<!DOCTYPE HTML PUBLIC "-//W3C//DTD HTML 4.0 Transitional//EN">
<!-- saved from url=(0065)http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html -->
<HTML><HEAD><TITLE>Association between Ozone and Asthma Emergency Department Visits in Saint John, New Brunswick, Canada</TITLE>
<META http-equiv=Content-Type content="text/html; charset=iso-8859-1">
<META content=document name=resource-type>
<META content=global name=Distribution>
<META 
content="David M. Stieb, Richard T. Burnett, Robert C. Beveridge, Jeffrey R. Brook" 
name=Author>
<META 
content="Association between Ozone and Asthma Emergency Department Visits in Saint John, New Brunswick, Canada" 
name=Description>
<META content="air pollution, asthma, emergency department, ozone" 
name=KeyWords>
<META content="MSHTML 6.00.2900.3086" name=GENERATOR></HEAD>
<BODY text=#000000 bgColor=#ffffff>  <BR><FONT 
face="Tekton, Times, Geneva, Arial"><FONT size=+0>Environmental Health 
Perspectives, Volume 104, Number 12, December 1996</FONT></FONT> 
<P>[<A 
href="http://v2626umcay001/cgi-bin/pubmedf.pl?linktype=citation&pmid=9118879">Citation 
in PubMed</A>] [<A 
href="http://v2626umcay001/cgi-bin/pubmedf.pl?linktype=related&pmid=9118879">Related 
Articles</A>] 
<H2>Association between Ozone and Asthma Emergency Department Visits in Saint 
John, New Brunswick, Canada</H2><B>David M. Stieb,<SUP>1</SUP> Richard T. 
Burnett,<SUP>2</SUP> Robert C. Beveridge,<SUP>3</SUP> and Jeffrey R. 
Brook<SUP>4</SUP></B> 
<P><SUP>1</SUP>Air Quality Health Effects Research Section, Health Canada; 
<SUP>2</SUP>Biostatistics Division, Environmental Health Directorate, Health 
Canada, Ottawa, Ontario, Canada; <SUP>3</SUP>Department of Emergency Medicine, 
Region 2 Hospital Corporation, Saint John, New Brunswick, Canada; 
<SUP>4</SUP>Atmospheric Environment Service, Environment Canada, Downsview, 
Ontario, Canada 
<UL>
  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#abstract">Abstract</A> 

  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#intro">Introduction</A> 

  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#methods">Methods</A> 

  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#results">Results</A> 

  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#discussion">Discussion</A> 

  <LI><A 
  href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/082714.html#conclusions">Conclusions</A> 
  </LI></UL>
<HR>

<P><A name=abstract></A><B>Abstract</B> 
<P>This study examines the relationship of asthma emergency department (ED) 
visits to daily concentrations of ozone and other air pollutants in Saint John, 
New Brunswick, Canada. Data on ED visits with a presenting complaint of asthma 
(n = 1987) were abstracted for the period 1984-1992 (May-September). Air 
pollution variables included ozone, sulfur dioxide, nitrogen dioxide, sulfate, 
and total suspended particulate (TSP); weather variables included temperature, 
humidex, dewpoint, and relative humidity. Daily ED visit frequencies were 
filtered to remove day of the week and long wave trends, and filtered values 
were regressed on air pollution and weather variables for the same day and the 3 
previous days. The mean daily 1-hr maximum ozone concentration during the study 
period was 41.6 ppb. A positive, statistically significant (p<0.05) 
association was observed between ozone and asthma ED visits 2 days later, and 
the strength of the association was greater in nonlinear models. The frequency 
of asthma ED visits was 33% higher (95% CI, 10-56%) when the daily 1-hr maximum 
ozone concentration exceeded 75 ppb (the 95<SUP>th</SUP> percentile). The ozone 
effect was not significantly influenced by the addition of weather or other 
pollutant variables into the model or by the exclusion of repeat ED visits. 
However, given the limited number of sampling days for sulfate and TSP, a 
particulate effect could not be ruled out. We detected a significant association 
between ozone and asthma ED visits, despite the vast majority of sampling days 
being below current U.S. and Canadian standards. <B>Key words</B>: air 
pollution, asthma, emergency department, ozone. <B>Environ Health Perspect</B> 
104:1354-1360 (1996) 
<P>
<HR>

<P>Address correspondence to D.M. Stieb, Air Quality Health Effects Research 
Section, Health Canada Building 8, (postal locator 0803C), Tunney's Pasture, 
Ottawa, ON K1A 0L2. 
<P>This study was supported by Health Canada (contract #H4078-4-C472/01-SS) and 
by the Environmental Trust Fund, Government of New Brunswick. The authors would 
like to thank Marc Smith-Doiron and Serge Beaulieu for data management and 
analysis and the Ministry of Environment of New Brunswick for providing air 
quality data. 
<P>Received 7 March 1996; accepted 16 August 1996. 
<P>
<HR>

<P><A name=intro></A>
<H3>Introduction</H3>Recent evidence has implicated ground-level ozone as a 
contributor to both mortality (1) and morbidity, including hospital admissions 
(2-4), emergency department visits (5-9), symptoms (10,11), pulmonary function 
changes (11,12), and inflammatory changes in the respiratory tract (13,14). Data 
on emergency department (ED) visits have been used to examine the effects of a 
variety of air pollutants including particulates, ozone, sulfur dioxide, 
nitrogen dioxide, sulfate, and hydrogen sulfide (5-9,15-21). As health 
endpoints, ED visits have the advantages of reflecting an adverse health event 
of clear clinical significance, which at the same time is more frequent in 
occurrence than death or hospital admission. Conversely, abstracted ED visit 
data in a form amenable to analysis together with environmental exposure data 
are not widely available. In this study, we employed ED visit data that had 
previously been abstracted from clinical records for administrative and 
quality-of-care assessment purposes. 
