:::
Methemoglobinemia in an Infant -- Wisconsin, 1992
=================================================
SOURCE: MMWR 42(12) DATE: Apr 02, 1993
Methemoglobinemia among infants is a rare and potentially fatal condition
caused by genetic enzyme deficiencies, metabolic acidosis, and exposure to
certain {*filter*} and chemicals. The most widely recognized environmental cause of
this problem is ingestion of nitrate-containing water. Ingestion of copper
causes abdominal discomfort, nausea, diarrhea, and in cases of high-level
exposure, vomiting. This report summarizes an investigation by the Division of
Health, Wisconsin Department of Health and Social Services, of
methemoglobinemia associated with ingestion of nitrate- and copper-containing
water in an infant during 1992.
A 6-week-old girl (birth weight: 7 lbs 9 oz) was hospitalized June 1 for
treatment of dehydration. On admission she weighed 6 lbs 10.5 oz and appeared
"dusky." She was afebrile and had no signs of infection. A history obtained
from her parents indicated that during her first 3 weeks she had appeared well
and had consumed approximately 20 ounces per day of soy-based formula
(consisting of a liquid concentrate diluted with 1 part water). During her 5th
week, she developed loose stools and began to vomit after eating.
Diagnoses on admission included vomiting with failure to thrive and
dehydration secondary to vomiting. She was treated and was discharged on June
2. On June 8, because of an acute weight loss (6 oz) and limited consumption
of formula (less than or equal to 3 oz) during the previous 24 hours, she was
readmitted to the local hospital. On admission, she weighed 6 lbs 12 oz and
appeared cachectic. Her hemoglobin level was 13 g/dL, with 21.4%
methemoglobin. She continued to vomit yellow- to blue-tinged liquid following
ingestion of fluids. Methemoglobinemia was diagnosed, and supportive
treatment, including {*filter*}fluids and oxygen, was initiated. Within 24 hours,
her methemoglobin level declined to 11.1%. Further evaluation at a referral
center did not identify any underlying medical problems. Since discharge, her
parents have used bottled water for drinking and for preparation of formula
and food.
The family's house was situated on a river between a river bank and
approximately 100 acres of corn and alfalfa. Water was supplied by a 28-foot
deep vacuum-sandpoint well located in a ba{*filter*}t pump room. Water used for
drinking and cooking.net">food preparation was filtered by a reverse-osmosis (R/O) unit
installed for nitrate removal when the family purchased the house in 1989.
Water samples collected from the R/O unit and from the well during the
infant's hospitalization contained 9.9 mg/L and 58 mg/L nitrate-N *,
respectively. During the investigation in late July, the well water contained
Health InfoCom Network News Page 15
Volume 6, Number 8 April 4, 1993
39.6 mg/L nitrate-N and was free of coliform bacteria. An early morning first
draw sample collected from the kitchen faucet contained 7.8 mg/L copper **.
Results of tests for corrosivity included a pH of 6.3 and an alkalinity of 16
mg/L (as CaCO superscript ((3))). Flushing the kitchen faucet for several
minutes reduced the copper level to 0.2 mg/L. A midday water sample from the
R/O system contained 0.6 mg/L copper.
Based on these analyses, the Wisconsin Division of Health recommended
that the family use bottled water for drinking and for preparation of food.
Reported by: L Knobeloch, PhD, K Krenz, H Anderson, MD, Environmental
Epidemiologist, Div of Health, Wisconsin Dept of Health and Social Svcs; C
Hovell, Trempealeau County Nursing Svcs, Whitehall, Wisconsin.
Editorial Note: In 1991 and 1992, a total of 1825 exposures to
nitrates/nitrites -- including 542 among children less than 6 years of age --
from environmental and other sources were reported to the Association of
Poison Control Centers (1,2). The most common environmental cause of
methemoglobinemia in infants in the United States is ingestion of water
contaminated with nitrates from agricultural fertilizers, barnyard runoff, or
septic-tank effluents. Acute toxicity may result after nitrate is reduced to
nitrite in the stomach and saliva (3). Nitrite reacts with the oxygen-carrying
protein, hemoglobin, reducing it to methemoglobin (Figure 1), which is unable
to transport oxygen to the tissues (4). Methemoglobin levels above 10% may
result in clinical anoxia (3), and levels above 60% can cause stupor, coma,
and death if the condition is not quickly treated.
