results. Blood lead values, 

determined by atomic absorption spectrometry, were transformed to log values for statistical 

analyses. 

Relationships between blood pressure and other variables were evaluated in two ways. 

First, men and women were stratified into normotensive and hypertensive categories and 

mean values for relevant variables contrasted across the categories. For ages 21-55 yr, 

diastolic high blood pressure (£90 mm Hg) male subjects (N = 475) had significantly 
P.16 

(p < 0.005) higher PbB levels, body mass index values, and calcium foods than did 

normotensive male subjects (N = 1,043). Similar results were obtained for aged 21-55 yr 

diastolic high blood pressure females (N = 263) in comparison to normotensive females 

(N = 1,316). For subjects aged 56-74 years, significantly (p <0.05) higher PbB levels were 

found for female subjects (but not males) defined as having isolated systolic high blood 

pressure (i.e., systolic t 160 and diastolic <90 mm Hg). Simple conelation analyses and 

step-wise multiple regression analyses were carried out as a second statistical evaluation 

approach; PbB values were entered into predictive models for systolic and diastolic pressure 

as well as several other pertinent variables (such as age, body mass index, etc.) entered 

sequentially according to greatest magnitude of variance explained for the dependent variable. 

The simple conelation analyses reported by Harlan et al. (1985) demonstrated statistically 

significant linear associations (p < 0.001) between blood lead concentrations and blood 

pressure (both systolic and diastolic) among males and females, aged 12 to 74 years. Using 

multiple regression analyses controlling for a number of other potentially confounding factors, 

however, the blood-lead/blood-pressure associations remained significant for males but not for 

women after adjusting for the effects of other pertinent variables. 

Additional analyses of NHANES II data reported by Pirkle et al. (1985) focused on 

white males (aged 40 to 59 years) in order to avoid the effects of collinearity between blood 

pressure and blood lead concentrations evident at earlier ages and because of less extensive 

NHANES II data being available for non-whites. In the subgroup studied, Pirkle et al. 

(1985) found significant associations between blood lead and blood pressure even after 

including in multiple regression analyses all known factors previously established as being 

conelated with blood pressure. The relationship also held when tested against every dietary 

and serologic variable measured in the NHANES U study. Inclusion of both curvilinear 

transformations and interaction terms altered Uttle the coefficients for blood pressure 

associations with lead (the strongest relationship was observed between the natural log of 

blood lead and the blood pressure measures). The regression coefficients for log blood lead 

versus systolic and diastolic blood pressure were 8.436 and 3.954, respectively. No evident 

threshold was found below which blood lead level was not significantly related to blood 

pressure across a range of 7 to 34 /ig/dl. In fact, the dose-response relationships 

characterized by Pirkle et al. (1985) indicate that large initial increments in blood pressure 
P.17 

occur at relatively low blood lead levels, followed by leveling off of blood pressure 

increments at higher blood: lead levels. Pirkle et al. (1985) also found lead to be a significant 

predictor of diastolic blood pressure greater than or equal to 90 mm Hg, the criterion blood 

pressure level now standardly employed in the United States to define hypertension. 

Additional analyses were performed by Pirkle et al. (1985) to estimate the likely public health 

implications of their findings concerning blood-lead/blood-pressure relationships. Changes in 

blood pressure that might result from a specified change in blood lead levels were first 

estimated. Then coefficients from the Pooling Project and Framingham studies (Pooling 

Project Research Group [1978] and McGee and Gordon [1976], respectively) of 

cardiovascular disease were used as bases to: (1) estimate the risk for incidence of serious 

cardiovascular events (myocardial infarction, stroke, or death) as a consequence of leadinduced blood pressure increases, and (2) predict the change in the number of serious 

outcomes as the result of a 37 percent decrease in blood lead levels for adult white males 

(aged 40-59 years) observed during the course of the NHANES II survey (1976-1980). 

The earlier Addendum (U.S. Environmental Protection Agency, 1986b) also noted that 

questions had been raised by Gartside (1985) and E. I. DuPont de Nemours (1986) regarding 

the robustness of the findings derived from the analyses of NHANES n data discussed above 

and as to whether certain time trends in the NHANES n data set may have contributed to (or 

account for) the reported blood-lead blood-pressure relationships. Gartside reported analyses 

of NHANES II data, which found that the obtained size and level of coefficient statistical 

significance varied depending upon specific data aggregations used in analyzing the data. The 

largest and most significant coefficients for blood lead versus blood pressure were obtained by 

Gartside for data aggregated by age groups that approximated that of the 40-59 yr male 

aggregation described by Pirkle et al. (1985), with coefficients for most younger cohort 

groups aggregated by varying 20-yr age intervals (e.g., 21-40, 22-41 yrs, etc.) or older 

groups not always being significant at p <0.05. As for the time trend issue, both blood lead 

and blood pressure declined substantially during the 4-yr NHANES JJ study and different 

geographic sites were sampled without revisitation of the same site over the survey period. 

Thus, variations in the sampling sites over time, coincident with changes in blood lead and/or 

blood pressure, might contribute to any observed associations between blood lead and blood 

pressure. E. I. DuPont de Nemours (1986) reported that multiple regression coefficients 
P.18 

decreased in magnitude and some became non-significant (p > 0.05) when adjusted for 

geographic site in analyses of NHANES JJ data, including analyses for the male group (aged 

12-74) reported by Harlan et al. (1985) and for males (aged 40-59) reported by Pirkle et al. 

(1985). For example, E. I. DuPont de Nemours reported unpublished reanalyses of 

NHANES JJ data confirming significant associations, for males aged 12-74 years as well as 

40-59 years, between log blood lead and systolic or diastolic blood pressure unadjusted for 

geographic site, but smaller coefficients (nonsignificant for diastolic) when geographic site 

was included in the analysis. However, neither the Gartside nor E. I. DuPont de Nemours 

analyses adjusted for all of the variables that were selected for stepwise inclusion in the 

Harlan et al. (1985) and Pirkle et al. (1985) published analyses by means of a priori decision 

rules for inclusion of variables having significant associations with blood pressure. Also, 

other differences existed in regard to specific aspects of the modeling approaches employed, 

making it extremely difficult to assess clearly the potential impact of variation in selection of 

age groups and geographic site adjustment on NHANES JJ analyses results. 

In order to address the "site" issue more definitively, Schwartz (1985a,b; 1986a,b) 

carried out a series of additional reanalyses of the NHANES JJ data. Those analyses 

confirmed that the regression coefficients remain significant for both systolic and diastolic 

blood pressure when site is included as a variable in multiple regression analyses. Of several 

different approaches used by Schwartz, the most direct was holding all aspects of the original 

Pirkle et al. (1985) analyses the same except for the addition of a variable controlling for the 

64 geographic sites sampled in NHANES IJ. Using this approach, the coefficients for log 

blood lead in relation to either diastolic or systolic blood pressure dropped somewhat from 

those of the original analyses when site was controlled for (i.e., from 8.44 to 5.09 for systolic 

and from 3.95 to 2.74 for diastolic blood pressure), but the coefficients for each still 

remained significant at p <0.05. When still other approaches were used to control for site 

along with variations in other variables included in the analyses, statistically significant results 

were still consistently obtained both for males aged 40-59 and for males aged 20-74. The 

results obtained by Schwartz (1985a,b; 1986a,b) via reanalysis of NHANES II data 

(unadjusted versus adjusted for geographic site) are presented in Table 1 in comparison to 

results reported by E. I. DuPont de Nemours (1986) and in relation to the findings presented 

by Pocock et al. (1984, 1985) for British men (also unadjusted versus adjusted for site). 
P.19 

TABLE 1. COEFFICIENTS FOR THE NATURAL LOG OF BLOOD LEAD 

CONCENTRATION (logPfoB) VS. BLOOD PRESSURE (BP) IN MEN WITH AND 

WITHOUT ADJUSTMENT FOR SITE VARIABLES 

Analysis 

Performed by 

Pocock et al. 

(1984, 1985) 

Schwartz (1985a,b) 

E. I. DuPont 

de Nemours (1986) 

Schwartz (1986a,b) 

E. I. DuPont 

de Nemours (1986) 

. i II.IHI^- ^ i^-,-,,.-,-,-,.-,.,,.-.,^ MIW m u i i l.ll . 1 1 

Study 

Group 

British Regional 

Heart Study 

White males aged 40-59 

Systolic (n=7371) 

Diastolic (n=7371) 

NHANES n 

Males aged 20-74 

Systolic (n=2254) 

Diastolic (n=2248) 

NHANES n 

Males aged 12-74 

Systolic (n=2794) 

Diastolic (n=2789) 

NHANESD 

White males aged 40-59 

Systolic (n=543) 

Diastolic (n=565) 

NHANESH 

White males aged 40-59 

Systolic (n=553) 

Diastolic (n=575) 

Coefficient of 

log PbB vs. BP 

Unadjusted 

for Site 

1.68** 

0.30 

5.23*** 

2.96*** 

3.43*** 

2.02*** 

8.44** 

3.95** 

6.27** 

4.01** 

Adjusted 

for Site 

2.09** 

1.81*** 

3.23** 

1.39* 

1.95* 

0.36 

5.01* 

2.74* 

3.46* 

1.93* 

*p <0.05 

**p <0.01 

***p < 0.001 

The 1986 Addendum (U.S. Environmental Protection Agency, 1986b) concluded that, 

overall, the analyses of data from the two large-scale general population studies (British 

Regional Heart Study and U.S. NHANES JJ Study) discussed above collectively provide 

highly convincing evidence demonstrating small but statistically significant associations 

between blood lead levels and increased blood pressure in adult men. The strongest 

associations appeared to exist for males aged 40-59 and for systolic somewhat more so than 

8 
for diastolic pressure. Virtually all of the analyses revealed positive associations for the 40-

59 aged group, which remain or become significant (at p <0.05) when adjustments are made 

for geographic site. Furthermore, the results of these large-scale studies were noted to be 

consistent with similar findings of statistically significant associations between blood lead 

levels and blood pressure increases as derived from other smaller-scale studies discussed in 

the Addendum, which also mainly found stronger associations for systolic pressure than for 

diastolic. 

The Addendum (U.S. Environmental Protection Agency, 1986b) further concluded that 

none of the observational studies in and of themselves can be stated as definitively 

establishing causal linkages between lead exposure and increased blood pressure of 

hypertension. However, the Addendum noted that the plausibility of the observed 

associations reflecting causal relationships between lead exposure and blood pressure increases 

was supported by: (1) the consistency of the significant associations found by numerous 

independent investigators for a variety of study populations, and (2) extensive toxicological 

data discussed in the 1986 Addendum which clearly demonstrated increases in blood pressure 

for animal models under well-controlled experimental conditions. The precise mechanisms 

underlying relationships between lead exposure and increased blood pressure, the Addendum 

further stated, appeared to be complex, and mathematical models describing the relationships 

still remained to be more definitively characterized. At the time of the Addendum, log bloodlead/blood pressure models appeared to fit best the available data, but linear relationships 

between blood lead and blood pressure could not be ruled out. The most appropriate 

coefficients characterizing these relationships also remained to be more precisely determined, 

although those listed in Table 1 obtained by analyses adjusting for site appeared to be the 

cunently best available and most reasonable estimates of the likely strength of the association 

for adult men (i.e., generally in the range of 2.0-5.0 for log blood lead versus systolic and 

1.4 to 2.7 for log blood lead versus diastolic blood pressure). The 1986 Addendum went on 

to note that the full range of blood lead levels that may be associated with increased blood 

pressure also remained to be more clearly defined. However, the collective evidence from 

the above studies points toward low or moderately elevated blood lead levels as being 

associated with blood pressure increases, with certain evidence (e.g., the NHANES II data 
P.21 

analyses and some other study results) also indicating significant relationships between blood * 

pressure elevations and blood lead levels ranging down, possibly, to as low as 7 jtg/dl. 

The earlier Addendum (U.S. Environmental Protection Agency, 1986b) further stated 

that the quantification of likely consequent risks for serious cardiovascular outcomes, as 

attempted by Pirkle et al. (1985), also remained to be more precisely characterized. The 

specific magnitudes of risk obtained for serious cardiovascular outcomes in relation to lead 

exposure, estimated on the basis of lead-induced blood pressure increases, depend crucially 

upon: the form of the underlying relationship and size of the coefficients estimated for bloodlead/blood-pressure associations; lead exposure levels at which significant elevations in blood 

pressure occur; and coefficients estimating relationships between blood pressure increases and 

specific more serious cardiovascular outcomes. It was noted that uncertainty still exists 

regarding the most appropriate model and blood-lead/blood-pressure coefficients, mading it 

difficult to resolve which specific coefficients should be used in attempting to project more 

serious cardiovascular outcomes. Similarly, it was indicated that it is difficult to determine 

appropriate blood lead levels at which any selected coefficients might be appropriately applied 

in models predicting more serious cardiovascular outcomes. Lastly, it was noted that the 

selection of appropriate models and coefficients relating blood pressure increases to more 

serious outcomes is also fraught with uncertainty. Questions exist regarding the general 

applicability of coefficients derived from the Pooling Projects and Framingham Study to men 

aged 40-59 in the general U.S. population. Further analyses of additional large-scale 

epidemiologic data sets, it was stated in the Addendum, would be necessary to determine 

more precisely quantitative relationships between blood-lead and blood-pressure, and more 

serious cardiovascular outcomes as well. 

The Addendum (U.S. Environmental Protection Agency, 1986b) went on to note further 

that the above findings, while pointing toward a likely causal effect of lead in contributing to 

increased blood pressure, need to be placed in broader perspective in relation to other factors 

involved in the etiology of hypertension. The underlying causes of increased blood pressure 

or "hypertension" (diastolic blood pressure above 90 mm Hg), which occurs in as many as 

25 percent of Americans, are not yet fully delineated (Frohlich, 1983; Kaplan, 1983). 

However, it is very clear that many factors contribute to development of this disease, 

including hereditary traits, nutritional factors, and environmental agents. The relative roles of 

10 
P.22 

various dietary and environmental factors in influencing blood pressure and the mechanisms 

by which they do so are a matter of intense investigative effort and debate (see proceedings of 

conference "Nutrition and Blood Pressure: Cunent Status of Dietary Factors and 

Hypertension," McCanon and Kotchen, 1983). The contribution of lead, compared to many 

other factors evaluated in various analyses discussed above, appears to be relatively small, 

usually not accounting for more than 1-2 percent of the variation explained by the models 

employed when other significant factors are controlled for in the analyses. 

Many of the above issues and questions have been further addressed since the 1986 

Addendum was completed, in part at the 1987 Lead-Blood Pressure Symposium and in other 

recent publications as well. Presentations and discussions at the 1987 Symposium (see 

Victery, 1988a) were contributed by (1) many of the key scientists who have carried out 

important investigations of the relationship of lead exposure and hypertension (including 

observational, epidemiologic, and experimental reports), as well as (2) other experts in 

hypertension and cardiovascular disease in general. Several speakers reviewed their analyses 

of the NHANES II data (Harlan, 1988; Schwartz, 1988; Gartside, 1988; Landis and Flegal, 

1988) and the British study (Pocock et al., 1988). Elwood et al. (1988a) also summarized the 

results of two new large-scale Welsh studies. Pocock further provided a new comparison of 

the relative magnitude of the contribution of lead exposure to blood pressure changes based on 

intercomparison of the results of the NHANES II, British, and Welsh studies. A number of 

other relevant papers described studies of lead-blood pressure relationships in selected 

populations of nonoccupationally exposed individuals, as well as workers (Kromhout, 1988; 

Moreau et al., 1988; Weiss et al., 1988; Wedeen, 1988; Cooper, 1988; Selevan et al., 1988; 

de Kort and Zwennis, 1988; Elwood et al., 1988; Neri et al., 1988; Staessen et al., 1988; 

Sharp et al., 1988). Other presentations addressed pertinent toxicological findings from in 

vivo or in vitro animal studies (Vander, 1988; Chai and Webb, 1988; Kopp et al., 1988; 

Boscolo and Carmignani, 1988; Weiler et al., 1988; Victery, 1988b). 