<P>The primary objective of this study was to further examine the relationship 
between daily ozone concentrations and emergency department visits for asthma, 
including analysis of lag periods ranging from 0 to 3 days, assessment of the 
impact of other pollution and weather variables on the ozone effect, and 
examination of differences in effects between children and adults. Unique 
features of our analysis include an examination of the shape of the 
dose-response function and assessment of the impact of repeat ED visits on the 
strength of the association. 
<H3><A name=methods></A>Methods</H3>This study was carried out in Saint John, 
New Brunswick, a city of approximately 75,000 people (surrounding metropolitan 
area with an additional 50,000) on Canada's Atlantic coast. Local air pollution 
sources include a large petroleum refinery, two oil-fired generating stations, 
and two pulp mills. The area is also subject to long range transport of air 
pollutants from eastern Canada and the eastern United States. The city has two 
hospital emergency departments, with a combined annual volume of approximately 
90,000 visits. Access to emergency services is universal under Canada's publicly 
provided health care system. 
<P>Data on ED visits were obtained from the Saint John Regional Hospital's ED 
visit database, which consists of electronic records created by the registration 
clerk at the time each patient enters the emergency department. Data were 
obtained for the period 1984-1992 (May-September only) for visits with a 
presenting complaint of asthma. At the time of registration, the patient 
provided a presenting complaint to the registration clerk, who then coded the 
complaint from a predefined menu, which included a category (other) for 
complaints that did not clearly fit the menu. In addition to the presenting 
complaint, extracted information included patient's age and sex and the date of 
visit. Original ED charts for all visits made between May and September 1987 
were also reviewed manually by research nurses, and data were abstracted for 
both presenting complaint and discharge diagnosis. This permitted an analysis of 
the relationship between presenting complaint and discharge diagnosis, as well 
as agreement between the database and manual reviewers on coding of presenting 
complaint. 
<P>Air pollution data [ozone (O<SUB>3</SUB>), sulfur dioxide (SO<SUB>2</SUB>), 
nitrogen dioxide (NO<SUB>2</SUB>), sulfate (SO<SUB>4</SUB><SUP>2-</SUP>), total 
suspended particulate (TSP)] and meteorological data (temperature, dewpoint, 
relative humidity, and humidex) were obtained from Environment Canada and the 
New Brunswick Ministry of Environment. Where data were available from more than 
one monitoring station, mean values were used. Ozone was measured continuously 
using instruments based on the UV absorption method. The instrument types 
employed were either a Monitor Labs (Model 8810; Englewood, CO) or a Dasibi 
(Model 1003; Glendale, CA). SO<SUB>2</SUB> was measured continuously using the 
pulsed fluorescence detection method (Monitor Labs-Model 8850) and 
NO<SUB>2</SUB> was measured continuously using the chemiluminescence detection 
method (Monitor Labs Model 8840). TSP samples were collected over 24-hour 
periods using standard high-volume samplers and Teflon-coated glass fibre 
filters (Pallflex, Putnam, CT). Particulate sulfate on the high volume samples 
was determined using ion chromatography. Daily O<SUB>3</SUB>, NO<SUB>2</SUB>, 
and SO<SUB>2</SUB> data were utilized, while only every sixth day 
SO<SUB>4</SUB><SUP>2-</SUP> and TSP data were available. No inhalable or 
respirable particulate data were available. Humidex is an index in which a 
certain number of degrees Celsius are added to the dry bulb temperature to 
account for the additional discomfort associated with excessive humidity. 
<P>Analysis was conducted on an IBM RS-6000 mini computer (IBM, Armonk, NY) 
using UNIX-based SAS statistical software (SAS, Cary, NC). Daily frequencies of 
asthma visits were filtered to remove day of the week and long wave (e.g. 