The symptoms described in this report appear to have been induced by
simultaneous exposure to copper and nitrates at levels close to the federal
drinking water standards for these substances; this phenomenon has not
previously been implicated as contributing to the development of
methemoglobinemia in infants. Copper is an effective emetic and
gastrointestinal irritant, and ingestion of water containing copper levels of
2.8-7.8 mg/L has been associated with vomiting and diarrhea among {*filter*}s and
school-aged children (5,6). Although the dose required to cause acute symptoms
in infants is unknown, children aged less than 1 year may be more sensitive to
copper than older persons (7). Elevated copper levels in water used to prepare
the infant's formula may have caused loose stools and vomiting after eating.
Repeated vomiting and diarrhea may have resulted in dehydration and weight
loss and, in turn, reduced gastric acidity sufficiently to enhance the growth
of nitrate-reducing bacteria and facilitate conversion of ingested nitrates to
nitrites. In addition, systemic copper poisoning has been reported to increase
methemoglobin levels independent of nitrate exposure (8) -- an effect
attributed to the ability of copper to inhibit red cell enzymes needed to
reduce endogenous methemoglobin (9).
The major source of dissolved copper in drinking water is copper pipes in
household plumbing. Water that stands overnight in copper pipes may contain
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Volume 6, Number 8 April 4, 1993
copper levels that exceed the federal drinking water standard. This problem is
most often associated with corrosive water supplies or with new copper pipes
and can usually be prevented by flushing the household plumbing before using
water for drinking or cooking.net">food preparation.
This report underscores that drinking water may be contaminated with
nitrates and/or copper in some areas of the United States. Accordingly, health
practitioners should routinely advise pregnant women that water from private
wells be tested for nitrate. In addition, copper exposure should be considered
in the differential diagnosis of unexplained gastrointestinal symptoms.
References
1. Litovitz TL, Holm KC, Bailey KM, Schmitz BF. 1991 Annual report of the
American Association of Poison Control Centers National Data Collection
System. Am J Emerg Med 1992;10:452-505.
2. Litovitz TL, Holm KC, Bailey KM, Schmitz BF. 1992 Annual report of the
American Association of Poison Control Centers National Data Collection
System. Am J Emerg Med (in press).
3. National Academy of Sciences. The health effects of nitrate, nitrite and N-
nitroso compounds. Washington, DC: National Academy Press, 1981.
4. Smith RP. Toxic responses of the {*filter*}. In: Klaassen CD, Amdur MO, Doull J,
eds. Casarett and Doull's toxicology. New York: MacMillan Publishing, 1986.
5. Spitalny KC, Brondum J, Vogt RL, Sargent HE, Kappel S. Drinking-water
induced copper intoxication in a Vermont family. Pediatrics 1984;74:1103-6.
6. Pettersson R, Kjellman B. Vomiting and diarrhea are the most common
symptoms in children who drink water with high levels of copper Swedish.
Lakartidningen 1989;86:2361-2.
7. Syracruse Research Corporation, ed. Toxicological profile for copper.
Atlanta: US Department of Health and Human Services, Public Health Service,
Agency for Toxic Substances and Disease Registry, 1990:46.
8. Chugh KS, Singhal PC, Sharma BK. Methemoglobinemia in acute copper sulphate
poisoning Letter. Ann Intern Med 1975;82:226-7.
9. Moore GS, Calabrese EJ. G6PD-deficiency: a potential high-risk group to
copper and chlorite ingestion. J Environ Pathol Toxicol 1980;4:271-9.
* The U.S. Environmental Protection Agency (EPA) maximum contaminant level
(MCL) for nitrate-N in drinking water is 10 mg/L.