With regard to new findings reported at the 1987 Symposium, a further alternative 

analysis of the data from NHANES JJ (using a generalized Mantel-Haenzel test) was reported 

by Landis and Flegal (1988). In order to more definitively assess the robustness of the earlier 

NHANES JJ results (Harlan et al. 1985; Pirkle et al., 1985) and, also, to evaluate possible 

time-trend effects confounded by variations in sampling sites, Landis and Flegal (1988) 

11 
P.23 

carried out further analyses for NHANES JJ males, aged 12-74, using a randomization mode.1-

based approach to test the statistical significance of the partial conelation between blood lead 

and diastolic blood pressure, adjusting for age, body mass index, and the 64 NHANES II 

sampling sites. Simple linear and multiple regression coefficients between log blood lead and 

diastolic blood pressure for all males (aged 12-74) were 0.15 and 4.90, respectively; for 

various groups broken out by age (<20, 21-39, >40 yrs) and body mass index levels, the 

respective coefficients ranged from 0.04 to 0.12 and from 1.29 to 3.55 (predominantly 

between 2.3 and 3.6), displaying considerable consistency across age-body mass comparison 

groups. Also, the most stringent or "conservative" approach used to calculate a randomized 

model statistic controlling for effects due to 64 sampling sites yielded a test statistic of 4.62 

(still significant at p <0.05). The authors noted that: (1) the association must be sufficiently 

robust to persist across 478 subgroups formed on the basis of factors also having an 

association with the levels of blood lead and blood pressure; and (2) nevertheless, even with 

the severe adjustments by sampling sites, age, and Quetelet index, the diastolic blood 

pressure/blood lead relationship remained statistically significant at the p <0.05 level. In 

summary, their analyses indicate that the significant linear association between blood-lead 

levels and diastolic blood pressure readings cannot be dismissed as due to concunent secular 

trends in the two variables across the 4-year survey period. Finally, the authors noted that, 

even though these partial conelations are not large, the magnitude of the regression 

coefficients suggests that elevated blood lead levels may be an important risk factor for 

elevated blood pressures, as developed in considerably greater detail for the restricted group 

of white men 40 to 59 years of age (as described in Pirkle et al., 1985). 

Also reported at the 1987 Symposium were two surveys in Wales that evaluated possible 

relationships between blood lead and blood pressure (Elwood et al., 1988a,b). The Welsh 

Heart Programme was carried out in 1985 throughout Wales, using a stratified cluster random 

sample of 21,000 households, with 2,010 male and female adult subjects being included in the 

survey population for whom blood lead determinations were made. Complete blood lead, 

blood pressure, and other key data were available for 865 men and 856 women. Mean blood 

lead values for men were 12.4 /xg/dl and for women 9.6 jtg/dl. Blood lead increased with 

age (1 /xg/dl every 15.6 years in men and every 13.1 years in women). The partial regression 

coefficients with standard enor were reported after the effect of age had been removed by 

12 
P.24 

cubic regression. In males, the coefficients for systolic and diastolic pressure were 0.82 ± 

1.49 and 1.29 ± 0.95, respectively. For females, these coefficients were 0.19 ± 1.46 and 

0.54 + 1.00. In neither sex was the relationship statistically significant. Categorizing the 

data by blood lead increments demonstrated no trend in the blood pressure levels. 

The second group studied was about half of the planned subjects in the Caerphilly 

Collaborative Heart Disease study; this study on ischemic heart disease was based on a cohort 

of men aged 45-59, living in Caerphilly, a small town in Wales. Several risk factors for 

ischemic heart disease were considered. Blood pressure was measured with a random-zero 

muddler sphygmomanometer after a 5-minute rest and during a cold pressor test. The 

conelation coefficients for resting systolic and diastolic pressure were 0.0183 and 0.0230, 

respectively, and for the rise in pressure with the cold pressor test, 0.0342 and 

0.0078 respectively. None were significant at p < 0.05. Ranking the blood pressure readings 

according to blood lead groups did not reveal any trend in the percentage of subjects with 

systolic blood pressure greater than 160 mm Hg. 

The authors (Elwood et al., 1988a,b) only conected for age as a confounding factor, 

because all other factors were positively conelated with both blood pressure and blood lead. 

They concluded that, in the event these other factors (e.g., cigarette use and alcohol 

consumption) were controlled for, they would probably reduce the already trivial (and 

nonsignificant) relationships found by them for blood lead and blood pressure measurements 

in the two Welsh studies. 

At the 1987 Symposium, Pocock compared the results of the NHANES JJ data analyses 

as discussed by Schwartz (1988), his analyses of the British study (Pocock et al., 1988), and 

the two Welsh studies (Elwood et al., 1988a,b). Figure 1 shows the magnitudes of effects 

obtained for adult men in these four large-scale studies, relating systolic blood pressure to 

blood lead concentration, as depicted by Pocock etal. (1988). The graph shows each study's 

estimated change in mean systolic blood pressure for each doubling of blood lead (e.g., from 

8 to 16 ftg/dl), together with 95% confidence limits. Pocock concluded that an overview of 

data from these large epidemiological surveys provides reasonably consistent evidence on lead 

and blood pressure. That is, whereas the NHANES JJ data on 2,254 U.S. men indicate a 

slightly stronger association between blood lead and systolic blood pressure than Pocock's 

13 
BRHS (N = 7371) 

NHANES II (N = 2254) 

Caerphilly (N =1164) 

Wales (N = 865) 

• l 

I 

| I 

r "- • 

• 1 

l 

1 

• C 

D 

— - ~ " i 

D 

A 

1 

n 1 

B 

i 

i 

i i 

r- 1 -

- 2 - 1 

ESTIMATED C 

(nunn 

0 

GE 

1 

*J MEAN SYSTOLIC BLOOD PRESSURE 

DOUBLING OF BLOOD LEAD 

Figure 1. Compaiiisom of study results from four larger-scale epidemiology studies of 

lead-blood pressure relationships in adult men. Depicted are each study's 

estimated changes in mean systolic blood pressure (in mm Hg) for each 

doubling of blood lead (e.g., from 8 to 16 jig/dl) together with 95% confidence 

limits. A = British Regional Heart Study (BRHS) analyses, described by 

Pocock et al. (1988); B = National Health and Nutrition Evaluation Survey 

(NHANES H) analyses described by Schwartz (1988); C and D = Caerphilly 

and Wales'studies described by Elwood et al. (1988a,b). 

Source: Pocock et al. (1988). 

8/16/90 14 
P.26 

British study, the data frqm the two Welsh studies on over 2,000 men also showed a small 

positive (but not statistically significant) association. Pocock noted that the overlapping 

confidence limits for all these studies suggest that there may be a weak positive statistical 

association whereby systolic blood pressure in adult men is increased by about 1 mm Hg for 

every doubling of blood lead concentration. 

There was an additional report of an analysis of cross-sectional data from Canada (Neri 

et al., 1988) presented at the 1987 Lead-Blood Pressure Symposium. This study was 

performed on data collected during 10 months of 1978-1979 for 2,193 subjects aged 25 to 64. 

The zero-order conelation between diastolic blood pressure and blood lead level was found to 

be 0.115. The authors concluded that, although this association is a weak one (as was also 

the case in the NHANES II data), its statistical significance is not in doubt (p <0.001). 

They further concluded that (1) the Canadian data were at least weakly supportive of the 

inference drawn from NHANES II, in that elevation of blood lead did seem to entail some 

risk of blood pressure elevation, but (2) it would be premature, in the absence of longitudinal 

data, to infer that this is a cause-and-effect relationship. Neri et al. (1988) then reported the 

findings for a group of subjects studied longitudinally for blood lead levels and blood 

pressure. In this study of lead foundry workers, an association was found between shortterm changes in an individual's blood lead level and contemporary changes in diastolic 

pressure, which remained significant after allowance for age (or time) trends and for effects 

attributable to changes in body weight. The authors also noted that short-term changes in 

urinary cadmium were similarly predictive of diastolic blood pressure levels. 

There were several other reports presented at the 1987 Symposium with regard to 

smaller-scale observational studies of blood lead and blood pressure in two types of 

occupationaliy lead-exposed groups: bus drivers (Shaip et al., 1988) and policemen (Weiss 

et al., 1988; Moreau et al., 1988). These studies all yielded findings that are also indicative 

of lead-induced increases in blood pressure. 

In other recent reports besides those presented at the 1987 Symposium, several new 

studies (with generally small cohorts) of occupationaliy exposed workers have yielded further 

results with regard to lead effects on blood pressure or other cardiovascular outcomes. For 

example, Parkinson et al. (1987) reported findings on data collected in 1982 from 270 lead 

workers and 158 nonexposed workers in Pennsylvania. Four measures of lead exposure were 

15 
P.27 

used: (a) employment at lead exposure vs. control plants; (b) cunent blood lead values at 

time of examination (exposed group mean = 39.9 /*g/dl vs. 7.4 /ig/dl for controls); (c) zinc 

protoporphyrin values at time of examination: (d) time weighted average (TWA) blood lead 

values for lead-exposed workers since date of hire. Other risk factors (age, years of 

education, etc.) were also included in the regression analysis. Of the three lead measures, 

only TWA blood lead was significantly, although modestly, conelated with blood pressure. 

After controlling for other predictive risk factors, including age, the effect of TWA on blood 

pressure was no longer significant and the authors did not find evidence of renal disease. 

Also, in 1987, de Kort and colleagues reported finding a statistically significant increase 

in systolic and diastolic pressure in 53 exposed males compared with 52 controls (de Kort 

et al., 1987). Blood lead averaged 47.4 jig/dl for the exposed workers vs. 8.1 for controls. 

Blood and urine cadmium were also higher than in controls; there were no adverse effects on 

kidney function. The prevalence of clinically defined hypertension (systolic greater than 160 

mm Hg and/or diastolic greater than 95 mm Hg and/or under treatment for hypertension) was 

higher in the exposed group, but the observed relative risk was not different. 

On the other hand, Sharp et al. (1989) did find a statistically significant (p <0.036) 

relationship between blood lead levels and blood pressure (diastolic but not systolic) levels 

among a subsample of bus drivers evaluated in their earlier study (Sharp et al., 1988) who 

were under beta-blocker treatment for hypertension. The association between diastolic blood 

lead and blood pressure was significant at the stated level even after controlling for other 

relevant risk factors, and none of the blood lead levels at the time of blood pressure testing 

exceeded 18 jtg/dl. 

In addition to the above study results which focused primarily on lead-blood pressure 

effects in adult males, some additional analyses of such relationships in females have become 

available since the 1986 Addendum and the 1987 Lead-Blood Pressure Symposium. For 

example, Schwartz (1989) has recently reported the results of still further analyses of the 

NHANES JJ data set. These latest analyses used variables identified as being important in 

earlier NHANES U analyses and examined the relationship between blood lead and blood 

pressure in both males and females, aged 20-74 years. The analyses were carried out in 

stages, with separate regressions first being performed for males and females. Diastolic blood 

pressure was regressed on age, age squared, Quetelet's body mass index, and the natural log 

16 
P.28 

of .blood lead. If, in this first step, lead was significant, then a stepwise regression was 

carried out taking into account a number of covariates or potentially confounding cofactors 

(e.g., dietary sodium, smoking), and final models were then ultimately estimated using 

SURREGR (a program conecting for study design effects as well). Statistically significant 

relationships were found between blood lead and blood pressure in both males (p <0.01) and 

females (p <0.01) in the first stage regressions, conecting for the above-noted covariates. 

Lead also continued to be significantly related to blood pressure in the separate stepwise 

regressions for males and females. The final models yielded a smaller but statistically 

significant (p <0.03) blood lead coefficient (1.640) for females than the coefficient for males 

(2.928; p < 0.006). These results therefore (1) tend to confirm the earlier findings of 

Schwartz and other investigators demonstrating significant associations between blood lead 

and blood pressure increases in adult males in the NHANES II study population, and 

(2) demonstrate analogous although somewhat smaller effects in women from the same study 

population. However, the strength of the relationship (blood lead coefficient) for women 

remains to be more precisely defined in terms of adjustments for geographic site effects 

analogous to those described above for earlier analyses for NHANES JJ and BRHS male 

subjects. 

In another study, Grandjean et al. (1989) evaluated lead-hypertension relationships in a 

cohort of 504 men and 548 women born in the same year and residing in the Glostrup area of 

Denmark. Both blood lead concentrations and blood pressure determinations were first 

obtained at age 40 and then, again, five years later for 451 men and 410 of the women. The 

average blood lead levels for men at age 40 and 45 years were 13 and 9 A*g/dl, respectively; 

for women they were 9 and 6 /*g/dl, respectively, at age 40 and 45 years. At age 40, women 

with systolic blood pressure above 140 mm Hg and/or diastolic above 90 mm Hg had slightly 

elevated blood leads compared to non-hypertensive women; but no differences were found for 

blood lead levels in men with increased versus normal (i.e., 140 or 90 mm Hg, respectively, 

for systolic and diastolic) blood pressure. Significant conelations were found between log 

blood lead and systolic blood pressure in both men and women and diastolic blood pressure in 

women only at age 40, but not at age 45. The relationship at age 40 indicated a doubling of 

blood lead being associated with a *3 mm Hg increase in blood pressure. When blood 

hemoglobin and alcohol intake (the only two of nine potential confounders assessed tiiat were 

17 
significantly related to both blood lead and blood pressure) were entered into multiple 

regression analyses, then all blood-lead blood-pressure associations become non-significant. 

Grandjean et al. (1989) noted that the initial association found between blood lead and blood 

pressure was similar or slightly weaker than those earlier reported for the BRHS (Pocock 

et al., 1985) or the NHANES JJ (Pirkle et al., 1985) data analyses. The impact of alcohol 

and blood hemoglobin on the subsequent multiple regression outcomes indicated clear 

confounding-with blood lead and an inability in this study to separate out relative 

contributions due to those factors versus lead (with blood lead increasing with number of umts 

of alcohol consumed as documented in the study cohort at age 40, for example). The authors 

(Grandjean et al., 1989) further noted that non-response and loss to follow-up, the very low 

range of blood lead values, and the limited statistical power of the study (with fewer than 

1,000 subjects) to detect small effects may help to account for their observed pattern of 

results. 

The issue of lead effects on more serious cardiovascular disease outcomes, possibly 

mediated by lead effects in blood pressure or via other lead-induced pathogenic processes has 

been further addressed since the 1986 Addendum to the EPA Lead Criteria Document was 

prepared. Some of the newer evidence has been derived from studies of occupationaliy 

exposed workers and other new results have emerged from still further analyses of the two 

larger-scale BRHS and NHANES II data sets. 

The causes of mortality of lead workers in the United Kingdom between 1926 and 1985, 

for example, have recently been updated in a case-control study reported by Fanning (1988). 

There were 867 deaths in men with relatively high occupational exposure to lead during these 

years, and 1206 deaths during the same period in men whose exposure had been low or 

absent. During the period between 1946 and 1965 there was a significant excess of deaths 

from cardiovascular disease; but there was no difference between the two groups over the past 

20 years. Also, no statistically significant excess in the number of deaths from malignant 

neoplasms was evident. Fanning concluded that the previous evidence of an increased risk of 

death from cerebrovascular disease (in lead workers) was therefore confirmed, but it would 

seem that with the introduction of stricter standards of lead control this increased risk has now 

disappeared, as has any marginal risk of death from malignant disease. 

18 
P.30 

' Another occupational exposure study, an analysis of causes of death in U.S. battery and 

lead production workers during the years 1947 to 1980, was presented at the 1987 LeadBlood Pressure Symposium by Cooper (1988). Mortality causes were coded according to the 

seventh (1955) revision of the International Classification of Diseases. Data were from deaths 

of 4,519 battery plant workers and 2,300 lead production or smelter workers during this time 

period. Cooper reported that there were significant excess deaths for "other hypertensive 

disease" (444-447) and "chronic nephritis" (592-594); but deaths from other hypertensionrelated diseases did not show comparable excesses and renal cancer deaths were fewer than 

expected. Also, Selevan et al. (1988) reported at the 1987 Symposium that an analysis of 

causes of mortality in 1,281 lead smelter workers employed at an Idaho smelter between 1940 

and 1965 did not suggest an association between occupational lead exposure and mortality 

from hypertension. On the other hand, the data do suggest an association between lead and 

renal disease and, possibly, renal cancer. 

The 1987 Symposium report by Pocock et al. (1988) not only reviewed relationships 

between blood lead concentration and blood pressure as determined in the British Regional 

Heart Survey, but also looked at relationships between lead and more serious cardiovascular 

outcomes. The results of these further analyses of the BRHS data set were reported by 

Pocock et al. (1988) to show that, by 6 years of follow-up, 316 of the men had major 

ischemic heart disease and 66 had a stroke. After controlUng for the confounding effects of 

cigarette smoking and town of residence, statistical analyses did not yield statistically 

significant associations between blood lead levels and such cardiovascular events. However, 

as the blood lead-blood pressure association is so weak, Pocock et al. (1988) noted, it is 

unlikely that any consequent association between lead and cardiovascular disease could be 

demonstrated from prospective epidemiological studies. 