seasonal) trends, which might otherwise confound the relationship with air 
pollution. Daily frequencies were related to air pollution and weather variables 
by the regression equation: 
<CENTER>
<P><IMG height=41 src="Stieb et al 1996_files/stiebe(yt).gif" width=263 
NOSAVE></CENTER>
<P>where E(y<SUB>t</SUB>) is the expected number of ED visits on the 
t<SUP>th</SUP> day of sampling; y<SUB>t</SUB> is the actual number of ED visits 
on the t<SUP>th</SUP> day of sampling; D<SUB>t</SUB> is the average number of ED 
visits by day of the week (seven distinct values); 
<CENTER>
<P><IMG height=45 src="Stieb et al 1996_files/stiebc.gif" width=104 
NOSAVE></CENTER>
<P>is a normalizing constant; 
<CENTER>
<P><IMG height=39 src="Stieb et al 1996_files/stiebst.gif" width=89 
NOSAVE></CENTER>
<P>is a 19 day symmetric linear filter, where <IMG height=11 
src="Stieb et al 1996_files/Phi.gif" width=13 NOSAVE><SUB>0</SUB> ... <IMG 
height=11 src="Stieb et al 1996_files/Phi.gif" width=13 NOSAVE><SUB>9</SUB> are 
unique weights given by 0.0874, 0.0857, 0.0807, 0.0729, 0.0629, 0.0518, 0.0404, 
0.0296, 0.0200, and 0.0123 with <IMG height=11 
src="Stieb et al 1996_files/Phi.gif" width=13 NOSAVE><SUB>l</SUB> = <IMG 
height=11 src="Stieb et al 1996_files/Phi.gif" width=13 NOSAVE><SUB>-l</SUB> 
(22); <IMG height=11 src="Stieb et al 1996_files/alpha.gif" width=11 
NOSAVE> and ß are the regression intercept and coefficient, respectively; and 
x<SUB>t</SUB> is the value of the environmental variable on the t<SUP>th 
</SUP>day. Because the daily frequency of visits was small, the variance was 
assumed to be proportional to the expected response (i.e., V(y<SUB>t</SUB>) 
= <IMG height=11 src="Stieb et al 1996_files/theta.gif" width=10 
NOSAVE>E(y<SUB>t</SUB>), where <IMG height=11 
src="Stieb et al 1996_files/theta.gif" width=10 NOSAVE> is a proportionality 
constant). The relationship between pollutant/meteorological variables, lagged 
0-3 days, and ED visit frequency was assessed using the SAS NLIN procedure (23), 
setting the regression weights equal to the inverse of the expected response. 
This is equivalent to a generalized estimating equation approach and is 
identical to Poisson regression when the variance is equal to the expected 
response (i.e., <IMG height=11 src="Stieb et al 1996_files/theta.gif" 
width=10 NOSAVE> = 1 above) (24). This procedure is more general than Poisson 
regression and allows for Poisson over and under dispersion (i.e., <IMG 
height=11 src="Stieb et al 1996_files/theta.gif" width=10 NOSAVE> >1 
or <IMG height=11 src="Stieb et al 1996_files/theta.gif" width=10 NOSAVE> 
<1) (24). Because the daily frequency of asthma ED visits may be serially 
autocorrelated (i.e., the frequency of visits on day t + 1 may be related to the 
frequency of visits on day t due, for example, to repeat visits by the same 
patient in relation to the same asthma episode or to multiday pollution 
episodes), first order autocorrelation was assessed using the autocorrelation 
coefficient and Durbin-Watson d statistic. These were derived from the SAS REG 
procedure, setting the regression weights equal to 
[(S<SUB>t</SUB>D<SUB>t</SUB>/C) (a + bx<SUB>t</SUB>)]<SUP>-1</SUP>, where a and 
b are, respectively, estimates of the regression intercept, alpha, and 
coefficient, ß, estimated based on the NLIN procedure. Failure to control for 
autocorrelation could result in biased estimates of the statistical significance 
of air pollution effects. Positive autocorrelation would result in 
underestimates of the standard errors of regression coefficients, while negative 
autocorrelation would result in overestimates (25). 
<H3><A name=results></A>Results</H3>A descriptive summary of air pollution and 
meteorological data is presented in Table 1, together with Pearson correlation 
coefficients. There was generally weak correlation among air pollution 
variables, although SO<SUB>4</SUB><SUP>2-</SUP> and TSP were moderately 
correlated, and as expected, maximum temperature and maximum dewpoint 
temperature were highly correlated with maximum humidex. Canada's National 
Ambient Air Quality Objective for ozone (80 ppb) was exceeded on 3.7% of study 
days, while the U.S. standard (120 ppb) was exceeded on 0.4% of days. Pearson 
correlations between daily 1-hr maximum and daily average concentrations were 
high for ozone, SO<SUB>2</SUB>, and NO<SUB>2</SUB> (0.9; p = 0.0001). 
<P><IMG height=244 src="Stieb et al 1996_files/stiebtable1.gif" width=722 
align=absBottom NOSAVE> <BR>During the study period, 1,163 individuals made a 
total of 1987 ED visits with a presenting complaint of asthma. Forty-seven 
percent of asthma ED visitors were male and 49% were 15 years of age or younger. 
Respectively, 1.5, 6.5, and 10.2% of asthma ED visitors (including admitted and 
discharged patients) made a return visit within 24 and 72 hr and 14 days. For 
the period May-September 1987, presenting complaints coded by research nurses 
based on a manual review of ED charts agreed with those contained in the ED 
database 97% of the time. Ninety-three percent of visits with a presenting 
complaint of asthma were assigned a discharge diagnosis of asthma (the remaining 
7% received discharge diagnoses of upper respiratory infection, pneumonia, 
bronchitis, chronic obstructive pulmonary disease, croup, and 
noncardiorespiratory conditions). Conversely, visits with a presenting complaint 
of asthma accounted for 43% of all visits with a discharge diagnosis of asthma. 
<P>The average number of visits per day with a presenting complaint of asthma 
(henceforth referred to as asthma visits) was 1.5 (minimum of 0 and maximum of 
8). There were no asthma visits on 27% of days, one visit on 30% of days, two 
visits on 24% of days, and more than two visits on 19% of days. Significant 
variability was observed in visit frequency by day of the week, ranging from a 
mean of 1.3 visits per day on Wednesday and Friday to 1.7 visits per day on 
Sunday. Variability was also observed by month, with the greatest number of 
visits in May and September, relative to the other months. The impact of 
filtering on temporal variability in visit frequency is revealed in Figure 1, 
which plots data for 1990, the year during which the greatest seasonal 
variability was observed. In Figure 1A, the value of the filter (S<SUB>t</SUB>), 
which smoothes daily oscillations in visit frequency, is overlaid on the 
unfiltered visit frequency data (y<SUB>t</SUB>). Figure 1B plots the filtered 
data, F<SUB>t</SUB> = y<SUB>t</SUB>/(S<SUB>t</SUB>D<SUB>t</SUB>/C), which 
reflect the removal of seasonal trends from the raw data. 