Health InfoCom Network News Page 17
Volume 6, Number 8 April 4, 1993
** The EPA MCL for copper in drinking water is 1.3 mg/L.
Health InfoCom Network News Page 18
Volume 6, Number 8 April 4, 1993
Classification of American Indian Race on Birth
and Infant Death Certificates -- California and Montana
=======================================================
SOURCE: MMWR 42(12) DATE: Apr 02, 1993
The accuracy of infant mortality rates and other indices of the health of
populations depends on the consistency of information collected from separate
sources (e.g., birth and death certificates). Inconsistent recording of basic
information such as race and ethnicity has resulted in underestimation of
mortality among minority populations, particularly minority populations other
than blacks (1). This report summarizes studies in California and Montana that
describe and measure the magnitude of differences in the recording of race for
American Indians/Alaskan Natives (AI/ANs) on birth and infant death
certificates. California
Reported infant mortality rates in California often have been based on
information obtained separately from birth and death certificates. To assess
the accuracy of the reported rates for the state's AI/ANs, the California Area
Office of the Indian Health Service (IHS) and the California Department of
Health Services used the state's Birth Cohort File in which birth and death
certificate information of individual infants are linked. Infant mortality
rates for 1984-1988 were calculated from the Birth Cohort File using two
methods for defining infant race: the pre-1989 algorithm from CDC's National
Center for Health Statistics * and the IHS algorithm **. With both methods,
information on parents' race was obtained from birth certificates. Numerators
for infant mortality rates were deaths linked to birth certificates that met
the CDC and IHS definitions for AI/AN births. Denominators were composed of
all AI/AN births meeting the CDC and IHS definitions using birth certificates.
All births and deaths were restricted to infants whose mothers were California
residents at the time of the birth. Record linkage for this population was
100% for this study.
The aggregate number of AI/AN births for 1984-1988 was 28,668 when
applying the IHS definition to unlinked birth certificate files (i.e., when
death certificate data were not linked to birth certificates). In the linked
Birth Cohort Files (which allow an additional year of data collection to
ensure the accuracy of the data) the number of AI/AN births was 27,588 when
tabulated using the CDC definition, and 29,030 using the IHS definition.
In every year, the number of AI/AN infant deaths in California was
greater when calculated from the linked file than from the unlinked death
certificate file, and greatest when the IHS definition was applied. The
unlinked Death Statistics Master File for 1984-1988 included data from a total
of 111 death certificates for AI/AN infants. In comparison, the linked file
included 298 deaths when the CDC definition was used (a difference of 168%)
and 315 when the IHS definition was used (a difference of 184%).
For 1984-1988, the infant mortality rate based on unlinked data was 3.9
deaths per 1000 live births. The rate was 2.8 times greater when calculated
Health InfoCom Network News Page 19
Volume 6, Number 8 April 4, 1993
using the linked files (10.8 and 10.9 per 1000 using the CDC and IHS
definitions, respectively). Both of these rates are comparable to that for
AI/AN infants in all areas served by the IHS (10.9 per 1000 for AI/AN
infants). Montana
In Montana, resident birth and infant death records linked by the state
vital registrar since 1980 were used to estimate errors in the reporting of
race (2,3). All linked birth and infant death records for 1980-1989 were
analyzed; less than 0.1% of these records in the state were not linked. Infant
race at birth was tabulated based on 1) the mother's race -- the procedure
used as of 1989 in statistics tabulated by CDC -- and 2) the algorithm used
before 1989. Infant race at death was tabulated by three methods: 1) race as
reported on the infant's death certificate -- the standard method, used in the
absence of linked records; 2) the mother's race as recorded on the birth
certificate; and 3) the infant's race derived from the CDC algorithm used
before 1989.
The number of AI/AN births based on the CDC definition (14,893) was
higher than that based on mother's race (12,749). For 1980-1989, there were
1285 infant deaths of all races in Montana. For 42 (3.3%) of these, the race
recorded on the death certificate differed from that on the birth certificate.
Of 1036 infants who were classified as white at birth, seven (0.7%) were
classified otherwise at death (one as black and six as AI/AN). In comparison,
of 232 infants classified as AI/AN at birth, 29 (12.5%) were classified as
white at death. The number of AI/AN infant deaths for the decade varied
according to the definition of the decedents' race: 210 were reported on the
death certificate, 232 were ascertained using the pre-1989 CDC algorithm, and
202 were ascertained by assignment of the mother's race.
Determination of the race at death based on the death record produced two
different estimates of AI/AN infant mortality (Table 1): 16.5 per 1000 when
mother's race was used to define race at birth and 14.1 per 1000 when the pre-
1989 CDC algorithm was used. Use of the pre-1989 CDC algorithm to define race
at death produced two higher estimates: 18.2 when the mother's race was used
to define race at birth and 15.6 using the pre-1989 CDC algorithm. Finally,
the rates were lowest when race of the infant at death was defined as the
mother's race: 15.8 when mother's race was used to define race at birth and
13.6 when the infant's race at birth was defined by the pre-1989 CDC
algorithm. When the mother's race determined from the birth record was used as
the denominator, estimated AI/AN infant mortality rates were always higher
than those obtained when the pre-1989 CDC algorithm was used, independent of
the method of assigning race to the infant (Table 1). The infant mortality
rate for the same period calculated by standard procedures with unlinked vital
records was 13.9 per 1000 live births.