On the other hand, Schwartz (1989) has recently reported results derived from new 

NHANES II data analyses, which provide evidence for significant associations between blood 

lead levels and electrocardiogram (ECG) abnormalities indicative of left ventricular 

hypertrophy (LVH). Such ECG abnormalities represent an early indicator of cardiovascular 

disease that is much more common than frank myocardial infarctions. The logistic regression 

analyses employed by Schwartz (1989) yielded a small (0.028) but statistically significant 

(p <0.01) coefficient for an association between blood lead levels and increased prevalence 

19 
P.31 

of LVH, taking into account age, race, and sex (all of which were significantly related to , 

LVH at p <0.01, < 0.0001, and < 0.0001, respectively). The interaction terms for sex or 

race differences, however, were non-significant (p >0.20) with regard to the cardiovascular 

effects of lead. 

The Schwartz (1989) results are consistent with previous reports of cardiovascular 

effects being associated with high levels of lead exposure, e.g., Kirkby and Gyntelberg's 

(1985) report of a 20% increase (p <0.01) in ischemic changes as coded on the Minnesota 

Codes in lead workers compared to controls matched on age, race, and several other pertinent 

factors. The newfindings by Schwartz, however, point toward small but not inconsequential 

increased risks for serious cardiovascular outcomes being associated with the relatively low 

range of blood lead levels encountered in the general U.S. adult population. It remains to be 

determined to what extent the observed lead effects on left ventricular hypertrophy or other 

cardiovascular functions are due to a lead-induced increase in blood pressure or to some other 

lead-related pathogenic mechanism. 

At the 1987 Symposium, there was extensive discussion concerning controversy in the 

published epidemiology literature about the nature of a relationship between blood lead and 

blood pressure. As summarized by Tyroler (1988), the general population studies discussed 

above were seen as pointing toward a causal relationship between increases in blood lead 

levels and significant increases in blood pressure, extending to values below those cunently 

considered to be clinically significant hypertension. Tyroler further noted that the association 

has not been found in all studies and, when present, has been such that the increase in blood 

pressure with increase in blood lead levels has been of small absolute magnitude and not 

constant across age, sex, and race subgroups. Nevertheless, despite the seemingly small 

elevations in blood pressure when viewed from the clinical perspective of each individual, 

Tyroler further noted that the potential public health importance of a blood-lead bloodpressure relationship is considerable due to the strong association of blood pressure with 

cardiovascular morbidity and mortality, the leading cause(s) of death in our society, and the 

large number of individuals exposed. 

Of importance for substantiating the plausibility of the blood-lead/blood-pressure 

associations observed in human populations representing causal relationships, the 

physiological and pharmacological regulation of blood pressure has been studied extensively 

20 
P.32 

by, experimental investigators. Much of the then available information on the subject was 

discussed in the 1986 Addendum (U.S. Environmental Protection Agency, 1986b). It is now 

understood that various hormonal regulatory systems, vascular smooth muscle, and heart 

contractility all contribute to the development of blood pressure changes. Several of the 

speakers at the 1987 Symposium (Victery, 1988a) also reviewed findings of changes in the 

cardiovascular system in animals exposed to lead. For example, Victery (1988b) reviewed 

the experimental studies that have been conducted for over 40 years on the effects of lead on 

blood pressure. Differences in animals species, age at beginning of exposure, level of lead 

exposure, and the effects of lead on blood pressure were described. It was noted that in 

several of the high-dose experiments, hypertension was observed, but nephrotoxicity of lead 

may have contributed to its development. In one experiment, high lead exposure may have 

reduced an elevated blood pressure. In contrast, lower dose experiments consistently 

demonstrated a hypertensive effect. The data suggest that a biphasic dose-response 

relationship may exist. Future research should be able to characterize in animals doseresponse relationships for blood pressure effects across low-level lead exposure ranges most 

applicable to the general human population. 

With regard to mechanisms potentially underlying lead-blood pressure relationships, 

Vander (1988) reviewed the chronic effects of lead on the renin-angiotensin system (one of 

the primary regulatory hormones for blood pressure). The changes observed in both animals 

and humans are highly variable and are dependent on the lead exposure level, the time of 

exposure, and other stimuli of the renin-angiotensin system. The human data are consistent 

with the tentative hypothesis that lead-exposed persons may have higher plasma renin activity 

than normal during periods of modest exposure but normal or depressed plasma renin activity 

following more chronic severe exposures. 

In another 1988 Symposium paper, the effects of lead on vascular reactivity were 

described by Chai and Webb (1988). There is convincing evidence that lead alters vascular 

reactivity in lead-exposed animals, which demonstrated a 15-20 mm Hg increase in systolic 

blood pressure. Increased pressor responsiveness to catecholamines has been demonstrated in 

chronically lead exposed animals and an enhanced contraction of isolated vascular smooth 

muscle to adrenergic agonists occurs in rats with lead-induced hypertension. Further 

experimental evidence suggests that there are alterations in the mechanisms that regulate 

21 
P.33 

intracellular calcium concentration and that this may contribute to the abnormal vascular 

function in lead-induced hypertension. There were further reports of morphological, 

biochemical, and functional alterations in cardiovascular tissue in the reports by Kopp et al. 

(1988) and Boscolo and Carmignani (1988). The paper by Kopp et al. (1988) in the 

Symposium volume reviews the cardiovascular actions of lead, including evidence that lead 

exposure (at least at high levels) leads to morphological, biochemical, and functional 

derangements of the heart. The experimental literature also confirmed findings of 

cardiovascular complications in experimental animals. Findings include myocarditis, 

electrocardiographic disturbances, heightened catecholamine anhythmogenicity, altered 

myocardial contractile responsiveness to inotropic stimulation, degenerative structural and 

biochemical changes affecting the musculature of the heart and vasculature, hypertension, 

hypercholesterolemia, atherosclerosis, and increased vascular reactivity to alpha-adrengergic 

agonists. The precise nature of the exposure-response relationships that apply are still poorly 

characterized, as well as the exact pathogenic mechanisms for the effects of lead on the 

cardiovascular system; but importantly, the experimental results provide clear evidence for 

lead causally affecting cardiovascular function. 

In the 1987 Symposium discussion session (Victery et al., 1988), one of the invited 

discussants, Dr. Anthony Johns proposed a number of critical experiments to understand how 

lead may be affecting blood pressure and whether blood pressure changes can be reversed by 

removing lead from the diet. Inhibitors of the angiotensin-converting enzyme (now used as 

antihypertensive therapeutic agents) could be tried to determine if this can prevent the blood 

pressure increase during lead exposure. The levels of intracellular calcium should be 

measured in vascular tissue and the pharmacologic agents that block calcium or potassium 

channels in cell membranes should be examined to determine if they might reduce the effects 

of lead. 

CONCLUSIONS . 

With regard to the effects of lead on blood pressure, the new information emerging 

since preparation of the 1986 Addendum, overall, substantiates further the main conclusions 

stated in that Addendum. Sufficient evidence exists from both the four large-scale general 

population studies discussed above (NHANES JJ, BRHS, and the two Welsh studies) and 

22 
P.34 

numerous other smaller-scale studies to conclude that a small but positive association exists 

between blood lead levels and increases in blood pressure. Quantitatively, the relationship 

appears to hold across a wide range of blood-lead values, extending possibly down to as low 

as 7 /ig/dl for middle-aged men and, furthermore, an estimated mean increase of about 1.5-

3.0 mm Hg in systolic blood pressure appears to occur for every doubling of blood lead 

concentration in adult males and something less than 1.0-2.0 mm Hg for adult females. The 

plausibility of these relationships observed in epidemiologic studies of human populations 

being of a causal nature is supported by controlled experimental animal studies demonstrating 

increased blood pressure effects clearly attributable to lead, with an apparent biphasic doseresponse relationship being involved (i.e., blood pressure elevations at low lead dose levels 

and possible blood pressure reductions at very high lead exposure levels). 

The implications of lead-induced blood-pressure increases with regard to potential 

increased risk for other, more serious cardiovascular outcomes still remain to be more clearly 

delineated. As noted by Tyroler and other discussants at the 1988 Symposium mentioned 

above, essentially any increase in blood pressure carries with it likely increased risk (albeit 

however small) for stroke, heart attack, and/or associated mortality. As such, projections of 

potential lead effects on such outcomes, as were modeled by Pirkle et al. (1985) and 

discussed in the 1986 Addendum, are not unreasonable in view of the potentially very large 

pubhc health impacts; however, much caution must be exercised in accepting the validity of 

any specific quantitative estimates derived from such projections in view of the uncertainties 

associated with selection of the specific coefficients used for (1) blood-lead blood-pressure 

relationships and (2) relationships between blood pressure increases and more serious 

cardiovascular outcomes. The difficulty in directly demonstrating associations between lead 

exposure and stroke, heart attacks, etc., lies in the very large study cohorts (many thousands 

of subjects) that would be necessary to have sufficient statistical power to detect the relatively 

small increased risk levels expected for the more serious cardiovascular outcomes. Some 

newly available results (i.e., those of Schwartz, 1989) from at least one large-scale study help 

to illustrate the possibility of detecting indications of such small increased risks in the general 

population. 

23 
fflo NEUROBEHAVIORAL AND GROWTH EFFECTS.M INFANTS AND 

CMILBREN 

A major advance in the epidemiological investigation of the health effects of lead 

occurred with the advent of independent but somewhat coordinated prospective studies of 

child development. Some of the results of four such studies were discussed in the 1986 

Addendum (U.S. Environmental Protection Agency, 1986b). Much of the same information, 

with updating, was also presented in a critical review and interpretation by Davis and 

Svendsgaard (1987). The four prospective studies in question were conducted in Boston, 

Cincinnati, Cleveland, and Port Pirie, Australia. 

Based on an assessment of these studies, the 1986 Addendum concluded that fetal lead 

exposure could have undesirable effects on infant mental development, length of gestation, 

and possibly other aspects of fetal development, with the most consistent evidence pertaining 

to neurobehavioral function. In particular, "All of these studies taken together suggest that 

neurobehavioral deficits, including declines in Bayley Mental Development Index scores and 

other assessments of neurobehavioral function, are associated with prenatal blood lead 

exposure levels on the order of 10 to 15 /*g/dl and possibly even lower, as indexed by 

maternal or cord blood lead concentrations" (U.S. Environmental Protection Agency, 1986b). 

This update summarizes and assesses the evidence from these and other, more recent 

studies, both individually and collectively, and draws conclusions regarding their implications 

for regulatory decision-making. 

As noted in the 1986 Addendum, a series of reports by Bellinger, Needleman, and their 

colleagues (Bellinger et al., 1984a, 1985, 1986a,b) described the results of a longitudinal 

study of early neurobehavioral development through the first two years of life in a cohort of 

Boston children. More recent updates on this work have since been published by Bellinger 

et al. (1987a, 1989a), covering the same period of development. The latter reports confirm 

that performance on the Bayley Mental Development Index (MDI) at 6, 12, 18, and 

24 months postnatally is inversely related to cord blood lead levels at birth and that the 

amount of performance deficit in infants with cord blood lead levels of 10-25 jxg/dl is 

0.25-0.5 standard deviations, or approximately 4-8 points on the MDI (Table 2). Moreover, 

24 
P.36 

TABLE 2, INFANTS' MENTAL DEVELOPMENT INDEX SCORES 

ACCORDING TO CORD BLOOD LEAD GROUP 

Cord Blood 

Lead Group 

Unadjusted score 

Low 

Medium 

High 

6 Months 

109.2 ± 12.9 

108.6 ± 12.0 

106.1 ± 11.1 

Controlled for potential confounders* 

Low 110.2 ± 1.3 

Medium 108.0 ± 1.3 

High 105.9 ± 1.4 

p value** 

No. of infants 

0.095 

201 

Mental Development Index Score 

12 Months 18 Months 

Mean ± Standard Deviation 

113.1 ± 12.5 

115.4 ± 12.9 

108.7 ± 12.8 

Mean ± Standard Error 

114.7 ± 1.6 

114.4 ± 1.5 

108.9 ± 1.6 

0.020 

199 

113.4 ± 15.5 

116.6 ± 16.7 

109.5 ± 17.5 

116.2 ± 1.9 

114.8 ± 1.9 

109.5 ± 2.0 

0.049 

187 

24 Months 

115.9 ± 17.2 

119.9 ± 14.4 

110.6 ± 16.5 

118.9 ± 1.8 

117.8 ± 1.7 

111.1 ± 1.8 

0.006 

182 

*Least-squares mean ± standard error, derived from regression equations that included 12 potential 

confounders and cord blood lead group coded as two indicator variables. 

""•Indicates p value associated with the F ratio that evaluates whether the mean Mental Development Index for 

any cord blood lead group differed significantly from the common mean after potential confounders were 

controlled for. 

Source: Bellinger et al. (1987a). 

Bellinger et al. (1989a) found that this association was evident in several different regression 

models and was not an artifact of the approach they used in analyzing the data. In addition, 

more detailed analyses showed that the average MDI deficit in high cord lead infants (blood 

lead levels of 10-25 /ig/dl) was not due to a disproportionate influence of results from infants 

at higher (e.g., > 15 /ig/dl) blood lead levels. That is, the MDI effect was evident across the 

entire range of blood lead levels starting at 10 /ig/dl, which reinforces the previous selection 

of 10-15 /tg/dl as a blood lead level of concern for early developmental deficits (U.S. 

Environmental Protection Agency, 1986b). 

25 
P.37 

Other recent analyses by Bellinger et al. (1988) have focused on the interaction of lead-* 

related Bayley MDI deficits with socioeconomic status (SES). Even in the relatively 

advantaged cohort of the Boston study, infants from less than the highest SES grouping 

tended to have lower covanate-adjusted scores on the MDI. Moreover, the second-year MDI 

performance of these "lower" SES children was adversely affected at lower blood lead levels 

than was the performance of the higher SES children. Specifically, within the lower SES 

grouping, infants with mid-range cord blood lead levels of 6-7 /ig/dl scored significantly 

worse on the Bayley MDI than infants with low cord blood lead levels of <3 /ig/dl. 

Bellinger et al. (1987b, 1989b) have also recently reported preliminary results of later 

testing of 170 children of the Boston cohort at about 5 years (57 months) of age on the 

McCarthy Scales of Children's Abilities. They found that the association between cord blood 

lead and cognitive performance, as measured by the General Cognitive Index (GCI) of the 

McCarthy Scales, was no longer statistically significant at 57 months. However, the 

relationship between blood lead level at 24 months postnatally and GCI scores was statistically significant, even after adjusting for covariates and confounders (Table 3). The mean 

24-month blood lead level was 6.8 /ig/dl (SD = 6.3). Other postnatal blood lead 

measurements at 18 and 57 months were consistent with this association but failed to achieve 

statistical significance. The size of the deficit amounted to approximately 3 GCI points for 

every natural log unit increment in blood lead level. 

Bellinger et al. (1989b) examined the change in cognitive performance of children from 

24 to 57 months of age in relation to pre- and postnatal lead exposure levels and various 

sociodemographic variables. Improvement in relative performance was associated with lower 

blood lead levels at 57 months, higher SES, higher HOME scores, higher maternal IQ, and 

female gender. Conversely, the risk of an early deficit persisting to 5 years of age was 

increased in children with higher prenatal lead exposure (10-25 /ig/dl) and either high 

postnatal exposure or less favorable sociodemographic factors. For example, "if two children 

with high cord blood lead achieved the same MDI score at 24 months, but one had a low 

blood lead level [<3 /ig/dl] at 57 months while the other had a high [> 10 /ig/dl] level, the 

child with lower exposure would be expected to have a GCI score that is 0.61 standard 

nonnal deviate units higher..., [which] conesponds to a difference of 9.8 points" (Bellinger 

et al., 1989b). Similar comparisons of males and females with high cord blood lead levels 

26 
TABLE 3. CHANGE IN McCARTHY GENERAL COGNITIVE INDEX (GCI) 

AND SUBSCALE SCORES ASSOCIATED WITH EACH 

NATURAL LOG UNIT INCREASE IN BLOOD LEAD LEVEL 

Age at 

Measurement 

of Blood 

Lead Level 

(months) 

6 

12 

18 

24 

57 

GCI 

0.28 ± 1.29* 

.83** 

-1.43 ± 1.25 

.25 

-1.62 ± 1.39 

.25 

-2.95 ± 1.42 

.040 

-2.28 ±1.88 

.23 

Verbal 

0.75 ± 0.93 

.42 

-0.98 ± 0.89 

.27 

-0.19 ± 1.01 

.85 

-0.41 ± 1.04 

.69 

-1.06 ± 1.38 

.44 

PerceptualPerformance 

-0.72 ± 0.77 

.35 

-0.64 ± 0.79 

.42 

-0.72 ± 0.87 

.41 

-2.58 ± 0.88 

.004 

-2.33 ±1.13 

.042 

Quantitative 

0.00 ± 0.76 

.99 

-0.09 ± 0.74 

.90 

-1.20 ± 0.82 

.15 

-1.45 ± 0.85 

.088 

0.13 ± 1.13 

.91 

Memory 

0.82 ± 0.83 

.33 

-0.09 ± 0.81 

.91 

-0.52 ± 0.91 

.57 

-0.66 ± 0.94 

.49 

0.49 ± 1.25 

.70 

Motor 

0.02 ± 0.81 

.98 

-0.82 + 0.80 

.31 

0.30 ± 0.89 

.74 

-0.90 ± 0.92 

.33 

-1.89 ± 1.15 

.10 

*Parameter estimate ± standard enor. 

p-value associated with the null hypothesis that the parameter estimate is zero. 