<P><A 
href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/stiebfig1a.GIF"><IMG 
height=388 alt="Figure 1A.GIF" src="Stieb et al 1996_files/stiebfig1a.gif" 
width=654 NOSAVE></A><A 
href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/stiebfig1b.GIF"><IMG 
height=388 alt="Figure 1B.GIF" src="Stieb et al 1996_files/stiebfig1b.gif" 
width=654 NOSAVE></A> 
<P><FONT size=-1><B>Figure 1.</B> (A) Unfiltered asthma visit frequency, 
y<SUB>t</SUB> and filter, S<SUB>t</SUB>; (B) filtered asthma visit frequency, 
F<SUB>t</SUB>, May-September 1990.</FONT> 
<P>Scatter plots were produced of the relationship of pollutant and meteorologic 
variables to filtered asthma ED visits in order to screen for nonlinearity. Only 
ozone appeared to have a nonlinear relationship with visit rates. Based on this 
screening analysis, for daily average and daily 1-hr maximum ozone 
concentrations, ED visit rates were regressed on an indicator variable 
representing days above and below the 95<SUP>th</SUP> percentile concentration, 
as well as linear, quadratic, and linear-quadratic ozone terms. Visit rates were 
regressed linearly on other individual pollutant and meteorologic variables. 
<P><IMG height=276 src="Stieb et al 1996_files/stiebtable2.gif" width=473 
align=absBottom NOSAVE> <BR>Of all pollutants considered, only ozone exhibited a 
consistently positive association with asthma visit rates 2 days later, which 
was statistically significant (p<0.05) or borderline significant in all model 
forms (see Table 2). Compared to the linear model, the nonlinear models revealed 
stronger associations between both daily average and maximum ozone and asthma ED 
visits, based on the model p-value. A plot of data collapsed into 
<30<SUP>th</SUP> percentile, 30-60<SUP>th</SUP> percentile, 
60<SUP>th</SUP>-95<SUP>th</SUP> percentile, and >95<SUP>th</SUP> percentile 
for daily 1-hr maximum ozone concentration reveals the apparent nonlinearity of 
the ozone effect (Figure 2) (a similar trend was observed for daily average 
concentration). The first order autocorrelation coefficient for the filtered 
daily visits series for both daily average and 1-hr maximum models was low and 
negative (-0.142 and -0.137, respectively), and the Durbin-Watson d statistic 
was close to 2 (2.284 and 2.274, respectively), indicating that there was no 
important autocorrelation in the filtered visit data. The frequency of asthma ED 
visits (filtered as above) was 33% higher (95% CI, 10-56%) when the daily 1-hr 
maximum exceeded 75 ppb (the 95<SUP>th</SUP> percentile). Ozone was not 
significantly associated with asthma visits for other lags (p>0.05). 
<P><A 
href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/stiebfig2.GIF"><IMG 
height=206 alt="Figure 2.GIF" src="Stieb et al 1996_files/stiebfig2.gif" 
width=334 align=absBottom NOSAVE></A> 
<P><FONT size=-1><B>Figure 2.</B> Filtered asthma emergency department (ED) 
visit frequency (F<SUB>t</SUB>) versus daily 1-hr maximum ozone concentration 
(percentiles <30, 30-60, 60-, >95).</FONT> 
<P><IMG height=223 src="Stieb et al 1996_files/stiebtable3.gif" width=473 
align=absBottom NOSAVE> <BR>To assess whether the ozone effect (lag 2 days) was 
confounded by other pollution and weather variables, days above and below the 
95<SUP>th </SUP>percentile for daily 1-hr maximum ozone concentration (lag 2 
days) were compared with respect to the other pollution and weather variables. 
As shown in Table 3, average SO<SUB>2</SUB> (lag 0 days) was lower on high ozone 
days, while maximum NO<SUB>2</SUB> (lag 2 days), SO<SUB>4</SUB><SUP>2-</SUP> 
(lag 3 days), TSP (lag 3 days), maximum temperature (lag 1 day), maximum humidex 
(lag 1 day), maximum dewpoint temperature (lag 2 days), and mean relative 
humidity (lag 1 day) were all higher on high ozone days. (The lag periods for 
the pollution and weather variables shown in this table were those for which the 
strongest association with asthma visits was observed in single variable 
regressions.) These variables were then added to the model for daily 1-hr 
maximum ozone indicator (lag 2 days) (the effects of TSP and 
SO<SUB>4</SUB><SUP>2-</SUP> were examined separately from the other pollution 
and weather variables because of the smaller sample size resulting from every 
sixth day sampling frequency). As seen in Table 4, the ozone coefficient 
increased slightly in both models and remained statistically significant in 
model 1, which had a larger sample size based on daily data. Of the other 
variables, only the effect of mean relative humidity (lag 1 day) was 
statistically significant. 