Reported by: CC Watson, MPH, California Area Office, Indian Health Svc,
Sacramento. T Bennett, DrPH, Dept of Maternal and Child Health, Univ of North
Carolina at Chapel Hill. FW Reed, PhD, WH McBroom, PhD, Center for Population
Health InfoCom Network News Page 20
Volume 6, Number 8 April 4, 1993
Research, Univ of Montana, Missoula. SD Helgerson, MD, Health Care Financing
Administration (Region X), Seattle. Div of Surveillance and Epidemiology,
Epidemiology Program Office, CDC.
Editorial Note: From 1973 through 1987, the reported U.S. AI/AN infant
mortality rate decreased from 22 to 11 infant deaths per 1000 live births (4).
In addition, during this period cause-specific mortality rates for AI/ANs
decreased 77% for gastrointestinal diseases, 76% for tuberculosis, 59% for
pneumonia and influenza, and 54% for unintentional injuries (4). Although
these trends in mortality rates suggest substantial improvements in public
health status for AI/ANs, the findings in this report indicate that race-
specific mortality rates can substantially underestimate disease burden when
race coding is inconsistent in vital records.
Race has not been consistently defined or ascertained in public health
and related data sets (5). For example, when infant mortality rates are
calculated, the denominator generally consists of the number of births in the
same year as that in which the deaths occurred. For many other mortality and
morbidity rates, denominators are commonly derived from U.S. Census data that
rely on respondents' self-identification. In comparison, race coding for
numerator data may reflect a wider range of methods, including designation of
race by next-of-kin, a coroner, or other person who certifies the death.
Similarly, race coding for reportable diseases may represent the independent
designation of a health-care provider.
The effect of the inconsistent race coding associated with these
practices may be to obscure race-specific disease and injury burdens. The
studies described in this report and in others (1,6) illustrate how the impact
of inconsistent race coding between data sets can be estimated by linking the
data sets. Previous studies linking IHS records to state death files, tumor
and injury registries, and a registry for end-stage renal disease have
documented underascertainment of the disease burden for AI/AN populations (7-
11).
Vital records and registries are essential data sources for monitoring
progress toward national health objectives for the year 2000 (12). The Indian
health amendments of 1992 *** require attention to 61 AI/AN health objectives.
Improved national and local surveillance of AI/AN mortality rates and other
health indicators is necessary for achieving these objectives. Improvement of
surveillance efforts will require assessment of fundamental concepts such as
"race," "ancestry," and "ethnicity" and the meaning of these terms in
different populations. These improvements also will require the use of
consistent categories and measurement procedures between data-collection
sources and agencies.
Different definitions of race identification may serve different
purposes. For example, the IHS definition corresponds to definitions of tribal
membership associated with health-care benefits. The revised definition of an
infant's race at birth as that of its mother (now used at the national level
Health InfoCom Network News Page 21
Volume 6, Number 8 April 4, 1993
for purposes of published statistics) is consistent for all races. Access to
national linked birth and infant death computer files, currently available for
infants born from 1983 through 1987, will facilitate the computation of vital
statistics with consistent definitions of race at birth and death (13).
References
1. Hahn RA, Mulinare J, Teutsch SM. Inconsistencies in coding of race and
ethnicity between birth and death in US infants: a new look at infant
mortality, 1983 through 1985. JAMA 1992;267:259-63.
2. Reed FW, McBroom WH, Sperry SH, Helgerson SD. Implications of inaccurate
infant mortality data. The IHS Primary Care Provider 1992;17:29-32.
3. McBroom WH, Reed FW, Sperry SH, Helgerson SD. Ethnic misclassification,
substance use, and their sequelae. Presented at the fifth Annual Indian Health
Service Research Conference, Tucson, Arizona, May 5-7, 1992.
4. Indian Health Service. Trends in Indian health, 1992. Rockville, Maryland:
US Department of Health and Human Services, Public Health Service, 1992:63.