Source: Bellinger et al. (1987b). 
P.39 

indicated that boys scored 7.7 points lower than girls on the GCI, and children in the lower« 

SES grouping scored 13.3 points lower than those in the higher SES grouping. 

Bellinger et al. (1987b) also evaluated the degree of disparity in McCarthy subscale 

performance, since learning disabilities are often highly specific to certain cognitive functions. 

A greater number of subscale discrepancies was associated with higher (> 10 /ig/dl) 

concurrent (57 month) blood lead levels. 

Interim results of a longitudinal study of inner-city children born in Cincinnati, Ohio 

were reported by Dietrich et al. (1986, 1987a) and summarized in the 1986 Addendum. 

Structural equation modeling, a statistical method of examining relationships among factors 

that may be both independent and dependent variables or mediators of effects, indicated that 

prenatal lead exposure had an indirect effect on 6-month MDI and PDI (Psychomotor 

Development Index) scores through its effects on gestational age and/or birth weight. Higher 

prenatal (maternal) blood lead levels were associated with reduced gestational age and reduced 

birth weight, which in turn were significantly associated with reduced MDI and PDI 

performance. 

Although the Cincinnati subjects were not grouped by blood lead levels for these 

analyses, separate analyses of birth weight effects by Bornschein et al. (1989) investigated the 

dose-response relationship by grouping into five 6-/ig/dl intervals. An increase in the 

percentage of low birthweight newborns between the 7-12 /ig/dl grouping and the £ 13 /ig/dl 

grouping implied the possibility of a threshold for low birthweight effects (categorically 

defined as less than 2750 g) in the vicinity of 12-13 /ig/dl, although a precise determination is 

not possible and could extend as low as 7 /ig/dl or perhaps as high as 18 /tg/dl. To the extent 

that low birthweight mediated the effect of lead on Bayley scores in the Cincinnati study, the 

infened information is consistent with the conclusion (U.S. Environmental Protection 

Agency, 1986b) that a blood lead level of 10-15 /ig/dl, and possibly lower, constitutes a level 

of concern for impaired performance on the Bayley MDI. 

Dietrich et al. (1987b, 1989a,b) later reported more complete results for various 

neurobehavioral developmental outcomes in approximately 300 infants from the Cincinnati 

study (Table 4). The children were tested on the Bayley Scales at 3, 6, 12, and 24 months of 

28 
P.40 

TABLE 4. RESULTS OF MULTIPLE REGRESSION ANALYSES EXAMINING THE 

RELATIONSHIP BETWEEN BLOOD LEAD LEVEL AND PERFORMANCE 

ON THE BAYLEY MENTAL INDEX AT 3-24 MONTHS OF AGE* 

Blood Lead 

Measure (/tg/dl) 

Prenatal 

Umbilical cord 

10 day 

3 month 

N Beta 

Standard 

Error 

228 

80 

261 

n.r. 

3-month MDI 

-.34 

-.60 

.06 

-.23 

0.17 

0.26 

0.22 

0.18 

*Betas adjusted for different sets of covariates for each age of MDI testing. 

**Two-tailed values. 

Sources: Dietrich et al. (1987b, 1988, 1989b). 

t 

Value 

-1.96 

-2.30 

0.26 

-1.30 

P 

Value** 

.05 

.02 

.79 

.20 

Prenatal 

Prenatal by 

Child Sex 

10 day 

10 day by SES 

Prenatal 

Umbilical cord 

10 day 

1 Prenatal 

1 10 day 

1 3 month 

1 Maximum 1st Year 

1 Maximum 2nd Year 

1 24 months 

249 

283 

258 

98 

257 

237 

270 

270 

270 

270 

270 

6-month MDI 

-0.89 

1.53 

-3.15 

0.16 

12-month MDI 

0.09 

-0.17 

-0.62 

24-month MDI 

0.51 

-0.02 

0.24 

0.24 

0.10 

0.13 

0.34 

0.51 

1.30 

0.08 

0.26 

0.36 

0.31 

0.22 

0.25 

0.21 

0.10 

0.07 

0.09 

-2.60 

2.98 

-2.43 

2.13 

n.r. 

n.r. 

n.r. 

2.31 

-0.07 

1.12 

1.39 

1.39 

1.45 

0.009 

0.003 

0.016 

0.034 

N.S. 

N.S. 

<0.04 

0.022 

0.948 

0.262 

0.166 

0.166 

0.149 

29 
P.41 

age. After adjusting for covariates, deficits in 3- and 6-month Bayley MDI scores were 

significantly associated with prenatal (maternal) and cord blood lead levels, confirming the 

earlier, preliminary analyses of Dietrich et al. (1987a). The magnitude of the deficit in the 

3-month MDI amounted to 6 points for every 10-/tg/dl increment in cord blood lead; the 

conesponding 6-month MDI deficit was nearly 7 points. Also, a conservative reanalysis 

(Dietrich et al., 1989b) of the 6-montii data using all potential covariates and confounders 

confirmed earlier findings, although interactive effects with gender and SES were evident (see 

below). However, neither prenatal nor cord blood lead measures were significantly related to 

12-month MDI scores, although the association between neonatal (10-day) blood lead levels 

and 12-month (as well as 6-month) MDI scores remained statistically significant (Dietrich 

et al., 1989a). It should be noted that analyses ofthe 12-month data are not yet final. 

By 24 months, no statistically significant negative relationships were detectable between 

lead exposure variables (either pre- or postnatal) and MDI scores (Dietrich et al., 1989b). 

Only one parameter estimate (for prenatal blood lead) achieved statistical significance (p = 

0.022), and it was positive rather than negative. 

In addition, the Bayley Infant Behavior Record (IBR) was administered at 12 and 

24 months to assess the infants' social and emotional development (Dietrich et al., 1989a,b). 

Factor analysis of the 30 IBR items at 24 months yielded principal factors of Sustained 

Attention, Activity Level, and Positive Mood, which were used in regression analyses to 

assess the effect of lead exposure. Lower scores on Sustained Attention and Positive Mood 

factors were both significantly associated with increased neonatal (10-day) blood lead levels, 

consistent with the 12-month MDI relationship to neonatal blood lead (Dietrich et al., 1989a). 

However, none of the 24-month IBR results achieved statistical significance. Bayley PDI 

results were incomplete. 

Dietrich et al. (1987b, 1989b) also found that gender and SES interacted with the lead 

exposure-MDI relationship. Male infants and children from the lower half of the SES 

distribution for this cohort were more sensitive to the effects of early lead exposure on 

neurobehavioral development. For example, male infants showed an 8.67-point deficit on the 

6-month MDI for every 10-/ig/dl increment in prenatal blood lead, and infants below the 

sample median SES score had a 7.57-point deficit for every 10-/ig/dl increment in neonatal 

blood lead (Dietrich et al., 1989b). 

30 
P.42 

Dietrich et al. (1989b) interpreted their failure to detect a persistent effect of fetal lead 

exposure on the 24-month Bayley Scales as probably due to a neurobehavioral catch-up 

response, similar to that observed in infant twins (Wilson, 1986) or other infants 

compromised during prenatal development (Tanner, 1981). Exploratory analyses indicated 

that the greatest percentage increase in MDI raw scores (number of items passed) from age 

3 months to 24 months was inversely related to prenatal lead exposure, birth weight, 

gestation, and head circumference (Dietrich et al., 1989b). Thus, those infants with the 

highest prenatal blood lead levels, lowest birth weight, shortest gestation, or smallest head 

circumference showed the greatest degree of catch-up in postnatal neurobehavioral 

development. 

As previously noted, the Cincinnati study has also produced evidence of direct effects of 

prenatal lead exposure on infant physical development at birth. Bornschein et al. (1989) 

reported that initial results from 202 infants showed an inverse relationship between maternal 

blood lead levels and both birth weight and length. Maternal blood lead samples were 

obtained at the first prenatal visit (between 6 and 28 weeks of gestation; mean = 16 wks). 

The mean maternal blood lead level was 7.6 /ig/dl (range: 1-26 /tg/dl). Exclusion criteria 

included birth weight less than 1500 g, less than 35 weeks gestational age at birth, twin birth, 

and serious medical conditions. Regression analyses considered 21 potential confounders and 

covariates. The final regression model for birth weight showed that log maternal blood lead 

was significantly related to birth weight, but interacted with maternal age to produce a 

significant (p < 0.007) reduction in covanate-adjusted birth weight. Thus, the effect of each 

natural log increment in blood lead varied from a birthweight reduction of 58.1 g for 18-

year-old mothers to a reduction of 600.1 g for 30-year-old mothers. Maternal blood lead and 

race interacted to produce a significant reduction in birth length in white infants (p < 0.025). 

Thus, the birth length of white infants decreased ~ 2.5 cm for each natural log increment in 

maternal blood lead. Maternal blood lead showed no significant relationship to covariateadjusted head circumference or gestation length. Obstetrical complications and Apgar scores 

also showed no relationship with lower birth weight or lead exposure. 

Separate analyses of 861 women (including the 202 subjects just described) from the 

Cincinnati cohort, but including pregnancies of not less than 20 weeks duration, also showed 

a highly significant (p < 0.0006) negative relationship between maternal blood lead level and 

31 
P.43 

covanate-adjusted birth weight (Bornschein et al., 1989). Overall, this cohort had a decrease* 

of approximately 114 g in birth weight for each natural log increment in maternal blood lead. 

Analyses of postnatal growth rates in a cohort of 260 children from the Cincinnati study 

suggested an interactive effect of prenatal and postnatal lead exposure (Shukla et al., 1987, 

1989). Splitting prenatal blood lead levels and the average increase in postnatal blood lead 

levels at the median (7.7 /tg/dl for prenatal, 3.4 /tg/dl for the postnatal increase) provided a 

matrix of low/low, low/high, high/low, and high/high exposure conditions. Analysis of data 

for 129 subjects in the high prenatal exposure classification indicated that covariate-adjusted 

growth rates for stature over 3-15 months of age were significantly (p = 0.006) and 

negatively related to the postnatal increase in blood lead level. No effect was seen for the 

low prenatal exposure group. Thus, for infants whose prenatal blood lead levels were greater 

than 7.7 /tg/dl, there would be, on average, a 2-cm difference in height at 15 months of age 

between those infants who experienced no increase in postnatal blood lead and those who 

experienced an increase of ~ 10 /tg/dl. 

Other neurobehavioral test results from the Cincinnati study have also been reported by 

Bhattacharya et al. (1988, 1989). Postural sway was assessed by an automated apparatus in 

33 children at 6 years of age (Bhattacharya et al., 1988). Initial results indicated that 

maximum blood lead level during the second year of life was significantly related to degree of 

postural sway or imbalance at 6 years. Blood lead levels in this sample peaked during the 

second year, averaging 25.6 /tg/dl (range: 9.3-49.4 /tg/dl), measured quarterly. The later 

report by Bhattacharya et al. (1989) covers 63 children and confirms the earlier results 

showing a significant relationship between sway and second year maximum blood lead (r = 

0.34, p = 0.02). Although more test results are needed to fully explore these effects, these 

preliminary findings provide additional indications of neurobehavioral dysfunction during 

early childhood exposure to lead. 

Cleveland 

As noted in the 1986 Addendum (U.S. Environmental Protection Agency, 1986b), the 

prospective study conducted by Emhart and her colleagues has provided some direct as well 

as indirect evidence of an effect of prenatal lead exposure on neurobehavioral development. 

Emhart et al. (1985a, 1986) reported significant associations between cord blood lead levels 

32 
P.44 

and measures of Abnormal Reflexes (on the Brazelton Neonatal Behavioral Assessment Scale: 

NBAS) and Neurological Soft Signs (on the Graham-Rosenblith Behavioral Examination for 

Newborns: G-R). Also, the G-R Muscle Tonus measure was significantly related to maternal 

blood lead levels at delivery. However, restricting the analyses to only 132 pairs of motherinfant data, Emhart et al. found only the G-R Neurological Soft Signs to be significantly 

related to cord blood lead. A brief report on later outcomes in this same cohort mentioned a 

significant association between performance on the G-R Neurological Soft Signs scale and 12-

month MDI scores (Wolf et al., 1985). Thus, it is possible to infer a relationship between 

cord blood lead levels and 12-month MDI performance in the Cincinnati study, although 

Ernhart et al. (1985a, 1986) did not conclude that such an association exists. Since the mean 

cord blood lead was 5.84 /tg/dl and the maximum was only 14.7 /tg/dl, any effect of prenatal 

lead exposure necessarily occuned at blood lead levels below 15 /tg/dl. 

Later reports by Ernhart and her colleagues (Ernhart et al., 1987, 1988; Ernhart and 

Monow-Tlucak, 1989) presented more complete results from their continuing longitudinal 

study (Table 5). The Bayley MDI was administered at 6, 12, and 24 months; the Bayley PDI 

and the Kent Infant Development Scale (KID) at 6 months; and the Stanford-Binet IQ test at 

36 months. Postnatal blood lead levels were measured at 6, 24, and 36 months. A total of 

285 children from the original cohort of 389 were sampled for blood lead levels; N's for 

individual analyses ranged from 109 to 165. After control for covariates, maternal blood lead 

accounted for a significant amount of the variance in 6-month MDI, PDI, and KID scores. 

Although these three relationships were all negative (higher PbB associated with lower 

developmental scores), concunent blood lead was positively related to the 6-month KID and 

accounted for nearly as much variance as did maternal blood lead. Maternal blood lead levels 

averaged 6.50 /tg/dl, with a maximum of 11.8 /tg/dl; 6-month blood lead levels averaged 

10.05 /tg/dl, with a maximum of 24.00 /tg/dl. 

More recent results from testing these children at age 4 years, 10 months on the 

Wechsler Preschool and Primary Scale of Intelligence (WPPSI) were reported by Emhart and 

Monow-Tlucak (1987). Analyses were based on N's ranging from 117 to 211. Although 

bivariate conelations were statistically significant for all PbB-WPPSI relationships except 

6-month PbB, in no case did PbB account for a significant amount of the variance (by 

two-tailed t test) after control for 13 covariates (Table 6). 

33 
TABLE SA. THE RELATIONSHIP OF FETAL (MATERNAL AND CORD) BLOOD LEAD 

(PbB) LEVELS WITH LATER DEVELOPMENTAL OUTCOMES 

Blood 

Lead 

Measure 

Developmental 

Measure N 

Covariate 

Variance 

Incremental Regression Model 

PbB With Race by PbB 

Covariate Control Interactions 

Variance Variance 

Maternal 

Cord 

*p <0.05, 

6 Month MDI 

6 Month PDI 

6 Month KID 

1 Year MDI 

2 Year MDI 

3 Year S-B IQ 

6 Month MDI 

6 Month PDI 

6 Month KID 

1 Year MDI 

2 Year MDI 

3 Year S-B IQ 

p <0.01, p 

127 

127 

119 

145 

142 

138 

113 

113 

109 

127 

125 

120 

= 0.002, 

-.24 

-.21* 

-.28** 

-.14 

-.14 

-.10 

-.14 

-.11 

-.10 

-.24** 

-.15 

-.21* 

two-tailed tests. 