<P><IMG height=221 src="Stieb et al 1996_files/stiebtable4.gif" width=473 
align=absBottom NOSAVE> <BR>The appropriateness of the 95<SUP>th</SUP> 
percentile as a cutpoint in the ozone models was examined by removing days with 
the top 5% of ozone concentrations and rerunning the regressions. Ozone no 
longer exhibited a statistically significant association with ED visits in this 
subset of the data [e.g., for the daily 1-hr maximum ozone linear model (lag 2 
days), ß = -0.0013 visits/ppb and p = 0.5758]. 
<P><A 
href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/stiebfig3.GIF"><IMG 
height=208 alt="Figure 3.GIF" src="Stieb et al 1996_files/stiebfig3.gif" 
width=334 align=absBottom NOSAVE></A> 
<P><FONT size=-1><B>Figure 3.</B> Filtered asthma emergency department (ED) 
visit frequency (F<SUB>t</SUB>) versus daily 1-hr maximum ozone concentration, 
by age group (percentiles <30, 30-60, 60-95, >95).</FONT> 
<P>Differences in ozone effect for adult (>15 years of age) and childhood 
(</=15 years of age) asthma visits were also examined. As shown in Figure 3, 
the rate of ED visits was slightly higher for adults than children, as was the 
proportional increase in visits when daily 1-hr maximum ozone concentration was 
above the 95<SUP>th</SUP> percentile (47% and 15%, respectively). In regression 
models, the association of ED visits with daily 1-hr maximum, as well as daily 
average ozone concentration, was statistically significant for adults but not 
for children (see Table 5). 
<P><IMG height=171 src="Stieb et al 1996_files/stiebtable5.gif" width=224 
align=absBottom NOSAVE> <BR>Specificity of the ozone effect was assessed by 
examining the relationship between ozone and visits with respiratory presenting 
complaints other than asthma (cough, congestion, wheeze, shortness of breath, or 
difficulty breathing). This constituted a heterogeneous group of 8238 visits, a 
significant proportion of which were assigned a nonrespiratory discharge 
diagnosis (30% in the case of shortness of breath and difficulty breathing). 
None of the pollutants exhibited a statistically significant association with 
this group of visits (p>0.05). 
<P>Although we did not detect important first order autocorrelation in the data, 
we did note (see above) that a small but nontrivial proportion of individuals 
making asthma visits made repeat visits within 14 days. Because this might 
spuriously increase the apparent association between asthma visits and ozone, we 
reran the NLIN procedure, excluding repeat visits within 24 and 72 hr and 14 
days of an earlier visit. As seen in Table 6, coefficients were reduced 
slightly, but remained statistically significant. The 95% confidence intervals 
on these coefficients overlapped those derived from analyses that included 
repeat visits. The largest reduction in coefficient occurred with the exclusion 
of repeat visits within 72 hr and 14 days, which reduced the coefficient for 
daily 1-hr maximum ozone by 13%. 
<H3><A name=discussion></A>Discussion</H3>We have found that asthma ED visits 
increased by 33% when the daily 1-hr maximum ozone concentration exceeded 75 
ppb. This is consistent with findings in New Jersey where asthma ED visits 
increased by 26% when the daytime mean ozone concentration (10 a.m. to 3 p.m.) 
exceeded 60 ppb (5), in Atlanta, Georgia, where pediatric asthma ED visits 
increased by 37% following days when the daily 1-hr maximum ozone concentration 
exceeded 110 ppb (7), and in Mexico City where pediatric asthma visits increased 
by 68% following 2 days on which the 1-hr maximum ozone concentration exceeded 
110 ppb (8). In other studies, however, ozone has not exhibited a significant 
association with respiratory or asthma visits (15-19). While there are a variety 
of factors that may account for these differences, in Vancouver (15) and 
Barcelona (16,18), ozone levels were relatively low compared to other studies in 
which significant ozone effects were observed. 
<P>Effects of daily average and daily 1-hr maximum ozone concentration were 
similar in magnitude and statistical significance in our study. It has been 
suggested that a measure of maximum cumulative exposure such as the daily 
maximum 8-hr moving average may be more relevant than daily average measures in 
terms of cumulative exposure (26). White et al. (7) found a high correlation (r 
= 0.95) between daily 1-hr maximum and daily maximum 8-hr moving average ozone 
concentrations, as well as apparently similar effects on asthma visits. Romieu 
et al. (8) also found a high correlation (r = 0.91) between these two ozone 
metrics, and on this basis only examined the effect of daily 1-hr maximum 
concentration on ED visits. We found a high correlation of average ozone 
concentration between 8 a.m. and 8 p.m. with both daily 1-hr maximum (r = 0.85) 
and daily average (r = 0.94) concentrations. Results of regressions relating 
average ozone concentration between 8 a.m. and 8 p.m. with asthma visits were 
consistent with results for the other metrics. The lack of a clearly stronger 
predictor of asthma visits among the three metrics probably reflects the high 
degree of collinearity. 
<P>In our study, a 2-day lag was observed between elevated ozone concentrations 
and increased asthma ED visits. In ED studies in New Jersey and Baton Rouge, 
Louisiana (5,6,9), the strongest effects were for same day ozone concentration 
(i.e., lag 0 days), while in the Mexico City study, the best fitting model 
included ozone concentration (lag 1 day) (8). In a study of air pollution and 
hospital admissions in Ontario (3), the largest ozone effects for all 
respiratory admissions were for lags of 1 and 2 days, while in a similar study 
in New York State (4), the largest effects differed by city and diagnosis, with 
the peak asthma admission effect at 3 days and 1 day in Buffalo and New York 
City, respectively. We were not able to detect an ozone effect for multiple lag 
periods, in contrast to other studies (3-6,8). It is not clear why various 
studies differ with respect to the observed lag between ozone concentrations and 
asthma ED visits or hospital admissions, although possible factors include both 
methodological differences and differences in study populations, exposure 
characteristics, and outcomes studied (e.g., hospital admissions vs. ED visits). 