5. Hahn RA. The state of federal health statistics on racial and ethnic
groups. JAMA 1992;267:268-71.
6. Hahn RA, Mendlein JM, Helgerson SD. Differential classification of American
Indian race on birth and death certificates, U.S. reservation states, 1983-
1985. The IHS Primary Care Provider 1993;18:8-11.
7. Kimball EH, Frost F. Estimate of racial misclassification error on
Washington State death certificates and Puget Sound Cancer Registry. Presented
at fourth Annual Indian Health Service Research Conference, Tucson, Arizona,
April 22-24, 1991.
8. Frost F, Taylor V, Fries E. Race misclassification of Native Americans in a
Surveillance, Epidemiology, and End Results registry. J Natl Cancer Inst
1992;84:957-62.
9. Bleed DM, Risser DR, Sperry S, Hellhake D, Helgerson SD. Cancer incidence
and survival among American Indians registered for Indian Health Service care
in Montana, 1982-1987. J Natl Cancer Inst 1992;84:1500-5.
10. Sugarman JR, Soderberg R, Gordon JE, Rivara FP. Racial misclassification
of American Indians: its effect on injury rates in Oregon, 1989 through 1990.
Am J Public Health (in press).
Health InfoCom Network News Page 22
Volume 6, Number 8 April 4, 1993
11. Sugarman JR, Lawson L. The effect of racial misclassification on estimates
of end-stage renal disease among American Indians and Alaska Natives, Pacific
Northwest, 1988-1990. Am J Kidney Dis 1993;24:383-6.
12. Public Health Service. Healthy people 2000: national health promotion and
disease prevention objectives -- full report, with commentary. Washington, DC:
US Department of Health and Human Services, Public Health Service, 1991; DHHS
publication no. (PHS)91-50212.
13. Prager, K, Flinchum GA, Johnson DP. The NCHS pilot project to link birth
and infant death records: stage 1. Public Health Rep 1987;102:216-23.
* Based on the pre-1989 CDC algorithm in use during the study period, infants
were determined to be AI/AN if on the birth certificate 1) the father was
coded AI/AN and the mother was not coded Hawaiian or 2) if the mother was
coded AI/AN and the father was coded AI/AN, white, or race unknown.
** According to the IHS procedure, infants were coded as AI/AN if the mother
and/or father were identified as AI/AN on the birth certificate.
*** Public Law 102-573.
Health InfoCom Network News Page 23
Volume 6, Number 8 April 4, 1993
Sliding-Associated Injuries in College and
Professional Baseball -- 1990-1991
==========================================
SOURCE: MMWR 42(12) DATE: Apr 02, 1993
Softball and baseball are among the most frequent causes of sports-
related emergency department visits in the United States, accounting for an
estimated 321,000 injuries in 1989 (1). Approximately 71% of softball-related
injuries are caused by sliding (2). The use of breakaway bases substantially
decreases the risk for or occurrence of sliding-related injuries among
recreational softball league players (3). This report summarizes the findings
of a study on the impact of breakaway base use on sliding injuries among
college and professional minor league baseball players (4).
During 1990 and 1991, 19 teams participating in the study used breakaway
bases on their home field and stationary bases during away games for one of
the two seasons. During the first season, the teams comprised one college and
six professional minor league teams; during the second season, seven college
and five minor league teams were added to the study. Base-sliding injuries and
comments about the bases were recorded on a standard form by team physicians,
athletic trainers, managers, or administrative staff for these teams.
During the 2-season period, the teams played an aggregate 498 away games
using stationary bases and 486 home games using breakaway bases. Ten sliding
injuries were recorded (2.0 per 100 games) during away games and two (0.4 per
100 games) during home games (relative risk=4.9; 95% confidence interval=1.2-
19.2). Of the 10 injuries involving stationary bases, seven were ankle sprains
(average participation time missed: 12 days), and three were knee injuries
(one medial collateral ligament sprain and two meniscus tears that required
surgery both were season-ending). Of the two injuries involving breakaway
bases, one was a minor shoulder contusion incurred when the player slid head
first into a base that did not release. The second injury occurred when a
player slid toward the base and sustained an ankle fracture; however, the
player did not make contact with the base.