0.2460 

0.2485 

0.1464 

0.1887 

0.3664 

0.3894 

0.2394 

0.1891 

0.1418 

0.2327 

0.3746 

0.4393 

0.0302 

0.0295 

0.0779 

0.0002 

0.0020 

0.0059 

0.0046 

0.0006 

0.0107 

0.0146 

0.0003 

0.0125 

-2.12* 

-2.12* 

-3.22 

-0.16 

+0.48 

+ 1.09 

-0.76 

-0.26 

-1.07 

-1.47 

-0.21 

-1.54 

0.0161 

0.0093 

0.0182 

0.0026 

0.0000 

0.0020 

0.0056 

0.0249 

0.0051 

0.0143 

0.0002 

0.0086 

-1.57 

-1.19 

-1.56 

-0.64 

-0.01 

-0.63 

-0.74 

+ 1.73 

+0.74 

-1.46 

-0.13 

-1.28 

TABLE SB. THE RELATIONSHIP OF PRIOR AND CURRENT POSTNATAL BLOOD LEAD 

Blood 

Lead 

Measure 

6 month 

2 year 

3 year 

(PbB) LEVELS WITH LATER DEVELOPMENTAL OUTCOMES 

Developmental 

Measure 

6 month MDI 

6 month PDI 

1 year MDI 

2 year MDI 

3 year S-B IQ 

2 year MDI 

3 year S-B IQ 

3 year S-B IQ 

N 

146 

146 

131 

126 

126 

165 

153 

167 

Simple 

Correlation 

r 

-.09 

.06 

.08 

-.03 

-.04 

-.25 

-.31 

-.27 

P 

.29 

.49 

.40 

.74 

.65 

Covariates 

.18 

.19 

.14 

.30 

.29 

.000 .32 

.000 .31 

.000 .36 

P 

.000 

.000 

.02 

.000 

.000 

.000 

.000 

.000 

Incremental Regression Model 

PbB with 

Covariate 

Control 

.01 

.01 

.01 

.00 

.00 

.00 

.01 

.00 

P 

.31 

.32 

.25 

.95 

.49 

.95 

.29 

.98 

Interaction p 

.01 

.00 

.00 

.00 

.00 

.01 

.00 

.01 

.34 

.48 

.51 

.63 

.87 

.07 

.32 

.08 

Two-tailed statistical test. 

Abbreviations: MDI = Bayley Mental Development Index; PDI = Bayley Psychomotor Development Index; KID = Kent Infant 

Development Scale; S-B IQ = Stanford-Binet Intelligence Quotient 

Source: Ernhart et al. (1987, 1988). 

34 
P.46 

TABLE 6. 

Blood 

Lead 

Measure 

Prenatal 

Maternal 

Cord 

Postnatal 

6 Months 

2 Years 

3 Years 

Mean 

THE RELATIONSHIP 01 F PRENATAL AND POSTNATAL L1 

BLOOD LEAD WITH IQ AT AGE FOUR YEARS, TEN MONTHS 

N 

134 

117 

121 

149 

154 

211 

Full Scale 

1> 

-.23** 

-.21* 

-.06 

-.38** 

-.32** 

-.26** 

t 

-0.27 

4).54 

-0.08 

-0.69 

+0.30 

+0.48 

IQ* 

Verbal Scale 

T 

-.25** 

-.22* 

-.07 

-.36** 

-.39** 

-.30** 

t 

-0.49 

-0.78 

-0.26 

-1.41 

-1.58 

-1.00 

OG 

Performance Scale 

r 

-.23** 

-.20* 

-.08 

-.34** 

-.30** 

-.26** 

t 

-0.48 

-0.78 

+0.10 

+0.22 

+0.96 

+0.56 

aWPPSI scores except for two S-B IQ scores; latter two cases were necessarily excluded from the tests of the 

Verbal and Performance Scales. 

r describes the unadjusted relationship of the PbB and IQ measures; t is the test of the increment in variance 

associated with the addition of PbB to the covariate variance in the IQ model. 

*p <.05, **p <.01, two-tailed tests. 

Source: Ernhart and Morrow-Tlucak (1987). 

Language development in the Cleveland cohort was also assessed at age 1,2, and 

3 years (Monow-Tlucak and Emhart, 1987). The Sequenced Inventory of Communication 

Development (SICD) was administered to assess expressive and receptive language 

development. In addition, productive speech (quantity, length of utterance, vocabulary, 

communicative intent, and intelligibility) was assessed at about 2 years. The number of 

subjects for each of the analyses was not reported, but the number of blood lead 

measurements over the period from delivery to 3 years ranged from 146 to 169. Alpha was 

set at 0.01. Several bivariate conelations achieved statistical significance, but no relationship 

achieved significance in multivariate F tests, although the regression analysis for cord PbB 

and mean length of utterance was significant at p = 0.03. By backward elimination, lead 

(cord PbB) remained in the model (at p ^0.05) for two measures of language development: 

mean length of utterance and one aspect of communicative intent ("expanded repetitions"). 

35 
P.47 

However, concurrent (2-year) blood lead was positively related to one aspect of intelligibility 

(a decrease in "unintelligible utterances"). 

Preliminary analyses of physical growth in the preschool period (up to 4 years, 

10 months) revealed no significant negative relationship between either cord blood lead or 

integrated postnatal blood lead and height or weight in an unspecified number of children 

from the Cleveland study (Marler and Emhart, 1987). Earlier analyses (Ernhart et al., 1986) 

of birth weight, length, and head circumference had shown no significant effect of either 

maternal blood lead (N = 185) or cord blood lead (N = 162). 

Vimpani et al. (1985) reported preliminary neurobehavioral results for the prospective 

study conducted in Port Pirie, South Australia. The Bayley MDI and PDI were administered 

to 592 children at age 24 months. Initial results indicated that a decline in 24-month MDI 

performance was significantly associated with postnatal blood lead levels both at 6 months and 

integrated over the 24 months after birth. Although several covariates were taken into 

account in these partial linear regression analyses, maternal IQ had been measured for only 

part of the cohort and HOME scores had not yet been included in the analyses. Nevertheless, 

as noted in the 1986 Addendum, the available information suggested that results from the Port 

Pirie study were consistent with the other prospective studies in pointing to deficits on the 

Bayley MDI as a function of perinatal lead exposure. 

Although postnatal rather than prenatal lead exposure appeared to play a greater role in 

the MDI deficits of the Port Pirie cohort, it has been noted that blood lead levels increased 

considerably after birth, particularly from 6 to 15 months of age, and that earlier testing on 

the Bayley Scales (e.g., at 6 months) might have revealed a more significant relationship with 

a measure of prenatal lead exposure (U.S. Environmental Protection Agency, 1986b; Davis 

and Svendsgaard, 1987). Indeed, the Port Pirie investigators have since stated that "the likely 

greater impact of the much higher levels of PbB encountered postnatally" may have accounted 

in part for the difference in their results versus the other prospective studies (Wigg et al., 

1988). 

Later reports from the Port Pirie study have noted that the effect of postnatal blood lead 

levels on the 2-year MDI was attenuated as more complete controls for maternal IQ and 

36 
. HOME scores were incorporated into the analysis, although all regression coefficients for 

postnatal blood lead measures remained negative (Vimpani et al., 1989; Wigg et al.9 1988; 

Baghurst et al., 1987). Elevations in 6-month blood lead continued to have the greatest 

measurable impact on 2-year MDI scores, although the regression coefficient for 6-month 

blood lead was only -0.16, with p = 0.07, after controlling for at least 15 covariates. The 

results indicated that "with other factors remaining constant, a child's MDI at 24 months will 

be 1.6 points (equivalent to 1.5%) lower for every 10 /ig/dl rise in PbB at 6 months of age. 

Moreover, in view of the fit of a linear model in multiple regression analysis, these findings 

provide no evidence of a threshold effect" (Wigg et al., 1988). 

Continuing study of the Port Pirie cohort has yielded results for neurobehavioral 

development at age four years (McMichael et al., 1988). The McCarthy Scales of Children's 

Abilities were administered to 537 children within 6 months of their fourth birthday; N's for 

individual analyses ranged from 463 to 534. Multiple regression analyses incorporating 

18 covariates indicated that scores on the McCarthy GCI were significantly related to logtransformed postnatal blood lead levels at 6, 24, and 36 months as well as an integrated 

average for the four-year postnatal period (Table 7). Similar effects were also evident for the 

McCarthy Perceptual-Performance and Memory Scales. 

The largest single coefficient (-15.0; p = 0.04) was for the GCI and integrated postnatal 

average blood lead relationship, which indicates that GCI scores decreased by 15 points for 

every 10-fold increment in blood lead. Alternatively stated, GCI scores declined 

approximately 7.2 points as blood lead levels increased from 10 to 30 /tg/dl. Further analyses 

indicated that the relationship between lead exposure and GCI was as strong or even stronger 

at blood lead levels below 25 /tg/dl than it was overall (geometric mean blood lead level 

peaked at 2 years: 21.2 /tg/dl; integrated postnatal average: 19 /ig/dl). Other analyses 

revealed no indication that GCI performance at four years was especially influenced by more 

recent blood lead levels; rather, the effect of lead on the GCI appeared to be cumulative 

across the entire postnatal period. 

Pregnancy outcomes were also evaluated in the Port Pirie cohort (McMichael et al., 

1986). A significantly elevated risk of preterm (<37 weeks) delivery was associated with 

maternal blood lead levels above 14 /tg/dl. Neither birth weight nor spontaneous abortions 

(<20 weeks) showed a significant association with blood lead levels. However, blood lead 

37 
TABLE 7. ESTIMATED COEFFICIENTS OF LOG BLOOD LEA© CONCENTRATION FROM SIMPLE 

AND MULTIPLE REGRESSION ANALYSES OF McCARTHY SCORES AT THE AGE OF FOUR YEARSC 

Blood Sample 

General Cognitive Index 

Simple Partial 

Perceptual-Performance Scale 

Simple Partial 

Memory Scale 

Simple Partial 

oo 

Prenatal average 

Delivery 

Cord 

6 Month 

15 Month 

24 Month 

36 Month 

48 Month 

Integrated postnatal 

average 

-15.1 ± 4.7 

(0.001) 

-10.4 ± 3.9 

(0.008) 

-6.1 ± 3.4 

(0.08) 

-15.1 ± 3.8 

(<0.001) 

-14.3 ± 3.8 

(< 0.001) 

-25.8 ± 4.3 

(<0.001) 

-25.1 ± 4.3 

(<0.001) 

-20.1 ± 4.1 

(<0.001) 

-28.9 ± 5.5 

(<0.001) 

-1.8 ± 5.7 

(0.75) 

-0.4 ± 4.5 

(0.93) 

3.3 ± 4.1 

(0.42) 

-8.5 ± 4.4 

(0.05) 

-3.5 ± 4.8 

(0.46) 

-11.1 ± 5.4 

(0.04) 

-13.1 ± 5.6 

(0.02) 

-5.5 ± 5.1 

(0.29) 

-15.0 ± 7.3 

(0.04) 

-9.5 ± 2.7 

(<0.001) 

-5.1 ± 2.3 

(0.03) 

-3.0 ± 2.0 

(0.13) 

-7.8 ± 2.2 

(< 0.001) 

-8.2 ± 2.2 

(< 0.001) 

-13.5 ± 2.5 

(<0.001) 

-13.6 ± 2.5 

(<0.001) 

-11.2 ± 2.4 

(<0.001) 

-15.1 ± 3.2 

(<0.001) 

-4.1 ± 3.3 

(0.22) 

0.2 ± 2.7 

(0.95) 

1.6 ± 2.4 

(0.50) 

-4.8 ± 2.5 

(0.06) 

-3.2 ± 2.8 

(0.26) 

-6.2 ± 3.2 

(0.05) 

-7.7 ± 3.3 

(0.02) 

-4.0 ± 3.0 

(0.18) 

-8.3 ± 4.1 

(0.05) 

-4.8 ± 2.7 

(0.07) 

-4.9 ± 2.2 

(0.03) 

-3.0 ± 2.0 

(0.12) 

-4.9 ± 2.2 

(0.03) 

-5.3 ± 2.2 

(0.01) 

13.4 ± 2.4 

(< 0.001) 

12.7 ± 2.4 

(<0.001) 

10.3 ± 2.3 

(<0.001) 

13.3 ± 3.1 

(<0.001) 

2.0 ± 3.3 

(0.54) 

-0.4 ± 2.6 

(0.89) 

2.6 ± 2.5 

(0.27) 

-2.4 ± 2.5 

(0.34) 

-0.5 ± 2.7 

(0.87) 

-7.3 ± 3.1 

(0.02) 

-7.2 ± 3.2 

(0.27) 

-4.1 ± 3.0 

(0.17) 

-7.9 ± 4.2 

(0.06) 

The partial regression coefficients are from analyses incorporating 18 covariates. The p values are for assessments of the two-sided hypothesis that the estimates 

are not zero. 

Source: McMichael et al. (1988). 
P.50 

levels of mothers who had stillbirths were significantly lower than those who had live births 

(7.9 vs. 10.4 /tg/dl). As discussed in the 1986 Addendum and by Davis and Svendsgaard 

(1987), this seemingly paradoxical relationship might reflect increased transfer of lead from 

the mother to the fetus, resulting in greater fetal toxicity and stillbirth. Vimpani et al. (1989) 

reported preliminary results of an analysis of tissue lead concentrations in cord and placental 

membrane and body that are consistent with this hypothesis. Placental membrane and body 

concentrations of lead were higher in cases of late fetal deaths and preterm births than for 

normal births. However, the number of cases analyzed was limited (e.g., 6 stillbirths, 

23 preterm births) and statistical analyses have not been completed. 

sydney 

Reports have recently started to emerge from a longitudinal study of children in Sydney, 

Australia (Cooney et al., 1989a, 1989b; McBride et al., 1989). Of an original cohort of 318, 

298 mothers and infants were sampled for blood lead levels at birth. After 3 years, 

215 children remained in the study. A second cohort of 123 children was also recruited 

because of "concern over the possible contamination of some of the early capillary blood 

samples" (Cooney et al., 1989b). However, the results reported by Cooney et al. (1989a) 

apparently pertain to the original cohort only. Geometric mean blood lead levels for mothers 

and infants at birth were 9.1 and 8.1 /tg/dl, respectively (overall range: 0-29 /tg/dl). 

Although blood lead measures were also taken at 6, 12, 18, 24, 30, and 36 months (McBride 

et al., 1989), the postnatal values were not considered in the analysis of neurobehavioral 

outcomes during the first three years. The Bayley MDI and PDI were administered at 6, 12, 

and 24 months, and the McCarthy GCI and Motor Scales were administered at 36 months. 

Other outcomes were also evaluated but the results have not yet been reported. 

Unadjusted bivariate conelations between blood lead levels (either maternal or cord) and 

cognitive and psychomotor outcomes (either Bayley or McCarthy Scales) were generally 

small, positive, and nonsignificant (Table 8). The only statistically significant simple 

conelations were for the relationship between cord blood lead and the 12-month Bayley MDI 

(r = 0.153, p <0.05, two-tailed) and PDI (r = 0.167, p <0.05, two-tailed). However, the 

direction of the relationship was positive, i.e., as blood lead increased, Bayley scores 

increased. After adjustment for covariates, the contribution of blood lead to the variance in 

39 
P.51 

TABLE 8. REGRESSION OF DEVELOPMENTAL INDICES ON MA' 

AND CORD BLOOD LEAD LEVELS 

Age 

6 Months 

Cognitive 

Psychomotor 

12 Months 

Cognitive 

Psychomotor 

24 Months 

Cognitive 

Psychomotor 

36 Months 

Cognitive 

Psychomotor 

Simple 

Maternal 

-.044 

.035 

.015 

.081 

.006 

.021 

.040 

.015 

Correlations 

Cord 

-.061 

.025 

.153* 

.167* 

.053 

-.060 

.045 

.010 

Effect of Covariates 

R* 

.136 

.153 

.177 

.106 

.219 

.081 

.165 

.081 

F p 

3.16 <.001 

3.63 <.001 

df = 13,261 

4.08 <.001 

2.23 <.01 

df = 13,245 

4.79 <.001 

1.49 >.10 

df = 13,222 

3.11 <.001 

1.39 >.10 

df = 13,204 

Incremental Effect 

of Lead 

R

z

 F p 

.008 1.22 >.25 

.003 0.46 > .60 

df = 2,259 

.008 1.17 >.30 

.019 2.60 .08 

df = 2,243 

.001 .19 >.70 

.013 1.54 >.20 

df = 2,220 

.001 .13 >.70 

.001 .10 >.90 

df = 2,202 

*p < .05, two-tailed test. 

Source: Cooney et al. (1989a). 

the regression model approached significance only for the positive relationship between cord 

blood lead and 12-month Bayley PDI (p = 0.08, two-tailed). 

Regression analyses and path models indicated that the greatest influences on MDI and 

PDI scores at 6 and 12 months were gestational age and HOME scores, while at later ages 

HOME and parental characteristics (maternal IQ and education) were the more important 

influences on cognitive measures. The fact that Bayley scores accounted for a significant 

amount of the variance in regressing HOME scores on prior HOME scores in two instances 

(12-month PDI for the 12- versus 6-month HOME and 36-month MDI for the 36- versus 

24-month HOME) raises the question of whether lead exposure (e.g., maternal blood lead) 

might have covaried with HOME scores. This particular relationship apparently was not 

40 
P.52 

examined by the investigators. However, the authors did examine the relationships between 

maternal/cord blood lead and gestational age, which were shown to be statistically 

nonsignificant (as were also birth weight, obstetrical complication, and postnatal risk factors). 