<P>To better understand the lag between exposure and effect in our study 
population, we conducted preliminary analyses of 3 months of enhanced ED data 
(July-September 1994) being collected in a subsequent phase of this study, in 
which ED visitors were interviewed in detail both at the time of their visit and 
in follow-up 2 weeks later. These analyses indicated that for asthma patients 
the median number of days between symptom onset and ED visit was 2.0 (mean of 
4.4) (27). Although this appears to be slightly longer than in some reports 
(28-31), it is consistent with one other study among children in Toronto (32) 
and corresponds to the 2 day lag effect we observed for ozone. While this is not 
sufficient evidence to infer a causal association, it does suggest that on 
average, the timing of the ozone effect is consistent with the temporal pattern 
of asthma exacerbations in this population. 
<P>In our study, quadratic, linear-quadratic, and indicator models consistently 
fit the data better than the linear model, suggesting that ozone effects are 
reduced or absent below a certain concentration. This is consistent with the 
finding of White et al. (7) that there appeared to be no effect of ozone on 
pediatric asthma visits when ozone concentrations were below 110 ppb. This is 
contrary to the findings in the Ontario hospital admission study in which there 
appeared to be no concentration at which ozone effects could not be detected 
(3). Again, interstudy differences could reflect a variety of factors. Emergency 
department visitors may, for example, be less sensitive to lower levels of air 
pollution than patients admitted to hospitals. 
<P>We did not detect significant effects of co-pollutants, either on their own 
or in terms of their impact on the ozone effect. Given the limited number of 
sampling days for sulfates and TSP, however, we cannot rule out an association 
between particles and asthma ED visits. The effects of these variables in other 
ED visit studies is inconsistent. As noted earlier, in some ED visit studies, 
other pollutants have had significant effects while ozone did not. In one study 
in Southern California (20), a significant ozone effect was noted in the region 
with the highest ozone levels, while sulfate effects were dominant in other 
regions. In Saint John, acid aerosols are a potentially important unmeasured 
co-pollutant in this analysis (33), and daily measurement of particle strong 
acidity is now under way to be used in future analyses with respect to ED 
visits. With respect to weather variables, temperature appears to be an 
inconsistent explanatory variable in ED visit studies. In one study, it was 
positively associated with ED visits for various respiratory conditions (20); in 
another study, a positive association was observed in winter and a negative 
association in summer (21), while in other studies, temperature was negatively 
associated (5,6,16,18,19). 
<P>We observed positive associations between ozone concentration and asthma ED 
visits for both adult and childhood asthma, although the subgroup analysis for 
children was not statistically significant. Burnett et al. (3) found that the 
age group with the largest proportion of asthma hospital admissions attributable 
to ozone was 0-1 year of age; however, asthma remains an unclear diagnosis in 
infants (34). In those above 2 years of age, the largest ozone effects were in 
those above 35 years of age (3), which is more consistent with our results. Age 
alone may be of limited usefulness as a subgrouping variable for asthmatics. 
Baseline severity, premorbid asthma management, and other coincident exposures 
(e.g. viral infections, allergens, tobacco smoke) may be more clinically 
relevant. These variables are currently being measured in a subsequent phase of 
this study. 
<P>We observed that a small proportion of ED visits were followed by repeat 
visits within 24 and 72 hr and 14 days. While relapse has not been consistently 
defined or applied in studies of asthma ED visits, our observations were 
consistent with comparably defined relapse rates reported in other studies 
(31,35,36). The strength of the association between ozone and asthma ED visits 
was reduced slightly, but remained statistically significant when we excluded 
repeat visits, the largest reduction in the coefficient (13%) being of marginal 
practical importance. To our knowledge, this issue has not been addressed in 
other time-series analyses of ED visits and air pollution, although in their 
study, White et al. (7) excluded repeat visits within 24 hr at the time of data 
collection. While repeat visits do not appear to have influenced our results, 
further examination of their influence in time-series studies would be of 
interest, particularly where relapse rates are higher than observed here. 
<P>Our results appeared to be specific for asthma visits. We were unable to 
demonstrate a relationship between ozone and ED visits for 
respiratory-presenting complaints other than asthma. This probably reflects the 
heterogeneity of this group of visits, in terms of their eventual discharge 
diagnosis. Although the vast majority of ED visits with a presenting complaint 
of asthma did in fact represent asthma exacerbations, having a presenting 
complaint of asthma was a relatively insensitive indicator of the presence of 
asthma as assessed by a physician, accounting for less than half of visits with 
a discharge diagnosis of asthma. It is unclear to what extent and in what 
direction this might bias our estimate of the effect of ozone on asthma. At this 
point we do not know how asthmatics who report asthma as their presenting 
complaint differ from those with some other presenting complaint such as 
shortness of breath or wheezing. In particular, it is unclear whether the former 
group of patients would tend to have more severe asthma and thus be more 
sensitive to air pollution and other triggers or alternatively better informed 
about their condition, better able to manage it, and thus less sensitive. We 
will be able to assess more accurately how these populations differ with respect 
to severity, appropriateness of management, and other parameters in subsequent 
phases of this study. 