Surveys of managers and trainers indicated that all teams planned to
continue using breakaway bases. Umpires reported that breakaway bases did not
complicate judgment calls (i.e., "safe" versus "out") when the bases released
(54 2.7% of 2028 total slides on breakaway bases).
Reported by: DH Janda, MD, D Mackesy, MD, Institute for Preventive Sports
Medicine, Ann Arbor, Michigan. R Maguire, Bucknell Univ, Lewisburg,
Pennsylvania. RJ Hawkins, MD, Steadman-Hawkins Clinic, Vail, Colorado. P
Fowler, MD, Univ of Western Ontario, London, Ontario, Canada. J Boyd, MD,
Orthopaedic Consultants, Minneapolis. Epidemiology Br, Div of Injury Control,
National Center for Injury Prevention and Control, CDC.
Editorial Note: The findings in this report suggest that breakaway bases
Health InfoCom Network News Page 24
Volume 6, Number 8 April 4, 1993
decrease the risk and severity of sliding injuries among college and minor
league baseball teams (4). The potential public health impact of increased use
of breakaway bases is important: in the United States, 712 college and 168
minor league teams compete in organized baseball. In addition, an estimated 40
million {*filter*}s participate in organized softball leagues that play
approximately 23 million games per year (3,5).
Most base-sliding injuries result from judgment errors of the runners,
poor sliding technique, poor timing, and/or inadequate physical conditioning
(3). Breakaway bases are a passive intervention that modifies the outcome of
these factors. The quick-release feature of the breakaway bases decreases the
impact load generated against the athlete's limb and subsequent trauma.
Additional studies should assess the usefulness of age-appropriate breakaway
bases in organized baseball and softball for children. Furthermore, such
studies should attempt to address the effect of potential biases (e.g.,
nonblinding with respect to the hypothesis being tested and the need for
uniform definitions of injury). The findings in this report suggest that
breakaway bases should be used at all levels of {*filter*} softball and baseball
play.
References
1. US Consumer Product Safety Commission. National Electronic Injury
Surveillance System: January-December 1989. NEISS Data Highlights 1990;13.
2. Janda DH, Hankin FM, Wojtys EM. Softball injuries: cost, cause and
prevention. Am Fam Physician 1986;33:143-4.
3. Janda DH, Wojtys EM, Hankin FM, et al. Softball sliding injuries: a
prospective study comparing standard and modified bases. JAMA 1988;259:1848-
50.
4. Janda DH, Maguire R, Mackesy D, Hawkins RJ, Fowler P, Boyd J. Sliding
injuries in college and professional baseball: a prospective study comparing
standard and break-away bases. Clinical Journal of Sports Medicine (in press).
5. CDC. Softball sliding injuries -- Michigan, 1986-1987. MMWR 1988;37:169-70.
Health InfoCom Network News Page 25
Volume 6, Number 8 April 4, 1993
Cigarette Smoking Among {*filter*}s -- United States, 1991
=====================================================
SOURCE: MMWR 42(12) DATE: Apr 02, 1993
From 1965 through 1985, smoking prevalence in the United States declined
at a rate of 0.5 percentage points per year (1), and from 1987 through 1990,
the rate of decline accelerated to 1.1 percentage points per year (2). CDC
monitors the use of tobacco in the United States to evaluate progress in
reducing smoking prevalence. To determine the prevalence of smoking among U.S.
{*filter*}s during 1991, the National Health Interview Survey-Health Promotion and
Disease Prevention (NHIS-HPDP) supplement collected self-reported information
on cigarette smoking from a representative sample of the U.S. civilian,
noninstitutionalized population aged greater than or equal to 18 years. This
report summarizes the results of this survey.
The overall response rate for the 1991 NHIS-HPDP was 87.8%. Participants
(n=43,732) were asked: "Have you smoked at least 100 cigarettes in your entire
life?" and "Do you smoke cigarettes now?" Current smokers were defined as
those who reported smoking at least 100 cigarettes and who were currently
smoking and former smokers as those who reported having smoked at least 100
cigarettes and who were not smoking now. Ever smokers included current and
former smokers. Current smokers were then asked: "Do you now smoke cigarettes
every day or some days?" Respondents reporting they smoked every day were
asked: "On the average, how many cigarettes do you now smoke a day?" Data were
adjusted for nonresponse and weighted to provide national estimates.