Cooney et al. (1989b) reported the results of testing the Sydney cohort at 4 years of age 

on the McCarthy GCI and Motor Scales. At 4 years, 207 children of the original cohort 

remained in the study. In addition to maternal and cord blood lead samples at delivery, 

postnatal blood lead samples were taken every 6 months. Geometric means for capillary and 

venous blood samples combined rose from 15 /tg/dl at 6 months to a peak average of 

16.4 /tg/dl at 18 months and then declined to 10.1 /tg/dl at 48 months. 

Bivariate and partial (conecting for venous versus capillary collection) conelations 

between 48-month McCarthy Scales and blood lead levels at different ages were generally 

quite small, mixed in sign, and uniformly nonsignificant. Analyses using composite blood 

lead measurements (averaged over 12-month periods) produced only one significant 

relationship, a. positive conelation (r = 0.160; p <0.05, two-tailed) between first year blood 

lead and GCI performance (Table 9). However, after allowing for covariates, regression 

analysis showed the relationship to be only marginally significant (p <0.07). Analysis of 

covariance did not indicate that change in developmental outcome from 36 to 48 months was 

significantly related to either cunent (48 month) or cumulative (cunent and prior) past lead 

exposure, although cumulative lead was a better predictor (p = 0.14) than cunent lead 

exposure alone (p = 0.36) for GCI scores. HOME score was stated to be the most important 

covariate for the GCI, apart from the 36-month GCI. 

Mexico City 

Preliminary results of a pilot study in Mexico City for a longitudinal investigation of 

developmental outcomes related to lead exposure and other factors have been reported by 

Rothenberg et al. (1989a,b). Approximately 50 mothers were sampled for blood lead levels 

at 36 weeks (M36) of pregnancy and delivery (MD); umbilical cord blood lead (UC) was also 

sampled at delivery. Mean blood lead levels were: M36, 15.0 /tg/dl; MD, 15.4 /tg/dl; and 

UC, 13.8 /tg/dl. The Brazelton Neonatal Behavioral Assessment Scale (NBAS) was 

administered to the infants at 48 hours and 15 and 30 days after birth. The data were 

analyzed by calculating the trend of the NBAS subscale scores over the first 30 days by linear 

41 
P.53 

TABLE 9. REGRESSION OF DEVELOPMENTAL INDICES AT 48 MONTHS 

ON CURRENT AND PRIOR BLOOD LEVELS 

Age 

Prenatal: 

Cognitive 

Motor 

First Year: 

Cognitive 

Motor 

Second Year: 

Cognitive 

Motor 

Third Year: 

Cognitive 

Motor 

Fourth Year: 

Cognitive 

Motor 

All Prior and Cunent: 

Cognitive 

Motor 

AR2 

.004 

.001 

.023 

.003 

.001 

.001 

.001 

.008 

,001 

.005 

.044 

.022 

Unadjusted 

df 

2,204 

2,204 

1,205 

1,205 

1,205 

1,205 

1,205 

1,205 

1,205 

1,205 

6,200 

6,200 

P 

.36 

.91 

.03 

.46 

.82 

.60 

.70 

.20 

.88 

.32 

.17 

.72 

AR2 

.009 

.002 

.013 

.001 

.002 

.004 

.003 

.006 

.001 

.001 

.028 

.018 

Adjusted 

df 

2,199 

2,199 

1,199 

1,199 

1,199 

1,199 

1,199 

1,199 

1,199 

1,199 

6,193 

6,193 

P 

.14 

.55 

.07 

.67 

.60 

.33 

.76 

.26 

.76 

.96 

.14 

.56 

Source: Cooney et al. (1989b). 

regression analysis and by computing the difference in M36 and MD values or M36 and UC 

values. The relationships among the various primary and secondary measures were then 

examined through bivariate conelations and multivariate regression analyses. 

Based on preliminary data analyses reported at a 1986 conference, Rothenberg et al. 

(1989a) initially reported that statistically significant bivariate conelations were found: 

between UC blood lead and the 30-day trend in NBAS Abnormal Reflexes (p <0.05); 

between the M36-MD blood lead difference and Regulation of States (p <0.05); and between 

42 
the MD-UC blood lead difference and Abnormal Reflexes (p <0.01) — all reflecting 

impairment of function. Also, stepwise regression modeling with all covariates entered before 

lead was reported to show that: blood lead differentials (M36-MD; MD-UC) accounted for a 

significant portion of the variance in the Abnormal Reflexes trend; the M36-MD differential 

accounted for a significant amount of variance in Regulation of States; but UC alone was no 

longer significantly related to Abnormal Reflexes. Rothenberg et al. (1989a) also reported 

that evaluations of physical development outcomes at birth in their cohort indicated that: 

M36-MD and M36-UC each accounted for a significant amount of variance in birthweight 

(p <0.05); M36-UC accounted for significant amount of variance in chest circumference 

(p < 0.054); and UC and M36-UC accounted for a significant amount of the variance in trunk 

length (p - 0.06) 

Upon further analysis of their dataset, however, Rothenberg et al. (1989b) reported that 

Abnormal Reflexes were not significantly associated with any of the lead measures and no 

simple lead measure yielded significant effects on any of the perinatal outcomes in multiple 

regression analyses. On the other hand, several outcome variables (Regulation of States, 

Autonomic Regulation, and Gestation Age) were statistically significantly related to composite 

lead measures (i.e., M36-MB, MB-UC, M36-UC). Generally, increases in the blood lead of 

the mother during the last month of pregnancy (indexed by the M36-MB and M36-UC 

differentials) or an umbilical cord lead higher than the mother's blood lead (MB-UC 

differential) were associated with deleterious changes in the last-noted outcome variables. 

Rothenberg et al. (1989b) interpreted their findings as indicating that either (1) the deleterious 

effects were related to increases in blood lead levels of the mother and fetus during the last 

several weeks of pregnancy or during delivery (regardless of initial maternal blood lead values 

before the increase) or (2) the negative outcomes were related to increased stress factors 

during late pregnancy or delivery that also raised maternal and fetal blood lead levels. 

Yugoslavia 

A longitudinal study involving two communities in Yugoslavia, Titova Mitrovica and 

Pristina, has been undertaken by Graziano et al. (1989a,b; Murphy et al., 1990). 

T. Mitrovica is a major lead smelter and industrial site, whereas Pristina, 40 km to the south, 

43 
P.55 

serves as a relatively non-exposed control community. The analyses completed thus far have, 

been retrospective as well as prospective. Only reproductive outcomes have been reported. 

Of the first 1032 women enrolled in the study, 639 (304 in T. Mitrovica; 335 in 

Pristina) had had at least one pregnancy (only first pregnancies were considered in the 

analysis) and had lived at their cunent address at least since their first pregnancy. Geometric 

mean blood lead levels at the time of initial interview were 15.9 /tg/dl in T. Mitrovica and 

5.1 /ig/dl in Pristina. The rate of spontaneous abortions (fetal loss prior to 7th month) was 

not significantly different in the two communities: 16.4% of T. Mitrovica women versus 

14.0% of Pristina women reported such loss. Graziano et al. (1989b) noted that they had 

systematically attempted to demonstrate an association between lead exposure and spontaneous 

abortion and that their failure to detect such an association suggested that it did not exist at 

the levels of exposure encountered in their samples. 

A preliminary analysis of prospective data from 907 births (401 in T. Mitrovica; 506 in 

Pristina) indicated that mean birthweights did not differ significantly between the two 

communities: 3308 g (SD = 566) in T. Mitrovica versus 3361 g (SD = 525) in Pristina. 

Geometric mean blood lead levels were 17.1 /tg/dl in the former and 5.1 /tg/dl in the latter 

mothers. Regression analysis controlling for several covariates also failed to show any 

significant relationship between mid-pregnancy blood lead levels and birthweight. 

Glasgow 

Following a cross-sectional duplicate diet study that showed, among other things, an 

inverse relationship between gestational age and lead exposure during pregnancy (Moore 

et al., 1982a), Moore et al. (1982b) initiated a prospective study of the neurobehavioral 

effects of lead in children bom in Glasgow, Scotland. A major source of lead exposure for 

this population was its plumbosolvent drinking water. However, subsequent to a successful 

program to control the plumbosolvency of the water supply for Glasgow, average blood lead 

levels declined substantially (Richards and Moore, 1984). Notwithstanding this complication 

in the design of their prospective investigation, Moore et al. (1989) undertook to assess 

whether prenatal or perinatal lead exposure was associated with birth outcomes or postnatal 

neurobehavioral development. 

44 
P.56 

Their study sample consisted of 151 subjects drawn from an initial pool of 885 families. 

Based on maternal blood lead levels during pregnancy, three groups, matched for social class, 

were created: high fe30 /tg/dl, mean = 33.05), medium (15-25 /tg/dl, mean = 17.73), and 

low (^10 /tg/dl, mean = 7.02). Infant blood lead levels were measured at 1 and 2 years of 

age, but were not included in the reported analyses because of incomplete records. 

Although birth weight appeared to be inversely related to maternal blood lead (a 

reduction of nearly 100 g for each increment in blood lead grouping), no statistical analysis of 

the data was reported. Birth length showed a slight trend in the opposite direction, and other 

pediatric measures (head circumference, Apgar scores, obstetric complications) showed no 

evident trend in either direction. 

Unadjusted Bayley scores (MDI, PDI, and mean) at 1 and 2 years generally decreased 

with increasing maternal blood lead grouping. However, stepwise linear regression analyses 

indicated that birth weight, social class, and HOME scores accounted for Bayley Scales 

performance better than lead exposure (as represented by maternal blood lead, water lead 

concentration, or reported history of pica). Since birth weight was significantly related 

inversely to lead exposure, it was removed from the model to see if the explanatory power of 

one of the lead variables could be improved. Only second year pica showed a notable 

improvement, although HOME score alone still had greater predictive ability for 2-year 

Bayley scores than pica coupled with HOME score. An analysis of the Bayley IBR revealed 

no consistent direction of association between lead exposure and IBR factors. Moore et al. 

(1989) concluded that their dataset provided "no firm evidence for either a direct or an 

indirect contribution of lead to decrements in cognitive development." 

Christchmrch 

Fergusson and his colleagues (Fergusson and Purchase, 1987; Fergusson et al., 

1988a,b,c) collected shed deciduous teeth from more than 1000 children in Christchurch, 

New Zealand to assess their long-term lead exposure and the relationship of such exposure to 

neurobehavioral outcomes. Blood lead levels were not measured. 

As reported by Fergusson et al. (1988b), IQ, reading ability, and school performance 

were assessed in relation to dentine lead levels in samples of 664-886 children from an 

original cohort of 1265 drawn from the Christchurch Child Development Study. The subjects 

45 
P.57 

were evaluated at ages 8 and 9 years on the WISC-R, the Burt Reading Test, and by teachers' 

ratings of reading, written expression, spelling, mathematics, and handwriting. All bivariate 

conelations between these outcomes and dentine lead levels were in the predicted direction 

(negative), and all but one (out of 18) were statistically significant at p <0.05 (one-tailed). 

After conecting for test unreliability, dentine lead measurement enor, sample selection 

factors, and several covariates, the coefficients for the reading test and all five teacher ratings 

at both ages 8 and 9 years remained significant. Further conection for pica (to test for the 

reverse causality hypothesis that reduced cognitive ability results in more pica, which in turn 

results in greater lead exposure) reduced the size of the conelations even more (ranging 

from -0.07 to -0.14), but 7/12 of the conelations remained statistically significant at p <0.05 

(Table 10). 

TABLE 10. UNEXPLAINED CORRELATIONS BETWEEN LOG DENTINE 

LEAD LEVELS TAKING INTO ACCOUNT TEST RELIABILITY, 

CONFOUNDING COVARIATES, SAMPLE SELECTION FACTORS, 

AND REVERSE CAUSALITY VIA PICA 

Measure 

Verbal IQ 

Performance IQ 

Total IQ 

Burt reading test 

Teacher ratings 

Reading 

Written expression 

Spelling 

Mathematics 

Handwriting 

8 years 

r 

-0.03 

-0.02 

-0.04 

-0.07 

-0.13 

-0.13 

-0.14 

-0.08 

-0.12 

p* 

N.S. 

N.S. 

N.S. 

<0.05 

<0.001 

< 0.001 

<0.001 

<0.10 

<0.05 

9 years 

r 

-0.02 

-0.02 

-0.03 

-0.08 

-0.08 

-0.08 

-0.09 

-0.10 

-0.08 

p* 

N.S. 

N.S. 

N.S. 

<0.05 

<0.10 

N.S. 

<0.10 

<0.05 

<0.10 

*One-tailed test. 

Source: Fergusson et al. (1988b). 

Fergusson et al. (1988c) also investigated the hypothesis that the demonstrated 

relationship between lead exposure and school performance was due to lead-related deficits in 

46 
P.58 

attentional processes. Mothers' and teachers' ratings of signs of restless activity and 

inattention at ages 8 and 9 years were conelated with dentine lead levels of 888 children 

(Table 11). Several aspects of the ratings were significantly conelated with dentine lead, 

with bivariate conelations for total rating scores ranging between 0.08 (p <0.05) and 0.14 

(p <0.01, one-tailed). Teachers' ratings were more highly conelated with lead and were 

also judged to be a more accurate measure of the children's behavior than were maternal 

ratings. After conecting for measurement enors, sample selection factors, covariates, and 

pica, the conelation between dentine lead and inattention/restless behavior was 0.08 

(p <0.01) at both 8 and 9 years of age. 

Although small, the conected conelations between dentine lead and various outcomes 

reflecting school behavior and performance were consistent and stable over a one-year interval 

in the Christchurch study. The uniformity and stability of these results across time provide 

compelling reason to judge the effects as real. However, this same study provided no 

evidence that intelligence, as measured by the WISC-R at 8 or 9 years of age, was related to 

dentine lead levels. 

The lack of blood lead measurements makes it difficult to interpret the Christchurch 

study for dose-response information. However, the mean dentine lead value in this study was 

6 /tg/g, wliich may be compared to a mean dentine lead level of ~ 14.5 /tg/g found by 

Needleman et al. (1979; see also Bellinger et al., 1984b) for a sample of 2335 Boston 

children. Given the comparability of the methods employed by Fergusson (Fergusson and 

Purchase, 1987; Fergusson et al., 1988a,b,c) and by Needleman et al. (1979), and an 

estimated mean blood lead level on the order of 30 /tg/dl at age 2-3 years in the study by 

Needleman et aL (1979), it is possible to infer that the average blood lead level of the 

Christchurch cohort could have been roughly on the order of <15 /tg/dl at age 2-3 years. 

Although this estimate is rather imprecise and based on several assumptions, it suggests that 

the level of lead exposure in the Christchurch study population was roughly comparable to 

populations in some of the other studies under consideration here. 

Norderiham - Stollberg 

As noted in the 1986 Addendum, Winneke et al. (1985a,b) enrolled 114 children in 

1982 from a population of 383 children bom 6-7 years previously in Nordenham, Federal 

47 
TABLE 11. PRODUCT MOMENT CORRELATIONS BETWEEN MATERNAL 

OR TEACHER BEHAVIOR RATINGS AND LOG DENTINE LEAD VALUES 

Measure 

8 years 

p* 

9 years 

Maternal ratings 

Activity 

Restless, overactive 

Excitable, impulsive 

Constantly fidgeting 

Always climbing 

Squirmy, fidgety 

Attention 

Short attention span 

Inattentive, easily distracted 

Can't settle to tasks 

Total Score 

Teacher ratings 

Activity 

Restless, overactive 

Excitable, impulsive 

Squirmy, fidgety 

Very restless 

Attention 

Inattentive, easily distracted 

Short attention span 

Poor concentration 

Total Score 

0.07 

0.07 

0.03 

0.07 

0.03 

0.09 

0.09 

0.06 

<0.05 

<0.05 

N.S. 

<0.05 

N.S. 

<0.01 

<0.01 

<0.05 

0.05 

0.05 

0.04 

0.04 

0.06 

0.08 

0.05 

0.04 

<0.10 

<0.10 

N.S. 

N.S. 

<0.05 

<0.01 

<0.10 

N.S. 

0.11 <0.01 0.08 <0.05 

0.09 

0.03 

0.11 

0.10 

0.16 

0.11 

0.13 

<0.01 

N.S. 

<0.001 

<0.001 

<0.001 

<0.001 

<0.001 

0.11 

0.10 

0.13 

0.13 

0.14 

0.12 

0.12 

<0.001 

<0.001 

<0.001 

<0.001 

<0.001 

<0.001 

<0.001 

0.13 <0.001 0.14 <0.001 

""One-tailed test. 

Source: Fergusson et al. (1988c). 