<H3><A name=conclusions></A>Conclusions</H3>We detected a significant 
association between ozone and emergency department visits for asthma 2 days 
later, despite the vast majority of sampling days being below current U.S. and 
Canadian standards. The effect of ozone appeared to be reduced or absent when 
ozone concentrations were below 75 ppb. It was not influenced by the addition of 
co-pollutants into multivariate models or by the removal of repeat visits. 
However, given the limited number of sampling days for sulfate and TSP, a 
particulate effect could not be ruled out. Subsequent phases of this study will 
shed further light on a number of aspects of the observed association. 
<P>
<HR>

<P><B><FONT size=-1>References</FONT></B> 
<P><FONT size=-1>1. Kinney PL, Ozkaynak H. Associations of daily mortality and 
air pollution in Los Angeles County. Environ Res 54:99-120 (1991).</FONT> 
<P><FONT size=-1>2. Bates DV, Sizto R. Air pollution and hospital admissions in 
southern Ontario: the acid summer haze effect. Environ Res 43:317-331 
(1987).</FONT> 
<P><FONT size=-1>3. Burnett RT, Dales RE, Raizenne ME, Krewski D, Summers PW, 
Roberts GR, Raad-Young M, Dann T, Brook J. Effects of low ambient levels of 
ozone and sulfates on the frequency of respiratory admissions to Ontario 
hospitals. Environ Res 65:172-194 (1994).</FONT> 
<P><FONT size=-1>4. Thurston GD, Ito K, Kinney PL, Lippmann M. A multi-year 
study of air pollution and respiratory hospital admissions in three New York 
State metropolitan areas: results for 1988 and 1989. J Expo Anal Environ 
Epidemiol 2:429-450 (1992).</FONT> 
<P><FONT size=-1>5. Weisel CP, Cody RP, Lioy PJ. Relationship between summertime 
ambient ozone levels and emergency department visits for asthma in central New 
Jersey. Environ Health Perspect 103(suppl 2):97-102 (1995).</FONT> 
<P><FONT size=-1>6. Cody RP, Weisel CP, Birnbaum G, Lioy PJ. The effect of ozone 
associated with summertime photochemical smog on the frequency of asthma visits 
to hospital emergency departments. Environ Res 58:184-194 (1992).</FONT> 
<P><FONT size=-1>7. White MC, Etzel RA, Wilcox WD, Lloyd C. Exacerbations of 
childhood asthma and ozone pollution in Atlanta. Environ Res 65:56-68 
(1994).</FONT> 
<P><FONT size=-1>8. Romieu I, Meneses F, Sienra-Monge JJL, Huerta J, Velasco SR, 
White MC, Etzel RA, Hernandez-Avila M. Effects of urban air pollutants on 
emergency visits for childhood asthma in Mexico City. Am J Epidemiol 141:546-553 
(1995).</FONT> 
<P><FONT size=-1>9. Jones GN, Sletten C, Mandry C, Brantley PJ. Ozone level 
effect on respiratory illness: an investigation of emergency department visits. 
South Med J 88:1049-1056 (1995).</FONT> 
<P><FONT size=-1>10. Oxman AD, Krueger PD, Zangwill LM, Julian JA, Pengelly LD, 
Silverman F, Torrance GW. The health effects of tropospheric ozone: a systematic 
overview. Report for the Environmental Health Directorate, Health and Welfare 
Canada. SSC Contract #H4078-1-C082/01-SS. Hamilton, ON:McMaster University, 
1992.</FONT> 
<P><FONT size=-1>11. Horstman DH, Folinsbee LJ, Ives PJ, Abdul-Salaam S, 
McDonnell WF. Ozone concentration and pulmonary response relationships for 
6.6-hour exposures with five hours of moderate exercise to 0.08, 0.10 and 0.12 
ppm. Am Rev Respir Dis 142:1158-1163 (1990).</FONT> 
<P><FONT size=-1>12. Spektor DM, Lippmann, M, Lioy PJ, Thurston GD, Citak K, 
James DJ, Bock N, Speizer FE, Hayes C. Effects of ambient ozone on respiratory 
function in active, normal children. Am Rev Respir Dis 137:313-320 
(1988).</FONT> 
<P><FONT size=-1>13. Crapo J, Miller FJ, Mossman B, Prior WA, Kiley JP. 
Relationship between acute inflammatory responses to air pollutants and chronic 
lung disease. Am Rev Respir Dis 145:1506-1512 (1992).</FONT> 
<P><FONT size=-1>14. Devlin RB, McDonnell WF, Mann R, Becker S, House DE, 
Schreinemachers D, Koren HS. Exposure of humans to ambient levels of ozone for 
6.6 hours causes cellular and biochemical changes in the lung. Am J Respir Cell 
Mol Biol 4:72-81 (1991).</FONT> 
<P><FONT size=-1>15. Bates DV, Baker-Anderson M, Sizto R. Asthma attack 
periodicity: a study of hospital emergency visits in Vancouver. Environ Res 
51:51-70 (1990).</FONT> 
<P><FONT size=-1>16. Castellsague J, Sunyer J, Sáez M, Antó JM. Short-term 
association between air pollution and emergency room visits for asthma in 
Barcelona. Thorax 50:1051-1056 (1995).</FONT> 
<P><FONT size=-1>17. Schwartz J, Slater D, Larson TV, Pierson WE, Koenig JQ. 