Confidence intervals (CIs) were calculated using standard errors generated by
the Software for Survey Data Analysis (SUDAAN) (3).
In 1991, an estimated 89.8 million (49.8%) {*filter*}s in the United States
were ever smokers, and 46.3 million (25.7%) were current smokers.
Approximately 43.5 million persons (48.5% of all ever smokers 95% CI=47.7%-
49.3%) were former smokers during 1991. The proportion of former smokers
among ever smokers was higher among men (51.6% 95% CI=50.4%-52.7%) than
among women (44.7% 95% CI=43.6%-45.8%) and increased with increased
education from 41.8% (95% CI=40.1%-43.6%) for those with less than 12 years of
education to 66.1% (95% CI=64.3%-67.9%) for those with greater than or equal
to 16 years of education.
Among men, 24.0 million (28.1%) were current smokers; among women, 22.2
million (23.5%) were current smokers (Table 1). The prevalence of smoking was
higher among men than among women for most sociodemographic groups (Table 1).
Smoking was most prevalent among persons aged 25-44 years. The prevalence of
smoking was highest among American Indians/Alaskan Natives and blacks, and
lowest among Asians/Pacific Islanders. Differences between black and white
{*filter*}s were mainly among men. The prevalence of smoking was lower among
Hispanics than non-Hispanics, reflecting the lower prevalence of smoking among
Hispanic women. Cigarette smoking prevalence decreased with increasing
education, and was higher among persons who lived below the poverty level *
Health InfoCom Network News Page 26
Volume 6, Number 8 April 4, 1993
(Table 1).
In 1991, the mean number of cigarettes smoked daily per smoker was 20.0
(95% CI=19.7-20.3). The mean was substantially higher for men (21.6 95%
CI=21.2-22.0) than women (18.3 95% CI=18.0-18.6), for whites (21.0 95%
CI=20.7-21.3) than blacks (15.0 95% CI=14.4-15.6), for non-Hispanics (20.4
95% CI=20.1-20.7) than Hispanics (13.4 95% CI=12.5-14.3), and for persons
at or above the poverty level (20.3 95% CI=20.0-20.6) than persons below the
poverty level (18.7 95% CI=18.1-19.3).
Reported by: Epidemiology Br, Office on Smoking and Health, National Center
for Chronic Disease Prevention and Health Promotion; Div of Health Interview
Statistics, National Center for Health Statistics, CDC.
Editorial Note: The findings in this report indicate that the estimate of
smoking prevalence in 1991 was the same as in 1990 (2). These findings are
consistent with national household surveys on drug abuse (4-6), and public
polls (7) that reveal similar patterns of declining prevalence until 1990
followed by a leveling during 1991. Among blacks and women, the prevalence of
current smoking during 1991 was slightly higher than during 1990 (2). Factors
that contributed to the leveling in smoking prevalence may include the steady
growth in market share of discount cigarettes (8) and the recent 10.4% annual
increase to an estimated $3.9 million in domestic cigarette advertising and
promotional expenditures (9).
Differences in prevalence among racial and ethnic groups may be
influenced by differences in educational levels and socioeconomic status, as
well as social and cultural phenomena that require further explanation. For
example, targeted marketing practices may play a role in maintaining or
increasing prevalence among some groups, and affecting the differential
initiation of smoking by young people (1). The national health objectives for
the year 2000 have established special population target groups for the
reduction of smoking prevalence including blacks, Hispanics, American
Indians/Alaskan Natives, and Southeast Asian men (10).
Acceleration of the decline in smoking prevalence will require
intensified efforts to discourage the use of tobacco by helping smokers break
the {*filter*}ion to nicotine, persuading children never to start smoking, and
enacting public policies that discourage smoking. Such policies include
increasing taxes on tobacco products, enforcing minors'-access laws,
restricting smoking in public places, and restricting tobacco advertising and
promotion (1).
References
1. CDC. Reducing the health consequences of smoking: 25 years of progress -- a
report of the Surgeon General. Rockville, Maryland: US Department of Health
and Human Services, Public Health Service, 1989; DHHS publication no. (CDC)89-
Health InfoCom Network News Page 27
Volume 6, Number 8 April 4, 1993
8411.
2. CDC. Cigarette smoking among {*filter*}s -- United States, 1990. MMWR
1992;41:354-5,361-2.
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HICN608 Medical News Part 2/4