Republic of Germany. An ongoing federal screening program of the Nordenham residents 

had taken blood samples from mothers and cords at delivery. The maternal geometric mean 

blood lead level was 9.3 /xg/dl (range: 4-31); for umbilical cord, it was 8.2 /tg/dl (range: 

4-30). When tested at age 6-7, the children's average blood lead level was again 8.2 /tg/dl, 

48 
but distributed differently (range: 4-23). In terms of accounting for a significant amount of 

variance in various measures of reaction time performance, maternal blood lead was better 

than cord; the combination of maternal and cord blood lead was better than either alone; and 

the combination was about as good as concunent blood lead. These relative standings are 

probably only rough comparisons, since differences in the quality of the samples from 

different sources and at different times could have affected these results (Winneke et al., 

1985a). 

Retesting of 76 of the Nordenham children at age 9, with blood lead levels then 

averaging 7.8 /tg/dl (range: 4-21), indicated some persisting deficits in reaction time test 

performance related to blood lead levels 3 years earlier (Winneke et al., 1989a,b). WISC-R 

performance was also significantly related to the preceding blood lead levels after conection 

for confounding. Concunent blood lead did not significantly account for either reaction time 

or WISC-R performance at age 9, although it had (for reaction time enors) at age 6-7. No 

results were reported for the relationship of 9-year outcomes to perinatal blood lead measures. 

Analogous results as for the Nordenham children have recently been reported by 

Winneke et al. (1989) for another cohort of 6 to 9 yr old children (N = 109) from Stolberg, 

West Germany whose blood lead levels averaged 7.4 /tg/dl (range 4.2-18.0 /tg/dl). 

Performance deficits on a serial choice reaction task were found to be significantly related to 

lead exposure after conection for confounders. The authors noted that certain features of the 

performance deficit resembled clinical observations for children with attention deficit 

disorder. 

BjrfffalQA prospective study was recently initiated in the Buffalo, NY area by Shucard et al. 

(1988a,b). Cord blood levels averaged ~4.4 /tg/dl for 802 newborns. Outcomes have not 

yet been reported. 

Other Recent Studies 

In addition to the longitudinal studies discussed above, several other studies of 

neurobehavioral function in lead-exposed children have been reported since the 1986 

Addendum. Although most appear to be well conducted, many of these studies are of 

49 
P.61 

high blood lead levels, or their primary reliance on tooth lead as an exposure indicator. 

Despite its limitations, blood lead is cunently the bioindicator of greatest utility for regulatory 

decision-making purposes. Consequentiy, the following studies will be discussed in less detail 

than the prospective longitudinal studies described above. 

A cross-sectional study of cognitive abilities and educational attainment in a population 

of school-age children from central Edinburgh, Scotland has been reported by Fulton et al. 

(1987) and Raab et al. (1989). The geometric mean blood lead level for the 501 children in 

the study sample was 11.5 /tg/dl (range: 3.3-34.0). Multiple regression analyses indicated 

significant relationships between log-transformed blood lead levels and composite scores on 

the British Ability Scales (p = 0.003) and between blood lead levels and attainment test 

scores for quantitative (p = 0.04) and reading (p = 0.001) skills, even after allowing for 

33 covariates. Grouping the subjects by blood lead levels showed a clear dose-response 

relationship without any evident threshold down to the lowest subgroup mean blood lead level 

of 5.6 /tg/dl. More recently, Thomson et al. (1989) reported that significant associations 

were found for the Fulton Edinborough study cohort between log blood lead levels and 

teacher's ratings on the Rutter scale of aggressive/anti-social hyperactive, and overall deviate 

behavior. 

Hatzakis et al. (1987, 1989) conducted neuropsychological testing on 509 children living 

near a lead smelter in Lavrion, Greece, and found impairments in WISC-R IQ scores and 

reaction time performance scores. These effects were significantly associated with blood lead 

levels after controlling for as many as 23 covariates in multiple regression models. Blood 

lead concentrations ranged from 7.4 to 63.9 /tg/dl and averaged 23.7 /tg/dl. Depending on 

the number of .covariates included in the model, full scale IQ decreased by 2.4-2.7 points for 

every 10-/ig/dl increase in blood lead concentration. Subjects were grouped by blood lead 

levels (10-/tg/dl increments) to analyze dose-response relationships for IQ as well as reaction 

performance scores. A nonlinearity in the dose-response pattern for IQ makes it difficult to 

interpret these data for a threshold. However, reaction time performance showed no evident 

threshold. 

Wolf et al. (1987) used structural equation modeling to evaluate the variables related to 

low-level lead exposure and infant mental development in an urban population in Costa Rica. 

Blood lead levels averaged 10.8 /tg/dl (range: 5.4-21.5 /tg/dl) in 182 subjects whose average 

50 
P.62 

Blood lead levels averaged 10.8 /tg/dl (range: 5.4-21.5 /tg/dl) in 182 subjects whose average 

age was 16.6 months (range: 12-23 months). Blood lead had neither direct nor indirect 

effects on the Bayley MDI. However, blood lead did have a significant negative relationship 

with birth weight (p = 0.05) and mother's height (p = 0.001), as well as a positive 

relationship with mother's age (p = 0.01), after adjustment for covariates. Birth weight, iron 

deficiency anemia, and child's age had significant direct effects on MDI. 

Vivoli et al. (1989) and Bergomi et al. (1989) measured lead concentrations in blood, 

teeth, and hair, as well as ALA-D activity, in 237 children from Sassuolo, Italy. At the time 

of the study, blood lead levels averaged ~ 11.5 /tg/dl. Although concunent blood lead 

measures showed no significant relationship to covariate-adjusted scores on any of six 

neurobehavioral tests, tooth lead was significantly related to full scale and verbal WISC-R 

scores. Moreover, ALA-D was significantly associated with one verbal subtest of the WISCR and with delayed reaction time performance. 

Tooth lead also accounted for a significant amount of the variance in WISC-R full scale 

and verbal IQ as well as Bender Gestalt visual motor test performance in 156 children from . 

Aarhus, Denmark (Hansen et al., 1989). Circumpupal tooth lead levels averaged 10.7 /tg/g. 

A study of children in Brussels (Cluydts and Steenhout, 1989) showed marginally significant 

(p <0.10) covariate-adjusted regression coefficients for tooth lead and neurobehavioral 

outcomes (WISC-R, reaction time and attentional performance), despite the small (N = 41) 

number of children evaluated. 

In separate but similar studies conducted in Rhode Island and Lavrion, Greece, Faust 

and Brown (1987) and Benetou-Marantidou et al. (1988), respectively, administered 

neurobehavioral test batteries to small groups (N's of 15-30) of children whose blood lead 

levels had been measured at ~ 30-60 /tg/dl. Significantly impaired performance was evident 

by comparison to matched controls, even after blood lead levels had been below 30 /tg/dl for 

at least one year (Faust and Brown, 1987) or after a 4-year intervening period before followup testing (Benetou-Marantidou et al., 1988). 

In analyses of data from the second National Health and Nutrition Examination Survey, 

Schwartz and Otto (1987) have also found evidence of retarded neurobehavioral.development 

and impaired neurosensory capability in relation to low-level lead exposure. The ages at 

which a child first sat up, walked, and spoke were significantly associated with blood lead 

51 

BH 
P.63 

with other evidence linking low-level lead exposure to electrophysiological changes in the 

auditory system (reviewed in U.S. Environmental Protection Agency, 1986a). 

With regard to fetal growth effects, a recent report by Ward et al. (1987) indicated that 

placental lead concentrations were highly significantly conelated with reductions in birth 

weight, head circumference, and placental weight in 100 obstetrically normal births in 

England. Gestational age was conelated at borderline statistical significance (p <0.10). 

Dividing the data into two birthweight groups, low (OOOO g) and high.(>4000 g), revealed 

a highly significant difference in mean placental lead concentrations: 2.349 /tg/g (S.D. = 

0.883) versus 1.122 /tg/g (S.D. = 0.361) for low and high weight groups, respectively. Of 

the 37 elements analyzed, lead and cadmium showed the most consistent negative 

relationships with these fetal outcomes. Other factors, such as parity, sex of neonate, social 

class, and history of miscarriage, did not appear to the authors to significantly confound their 

results. 

Cottdtosions 

As noted in the 1986 Addendum (U.S. Environmental Protection Agency, 1986b), 

prospective studies offer a major advantage over cross-sectional and even many retrospective 

studies in that they provide a better history of lead exposure. This key difference is the 

reason why more weight is placed on findings from prospective studies. However, 

notwithstanding this advantage, prospective studies may have various types of shortcomings in 

common with other epidemiological studies. One problem of particular importance is 

statistical power, especially in longitudinal studies that typically experience attrition and 

declining sample sizes over the course of the study. 

The studies reviewed here differ considerably in population size and, even within some 

studies, in the number of subjects included in individual analyses. To illustrate, the first 

results from the Boston prospective study were based on analyses involving 201 infants out of 

a cohort of 249 (Bellinger et al., 1987a). By 5 years, the analyses were based on 170 

children (Bellinger et al., 1989b). While this rate of attrition is certainly not high, it does 

nevertheless make it increasingly difficult to detect an effect of low-level lead exposure. 

According to Cohen (1977), an N of over 400 subjects would be required to detect an effect 

size (AR2) of 0.01 with a power of 0.80 at an alpha of 0.05, one-sided. This calculation 

52 
, assumes that the analysis includes 13 covariates having an R2

 of 0.30. If the covariate R2 is 

overestimated in the analysis for some reason (e.g., because of spurious sample conelations 

between covariates and blood lead or because the covariates are involved in causal pathways 

linking lead exposure to development), then regression methods may fail to detect a truly 

significant association between lead and developmental outcomes. 

Another factor that can affect the power to detect a significant relationship in these 

studies is the amount of variance in the sample blood lead measures. A small standard 

deviation in the independent variable (e.g., blood lead) will necessarily reduce any conelation 

between it and a dependent variable (e.g., MDI scores). Also, the presence of a significant 

effect of another variable on the outcomes under consideration will make it more difficult to 

detect the effect of lead exposure by regression analysis. For example, over 50 percent of the 

mothers enrolled in the Cleveland study were determined to be alcoholic, and significant early 

developmental effects were shown to be alcohol-related in the Cleveland study population 

(Ernhart etal., 1985b). 

Perhaps the most surprising outcome of the prospective studies as a whole, then, is that 

so many of them are able to detect any effect of lead at all. Given the limitations of sample 

size and power of most of the studies under consideration, one's confidence in the reality of 

any detected effects that achieve statistical significance is enhanced. Conversely, it is difficult 

to interpret a failure to detect a lead effect as suggesting the absence of an effect, if the 

analyses in question were based on fewer than 400 subjects, as was generally the case. 

Despite these limitations, some important conclusions emerge from the findings reported 

above. Various lines of evidence still relate neurobehavioral effects to blood lead levels of 

"10-15 /tg/dl, and possibly lower," as was previously concluded in the 1986 Addendum (U.S. 

Environmental Protection Agency, 1986b). Further analyses from the Boston study, which 

has provided the most direct information bearing on dose-response relationships for 

neurobehavioral effects, not only supported the 10-15 /tg/dl level of concern but indicated that 

MDI deficits can be detected in relation to cord blood lead levels of 6-7 /tg/dl in lower SES 

children (Bellinger et al., 1988). Since the Boston cohort was mostly middle to uppermiddle class, "lower" SES merely refers to less than the highest SES levels and is probably in 

fact much closer to the median of the U.S. population than the term suggests. Although the 

postnatal lead exposure levels were somewhat higher in the Port Pirie study, analyses of the 

53 
P.65 

relationship between postnatal blood lead levels and covariate-adjusted MDI scores provided 

"no evidence of a threshold effect" (Wigg et al., 1988). Indeed, restricting the analysis to 

children with blood lead levels below 25 /tg/dl in the Port Pirie study yielded an even 

stronger association between covariate-adjusted McCarthy GCI scores and integrated postnatal 

blood lead measures (McMichael et al., 1988). 

Supporting evidence for the stated level of concern may also be derived from other 

studies. Although McCarthy Scale results from the Boston study have not been analyzed in a 

manner to allow direct extraction of dose-response information, the average blood lead level 

significantly associated with covariate-adjusted performance on the McCarthy Scales (GCI and 

Perceptual-Performance subscale) was 6.8 /tg/dl (SD = 6.3) (Bellinger et al., 1987b). 

Similarly, analyses relating cord blood lead levels to the G-R Neurological Soft Signs and, 

indirectly, to 12-month MDI scores were previously reported for the Cleveland study (Emhart 

et al., 1986; Wolf et al., 1985), along with more recent significant results, by more than one 

analysis, relating cord blood lead to Length of Utterance, a measure of language development 

in 24-month-old infants (Monow-Tlucak and Emhart, 1987). Significant relationships were 

also indicated for maternal blood lead and deficits in 6-month MDI, PDI, and KID scores in 

the Cleveland cohort, but results of other analyses were mixed (Emhart et al., 1987). Any 

significant neurobehavioral effects associated with cord blood lead in the Cleveland study 

necessarily occuned at levels below 15 /tg/dl since the maximum single cord blood lead level 

measured was only 14.7 /tg/dl (mean: 5.84 /tg/dl). Moreover, the fact that significant 

evidence of lead-associated impairments was found in a relatively small cohort (N's <200) 

suggests that the work was carefully conducted and that the results are credible. 

Some evidence from recent cross-sectional studies is also consistent with the identified 

level of concern. The Edinburgh study (e.g., Fulton et al., 1987) shows a significant 

relationship between blood lead levels averaging as low as 5.6 /tg/dl and covariate-adjusted 

scores of cognitive ability and educational attainment. As noted by Grant and Davis (1989), 

this finding appears to closely parallel the results of Schroeder and Hawk (1987), whose 

North Carolina cross-sectional study population showed IQ deficits in relation to blood lead 

extending to levels as below 10 /tg/dl. The latter study was described in detail and evaluated 

in the 1986 Air Quality Criteria for Lead (U.S. Environmental Protection Agency, 1986a). In 

54 
P.66 

addition, the reaction time performance results of Hatzakis et al. (1989) showed no evident 

threshold over a blood lead range of 7.4 to 63.9 /tg/dl. 

Based on all of the above considerations, a blood lead concentration of 10-15 /tg/dl, and 

possibly lower, remains the level of concern for impaired neurobehavioral development in 

infants and children. Given the fact that such effects have been associated with blood lead 

measures in pregnant women, umbilical cords, and infants up to at least 2 years of age, there 

is no apparent distinction at present as to whether this level of concern applies to only fetuses 

or infants or preschool-age children. Thus, a blood lead level of 10-15 /tg/dl, and possibly 

lower, ought to be avoided in pregnant women, fetuses, infants, and young children, although 

it is recognized that pregnant women per se are not necessarily a population at risk. 

Various lines of evidence suggest that lower SES and male gender may be additional 

risk factors for the developmental effects of low-level lead exposure. Increased vulnerability 

in lower SES children has been indicated in analyses of MDI scores (Bellinger et al., 1988) 

and MDI-GCI change scores in the Boston study (Bellinger et al., 1989b) and in the MDI 

results from Cincinnati (Dietrich et al., 1987b, 1989b). Also, some IQ results from crosssectional investigations are consistent with the view that lower SES children are more 

vulnerable to lead-induced cognitive impairment (Harvey et al., 1984; Lansdown et al., 1986; 

Schroeder and Hawk, 1987). Greater susceptibility of male infants to lead developmental 

toxicity has been evident in analyses of MDI-GCI change scores in the Boston study 

(Bellinger et al., 1989b), in MDI results from the Cincinnati study (Dietrich et al., 1987b, 

1989b), in reanalyses of IQ data from the Southampton cross-sectional study (Pocock et al., 

1987), and in some early data on sex ratios of stillbirths in Port Pirie and other locations 

(Scragg et al., 1977). However, McMichael et al. (1988) found that the blood lead-GCI 

regression coefficient was slightly higher for girls than for boys, although stratification by 

gender did not significantly improve their model in repeated analyses. 

The evidence regarding pregnancy outcomes and physical growth effects related to 

prenatal lead exposure is less consistent than that for neurobehavioral outcomes. Such was 

the case at the time of the 1986 Addendum (U.S. Environmental Protection Agency, 1986b) 

and is still the case simply because relatively little additional information pertaining to 

pregnancy outcomes and growth has appeared since the 1986 assessment. 

55 
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As far as growth effects are concerned, the Cincinnati study has shown a significant 

covariate-adjusted reduction in birth weight associated with prenatal (maternal) blood lead 

levels (Bornschein et al., 1989). In addition, early postnatal growth rates (over 3-15 months) 

have also been associated with lead exposure pre- and postnatally in the Cincinnati study 

(Shukla et al., 1987, 1989). (Interactions with other variables were evident in both of these 

cases, which will be discussed further below.) 