Particulate air pollution and hospital emergency room visits for asthma in 
Seattle. Am Rev Respir Dis 147:826-831 (1993).</FONT> 
<P><FONT size=-1>18. Sunyer J, Antó JM, Murillo C, Sáez M. Effects of urban air 
pollution on emergency room admissions for chronic obstructive pulmonary 
disease. Am J Epidemiol 134:277-286 (1991).</FONT> 
<P><FONT size=-1>19. Samet JM, Bishop Y, Speizer FE, Spengler JD, Ferris BG. The 
relationship between air pollution and emergency room visits in an industrial 
community. J Air Pollut Control Assoc 31:236-240 (1981).</FONT> 
<P><FONT size=-1>20. Goldsmith JR, Griffith HL, Detels R, Beeser S, Neumann L. 
Emergency room admissions, meteorologic variables, and air pollutants: a path 
analysis. Am J Epidemiol 118(5):759-778 (1983).</FONT> 
<P><FONT size=-1>21. Rossi OVJ, Kinnula VL, Tienari J, Huhti E. Association of 
severe asthma attacks with weather, pollen, and air pollutants. Thorax 4:244-248 
(1993).</FONT> 
<P><FONT size=-1>22. Shumway RH, Tai R, Tai L, Pawitan Y. Statistical analysis 
of daily London mortality and associated weather and pollution effects. 
Technical Report 53. Davis, CA: Division of Statistics, University of 
California, Davis, 1983.</FONT> 
<P><FONT size=-1>23. SAS Institute Inc. SAS/STAT user's guide, version 6. 4th 
ed, vol 2. Cary, NC:SAS Institute Inc., 1989.</FONT> 
<P><FONT size=-1>24. Wedderburn RWM. Quasi-likelihood functions, generalized 
linear models and the Gauss-Newton method. Biometrika 61:439-447 (1974).</FONT> 
<P><FONT size=-1>25. Box GEP, Jenkins GM. Time series analysis: forecasting and 
control. San Francisco, CA:Holden-Day, 1976.</FONT> 
<P><FONT size=-1>26. WHO Regional Office for Europe. Update and revision of the 
Air Quality Guidelines for Europe. Copenhagen:World Health Organization Regional 
Office for Europe, 1995.</FONT> 
<P><FONT size=-1>27. Stieb DM, Beveridge RC, Brook J, Burnett RT, Anis AH, Dales 
RE. The Saint John particle health effects study: measuring health effects, 
health costs and quality of life impacts using enhanced administrative data. In: 
Particulate matter: health and regulatory issues (VIP-49), proceedings of an 
international specialty conference, 4-6 April 1995, Pittsburgh, PA. 
Pittsburgh:Air and Waste Management Association, 1995;131-142.</FONT> 
<P><FONT size=-1>28. Chidley KE, Wood-Baker R, Town GI, Sleet RA, Holgate ST. 
Reassessment of asthma management in an accident and emergency department. 
Respir Med 85:373-377 (1991).</FONT> 
<P><FONT size=-1>29. Reed S, Diggle S, Cushley MJ, Sleet RA, Tattersfield AE. 
Assessment and management of asthma in an accident and emergency department. 
Thorax 40:897-902 (1985).</FONT> 
<P><FONT size=-1>30. Fitzgerald JM, Hargreave FE. Acute asthma: emergency 
department management and prospective evaluation of outcome. Can Med Assoc J 
142(6):591-595 (1990).</FONT> 
<P><FONT size=-1>31. O'Halloran SM, Heaf DP. Accident and emergency department 
attendance by asthmatic children. Thorax 44:700-705 (1989).</FONT> 
<P><FONT size=-1>32. Canny GJ, Reisman J, Healy R, Schwartz C, Petrou C, Rebuck 
AS, Levison H. Acute asthma: observations regarding the management of a 
pediatric emergency room. Pediatrics 83(4):507-512.</FONT> 
<P><FONT size=-1>33. Brook JR, Wiebe HA, Stieb DM, Burnett RT. Saint John, NB, 
particle health effects study: determination of ambient exposures to respirable 
particles. In: Particulate matter: health and regulatory issues (VIP-49), 
proceedings of an international specialty conference, 4-6 April 1995, 
Pittsburgh, PA. Pittsburgh: Air and Waste Management Association, 
1995;347-357.</FONT> 
<P><FONT size=-1>34. Martinez FD, Wright AL, Taussig LM, Holberg CJ, Halonen M, 
Morgan WJ, Group Health Medical Associates. Asthma and wheezing in the first six 
years of life. N Engl J Med 332(3):133-138 (1995).</FONT> 
<P><FONT size=-1>35. Yaacob I, Omar R, Ahmad Mustafa WN. The outcome of patients 
with acute bronchial asthma presenting to the emergency room. Singapore Med J 
32:166-168 (1991).</FONT> 
<P><FONT size=-1>36. Newcomb RW, Akhter J. Outcomes of emergency room visits for 
asthma. J Allergy Clin Immunol 77:309-314 (1986).  
<HR>
[<A href="http://v2626umcay001/crib2/ecrib/082001-083000/082714/toc.html">Table 
of Contents</A>] [<A 
href="http://v2626umcay001/cgi-bin/pubmedf.pl?linktype=citation&pmid=9118879">Citation 
in PubMed</A>] [<A 
href="http://v2626umcay001/cgi-bin/pubmedf.pl?linktype=related&pmid=9118879">Related 
Articles</A>]</FONT> 
<P><FONT size=-1>Last Update: February 24, 1997</FONT> </P></BODY></HTML>