Some supporting evidence of lead-related reductions in birth weight also comes from the 

Boston study and, possibly, the Glasgow study. Although birth weight per se showed no 

relationship to cord blood lead in the Boston study, there was an exposure-related trend in the 

percentage of small-for-gestational age infants that approached statistical significance 

(Bellinger et al., 1984a). In Glasgow, the high maternal blood lead infants (mean: 33.05 

/tg/dl) weighed 3.32 kg, on average, at birth, whereas the medium blood lead group (mean: 

17.73 /tg/dl) weighed 3.43 kg and the low group (mean: 7.02 /tg/dl) weighed 3.51 kg (Moore 

et al., 1989). However, these data were not adjusted for covariates and were not tested for 

statistical significance. 

No other prospective study has shown a significant association between reduced birth 

weight and lead exposure. The Yugoslavian study (Graziano et al., 1989b; Murphy et al., 

1990), in particular, has failed thus far to yield any evidence of lead-related birthweight 

reductions in more than 900 births, even at relatively high blood lead levels. Also, analyses 

by Ernhart et al. (1986) showed no significant effect of lead on birth weight, birth length, or 

head circumference; nor, according to preliminary analyses, was any effect evident on 

postnatal growth (Marler and Emhart, 1987). However, the cross-sectional study by Ward et 

al. (1987) did indicate highly significant simple relationships between placental lead 

concentrations and reduced birth weight and head circumference. 

The Port Pirie study provided strong evidence relating prenatal lead exposure to 

increased risk of preterm delivery (< 37 weeks gestation) in a sample of 749 pregnancies 

(McMichael et al., 1986). Also, a small but significant relationship between prenatal lead 

and gestational maturity was observed in the structural analyses of the Cincinnati study (e.g., 

Dietrich et al., 1987b). However, regression analyses of a different sample from the 

Cincinnati study did not reveal a significant association between prenatal lead and preterm 

(<35 weeks) deliveries (Bornschein et al., 1989). Also, as previously discussed in the 1986 

56 
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Addendum, the Boston prospective study showed a positive but nonsignificant relationship 

between cord blood lead and gestation length (Bellinger et al., 1984a), whereas the 

cross-sectional study of Moore et al. (1982a) found significant negative associations between 

gestational age and maternal as well as cord blood lead after allowing for a number of 

covariates. A reduction in gestational age was associated with an increase in maternal blood 

lead level between the 36th week of pregnancy and delivery in the Mexico City pilot study 

(Rothenberg et al., 1989b). Placental lead concentrations appeared to be inversely related to 

gestation length in the analyses of Vimpani et al. (1989) and Ward et al. (1987), but were of 

uncertain statistical significance. 

There are many possible explanations for these apparent inconsistencies among studies. 

The lack of a significant difference in pregnancy outcomes between the high exposure and 

reference communities in Yugoslavia (Graziano et al., 1989a,b; Murphy et al., 1990) could 

reflect the difficulty of assessing fetal lead exposure by means of maternal blood lead levels. 

With regard to gestation length, Bornschein et al. (1989) noted that their analysis was 

restricted to pregnancies of at least 35 weeks. A similar cutoff of 34 weeks was used in the 

Boston prospective study (Bellinger et al., 1984a). Constraining the data in this fashion could 

make it more difficult to detect an effect of lead on gestation. Also, there were interactions 

involving mother's age and race evident in the Cincinnati study (Bornschein et al., 1989). 

Thus, differences in the age of the mothers, racial make-up, other population characteristics, 

sample sizes, level of lead exposure (past as well as cunent), and approaches to analyzing 

data could underlie the varying results of different studies. 

Based on the evidence reviewed here and in the 1986 Air Quatity Criteria for Lead 

(U.S. Environmental Protection Agency, 1986a), it seems likely that prenatal lead exposure 

poses a potential hazard to the developing fetus in terms of reduced gestational length and 

possibly other aspects of fetal growth (in addition to postnatal neurobehavioral development, 

as already noted above). It is difficult, however, to derive a definitive dose-response 

relationship for fetal outcomes from the available data, although some indications point to a 

level of concern starting in the region of 10-15 /tg/dl. The average maternal blood lead levels 

in the studies where the pre-term delivery effect was clearest (Port Pirie and Glasgow) were 

in the 10-15 /tg/dl range, in contrast to somewhat mixed findings in the Cincinnati, Boston, 

and Cleveland studies where the maternal or cord blood lead levels averaged below 10 /tg/dl. 

57 
P.69 

A similar pattern seems to hold for birth weight as well. The strongest evidence of a 

birthweight effect comes from the Cincinnati study, with some of their analyses suggesting 

that such an effect could start in the region of 12-13 /tg/dl, but possibly extending from 7 to 

18 /tg/dl. However, other prospective studies provide no support for this conclusion, and so 

it must be considered an open issue awaiting more definitive resolution. 

The issue of the persistence of the neurobehavioral effects needs to be considered in an 

assessment of the risk of low-level lead exposure. If decreased scores on the Bayley Scales 

reflected merely a transient delay in children's neurobehavioral development - a minor 

perturbation that children could quickly "grow out of - then the public health significance of 

lead exposure at blood lead levels of 10-15 /tg/dl would perhaps be diminished. Some recent 

results from two of the studies that had convincingly demonstrated a link between prenatal 

lead exposure (either matemal or cord blood lead levels) and early deficits on the Bayley 

Mental Development Index now suggest that the association between prenatal lead exposure 

and cognitive development may not hold up at later ages. The Cincinnati study found a 

declining influence of prenatal exposure indicators on MDI scores at 12 months and 

24 months: only 10-day blood lead measures were significantly associated with MDI scores 

at 12 months, and no significant negative PbB-MDI relationships were evident at 24 months 

(Dietrich et al., 1989b). Although the Boston study did continue to find a significant 

relationship between prenatal lead exposure (cord blood lead) and MDI performance at 

24 months, no relationship could be shown for prenatal lead exposure and cognitive abilities 

at 57 months on the McCarthy Scales (Bellinger et al., 1987b). Thus,, one implication of 

these findings might be that the effects of prenatal lead exposure on neurobehavioral 

development are not permanent. 

Such a conclusion could be valid, but the evidence available to support it is not 

adequate. The inability to detect a continuing significant association between prenatal lead 

exposure and neurobehavioral function at age 2 in the Cincinnati study or age 5 in the Boston 

study could result from, among other things, a lack of adequate statistical power, as noted 

above. Other factors could also interfere with detecting such a relationship and possibly 

account for differing results in separate studies. For example, it is not clear which measure 

of blood lead provides the best indicator of exposure during critical periods of organogenesis. 

The Cincinnati study primarily used maternal blood samples obtained during the first or 

58 
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second trimester of pregnancy to indicate prenatal lead exposure. Prenatal exposure was also 

reflected in samples taken from the Cincinnati infants 10 days after birth (gestationally 

conected), but comparatively few cord blood samples were available for data analysis. The 

Boston study relied exclusively on cord blood samples while other studies have also sampled 

the mother's blood around the time of delivery. As suggested previously (U.S. 

Environmental Protection Agency, 1986b; Davis and Svendsgaard, 1987) and supported by 

preliminary results from the Mexico City study (Rothenberg et al., 1989a,b), there may be 

differences in these measures, particularly during the last month or so of pregnancy, that 

reflect important biokinetic transfers of lead between compartments, both within the mother 

and infant individually and between them. 

In addition to the nature and extent of such biokinetic transfers, their timing (and the 

point at which blood lead is sampled) could make an important difference in the ontogenesis 

of neural structures and, consequently, later neurobehavioral function. It is interesting to 

note, for example, that almost no significant negative associations have been found, after 

covariate adjustment, between concunent blood lead levels and postaatal outcomes in any of 

the prospective studies reviewed here. Rather, cumulative past exposure (e.g., average 

postnatal blood lead levels) or, in several instances, blood lead levels several months or years 

prior to a given outcome have shown the strongest relationship to postnatal neurobehavioral 

effects. Such a lagged effect has been suggested by several findings, as shown in Table 12. 

When significant positive (unpredicted) relationships between a measure of blood lead and 

some outcome have occasionally been found, they generally involved concunent blood lead 

levels. 

Thus, it may well be that the ability to detect a significant relationship between lowlevel lead exposure and neurobehavioral outcomes depends, at least in part, on where and 

when the measure of blood lead is obtained. Since the fetus and infant constitute the 

population at greatest risk, it would be preferable to measure a direct indicator of their 

exposure. However, that approach has not been as feasible as measuring a related indicator, 

such as maternal or cord blood lead. (Even cord blood is sampled from the placental rather 

than the fetal side and may therefore not fully reflect fetal exposure.) Consequently, 

measures such as matemal and cord blood lead levels, while reasonably good indicators of 

59 
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TABLE 12. STRONGEST RELATIONSHIPS BETWEEN BLOOD LEAD MEASURES' 

AT SPECIFIED TIMES AND LATER NEUROBEHAVIORAL OUTCOMES 

AS DETECTED BY PROSPECTIVE STUDIES 

Study 

Boston 

(Bellinger 

etal., 1987a,b) 

Cincinnati 

(Dietrich 

etal., 1987b, 

1988; Bhattacharya etal., 

1988, 1989) 

Cleveland 

(Ernhart 

etal., 1986, 

1987; Wolf et 

al., 1985; 

Monow-Tlucak 

& Ernhart, 1987) 

Pt. Pirie 

(Wigg et al., 

1988; McMichael 

etal., 1988) 

Sydney 

(Cooney etal., 

1989b) 

Time/Type 

of Blood Lead Measure 

Delivery/cord 

2yr 

Prenatal (x: 16 wk)/maternal 

10 day 

2yr 

Delivery/cord 

Delivery/maternal 

Delivery/cord 

6 month 

6, 24, 36 month and 

integrated postnatal 

Cumulative (prior plus 

cunent) 

Outcome 

6-, 12-, 18-, and 

24-month MDI 

5-yr McCarthy 

3- and 6-month MDI 

12-month MDI 

6-yr postural sway 

12-month MDI (via 30-day 

G-R soft signs) 

6-month MDI, PDI, KID 

2-yr language acquisition 

(2 measures) 

2-yr MDI 

4-yr McCarthy 

36 to 48 month change in 

cognitive development (p = 

0.14, two-tailed) 

prenatal lead exposure, may not afford the most accurate predictors of later neurobehavioral 

outcomes. 

Another factor that could obscure a relationship between prenatal lead exposure and 

postnatal neurobehavioral function and that could account for some differences in results 

60 
P.72 

among studies is the rather precipitous increase in lead exposure observed in most of the 

prospective studies during the first 2-3 years of life. Sizeable rises in postnatal blood lead 

levels were noted in the Cincinnati, Cleveland, and Port Pirie studies, but not in the Boston 

study. None of the first three studies showed a significant association between prenatal lead 

exposure indicators and 2-year MDI scores, but the Boston study did. While analyses from 

the Boston study also indicated that continuing "high" lead exposure (> 10 /tg/dl) contributed 

to a persistence of neurobehavioral deficits at later ages, a high postnatal blood lead in the 

Boston cohort was not as high as the average for any of the other prospective studies at age 

2 years. Thus, the relationship between lead exposure at a critical stage of early development 

and subsequent neurobehavioral function could be obscured by differential lead exposure 

(most often, but not always, increased exposure) in the intervening period. 

By this line of reasoning, one would perhaps expect that increased postnatal blood lead 

measures should then show a significant relationship to later neurobehavioral outcomes. This 

was in fact the case in the Port Pirie study, where sample size was more clearly large enough 

to afford relatively good statistical power. (This also points up, as discussed above, why it is 

not possible at present to discriminate between prenatal and postnatal lead exposure in stating 

a blood lead level of concern for developmental neurobehavioral effects.) 

It should also be kept in mind that scores on tests such as the Bayley or McCarthy 

Scales are only indicators or reflections of neurobehavioral function. Such variables may be 

valid and reliable measures, but they do not fiilly represent all aspects of a child's cognitive, 

social, and emotional development. Thus, considerable caution must be exercised in drawing 

conclusions from findings of "no effect." Not only are there many facets to a child's 

development that need to be assessed, but these facets may interact in complex ways that may 

be quite difficult to detect or evaluate. For example, a child's emotional and social adaptation 

(including such notions as "self-esteem") may be influenced in subtle as well as obvious ways 

by his or her cognitive abilities. Findings of lead-related aggressive/antisocial behavior in the 

Edinburgh cohort (Thomson et al., 1989) support this view. 

Such complexities make it difficult to presume that a failure to detect a continuing 

association between an indicator of prenatal lead exposure and, for example, scores on the 

McCarthy Scales at 5 years is evidence of no permanent effect. It is well known that the 

nervous system is capable of adapting to and even compensating for various insults during 

61 
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early development. But it is also true that the full realization of developmental potential can 

very much depend on events during critical stages of ontogeny. A parallel may be drawn 

with impaired language acquisition in children whose hearing has been affected by chronic 

otitis media (e.g., Kavanagh, 1986). Although otitis media itself may be transient and fully 

reversible, its secondary effects on language development in young children may be much 

longer lived. As noted by Jenkins (1986), "With a fluctuating hearing loss, such children 

may receive inconsistent and inadequate information, or may have to devote so much attention 

to the decoding process itself that there is little capacity left over for higher-order cognitive 

operations. If children are, in fact, doing something different or expending more resources 

on lower levels of speech perception, we may see deficits in other processes at a later 

period." 

Important new evidence for the long-term persistence of lead-induced intellectual and 

behavioral deficits has recently been reported by Needleman et al. (1990). In an 11-year 

follow-up of children studied earlier at age 6-7 years, those children originally having high 

dentin lead levels (>20 ppm) had markedly higher rates of having a reading disability and of 

dropping out of high school than lower lead children (dentin lead < 10 ppm). Also, higher 

lead levels in earlier childhood were significantly related to lower class standing, increased 

absenteeism, lower verbal scores, and poorer sensory-motor performance. These persisting 

effects were detected a decade later when blood lead levels had declined to 7 /tg/dl or lower 

(in contrast to the high-lead dentin group with blood lead levels averaging approximately 

34 /tg/dl at the time of the original study). 

It is also important to note the convergence of animal findings, particularly those 

showing impairments in higher-level behavioral processes such as discrimination reversal 

learning in primates as well as rodents at blood lead levels below 20 /tg/dl. Although it is 

difficult to precisely equate blood lead leads in different species, the parallels between humans 

and animals in the developmental neurotoxicity of lead are striking in many respects (Davis 

et al., 1990). Indeed, the similarity of the animal findings in the absence of socioeconomic 

and other complex variables that sometimes complicate interpretation of the human data 

provide compelling support for concluding that low level lead exposure and developmental 

neurotoxicity in children are causally related and not merely associated artifactually. 

62 
A remaining issue is whether the effects discussed here are large enough to constitute a 

significant risk to public health. Valuable information has been provided by various 

prospective studies on the magnitude of neurobehavioral deficits relative to increments in 

blood lead levels. Despite different approaches to treating their data, three of the prospective 

studies provide results suggesting that Bayley MDI scores decline by 2-8 points for 

approximately every 10-/tg/dl increase in blood lead level. The Boston study found 4-8 point 

differences in 6- to 24-month MDI scores between high (mean: 14.6 /tg/dl) and low (mean: 

1.8 /tg/dl) cord blood lead groups. The Cincinnati study showed as much as an 8.4-point 

decline in boys' 6-month MDI scores for every 10-/tg/dl increment in matemal blood lead. 

Also, the Port Pirie study demonstrated a 1.6-point decrease in the 24-month MDI per 

10 /tg/dl of blood lead at 6 months, and a ~3.5-point decrease in GCI scores for every 

10 /tg/dl in average postnatal blood lead. 

The implications of such deficits are more apparent when considered in population rather 

than individual terms (Needleman, 1983; Davis, 1990). Figure 2 illustrates the effect of 

shifting a normal distribution of MDI scores downward by 4 points. The lighter shaded area 

is 50% of the darker area, which represents the children in a population who score 80 or 

lower. Thus, even a 4-point shift in average performance results in a considerable increase in 

subnormal scores. Since the MDI has a standard deviation of 16, a score of 80 is one 

standard deviation below a mean of 96 and provides a convenient reference point for 

illustration purposes. However, the impact of such a shift appUes across the entire 

distribution of scores, reducing the number of children scoring above the norm as well as 

increasing the number scoring below the norm (Weiss, 1988). 

63 
CTl 

-P. 

25 -

20 

X 

H5 

no -

I0L, 5 

40 60 80 100 H20 MO •J S O 

Figure 2. Normal distributions with means of 96 and 100 and standard deviations of 16. 

Source: Davis (1990). 
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