1 Laboratory of Industrial Hygiene, Ministry of Public Health, Beijing, China.
2 Division of Cancer Epidemiology and Genetics, National Cancer Institute, Bethesda, MD.
3 Ministry of Health, Beijing, China.
4 International Epidemiology Institute, Rockville, MD.
5 Vanderbilt University, Nashville, TN.
6 Indoor Environments Division, US Environmental Protection Agency, Washington, DC.
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ABSTRACT |
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environment and public health; lung neoplasms; radiation; radon
Abbreviations: CI, confidence interval; GSD, geometric standard deviation
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INTRODUCTION |
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Several case-control studies of residential radon have been conducted (314
). Some studies have found no risk with indoor radon exposure, while others are consistent with increasing risk with increasing indoor exposure. Meta-analyses report a statistically significant excess risk from radon exposure (15
, 16
) but also indicate significant heterogeneity among studies, although such variability is expected (17
, 18
).
Low levels of exposure to residential radon, resulting in small excess risks, and uncertainties from reconstructing historic exposures have hampered evaluation of risk (3, 11
, 19
, 20
). To address these limitations, we conducted a case-control study in an area of China where indoor radon concentrations are high and residential mobility is low.
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MATERIALS AND METHODS |
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Underground dwellings consist of several rooms, each a tunnel 510 m long, constructed around an excavated courtyard. There are four basic designs: underground cave dwellings, open-cut cave dwellings, ground cave dwellings, and aboveground cave dwellings. Aboveground cave dwellings are constructed on the surface and have thick walls, high ceilings, and other characteristics that mimic underground types (2123
). People also live in standard aboveground dwellings with one or two stories, a single ridged roof and rectangular rooms, and multilevel apartments. A stove, which burns coal or other biomass, provides heating. The chimney is routed under a brick sleeping platform called a kang, then vented outside. Fuel is added to the firebox through an access door located either inside or outside the house.
Study subjects
Beginning in June 1995 and following approval by institutional review boards, we identified all persons aged 3075 years who were diagnosed with lung cancer between January 1994 and April 1998 and lived in the two prefectures. Cases were ascertained from each prefecture hospital, a company hospital located at a nearby oilfield, 15 county hospitals, and local clinics. We also reviewed records from antituberculosis stations and from hospitals in the large nearby cities of Lanzhou, Xian, Baoji, and Yinchuan to identify lung cancers diagnosed in residents of the two prefectures.
On the basis of clinical/radiologic symptoms suggestive of lung cancer or pathologic evidence, 1,209 possible cases were identified. An expert panel of pathologists, radiologists, and clinicians from the Gansu Department of Health reviewed all diagnoses. The panel excluded 271 subjects because of insufficient supporting evidence or incorrect diagnosis, leaving 938 cases. Of those cases, 43 could not be located, 6 were outside the age range, and 3 had moved from the area; therefore, 886 cases remained (656 males, 230 females). Diagnoses of lung cancer were based on clinical/radiologic criteria for 533 cases (60 percent) and on pathologic evidence for 353 cases (40 percent). Among the clinically/radiologically identified cases, 414 (78 percent) died before the study period ended.
We randomly selected 1,968 controls from a list of all persons included in the 1990 population census and frequency matched on age in 1995 to cases in 5-year age groups, within categories of sex and prefecture. The number of controls in each stratum was based on twice the expected number of lung cancers derived from a 1991 medical records review. Among controls, 6 refused to be interviewed, 23 had moved from the area, 62 could not be located, 73 died before 1994, 4 became cases, and 35 were not interviewed for other reasons. The study enrolled 1,310 male and 455 female controls.
After informed consent was obtained, interviews were conducted at home or at the hospital by trained interviewers using a closed-form, structured questionnaire. We asked questions on demographic characteristics, smoking habits, diet and cooking practices, and occupational, residential, and medical histories. If a subject was deceased or was too ill to participate, the interview was conducted with his or her next of kin, usually the spouse. Surrogates provided information for 481 (54 percent) cases and 71 (4 percent) controls.
Radon measurements
Interviewers placed two 1-year alpha-track detectors in each respondent's house (Track-etch; TechOps-Landauer, Glenwood, Illinois), one in the living area and one in the sleeping area. Detectors were placed in all former houses in the study area that the subject occupied for 2 or more years during the previous 30 years. For quality assurance, we placed colocated detectors in 20 percent of the houses.
We conducted a substudy to investigate variation in radon levels within and between rooms, between dwellings, and over time to provide data to adjust for exposure variability. We placed six 1-year alpha-track detectors in each room (two each at the front, middle, and back) of 55 houses during 3 consecutive years, starting in July 1996. A total of 1,654 detectors were placed in one to five rooms of each house (mean, 2.3 rooms/house).
Assignment of radon exposure
We defined reference age as age at diagnosis for a case and at interview for a control. We designated 530 years prior to the reference age as the time-relevant exposure period most related to lung cancer risk (1). For nearly all cases (881 of 886) and controls (1,761 of 1,765), at least one radon measurement was available. For 88 percent (775 of 881) of cases and 95 percent (1,669 of 1,765) of controls, at least one measurement was within the exposure window from 1.9 (for cases) and 1.6 (for controls) mean eligible residences per subject.
For analysis, we used time-weighted average radon concentration within the exposure window measured in becquerels per cubic meter (Bq/m3), using number of years resident as weights. (Becquerel is an international unit of radioactivity; 1 Bq = 1 disintegration per second.) Two controls had elevated radon values (1,554 and 1,676 Bq/m3) that were more than 40 percent higher than the next-largest value and were omitted from our analyses, although this exclusion had little impact on inference. We imputed values for gaps in residential histories due to missing measurements or for less than 2 years of occupancy by using mean radon concentration in the houses of controls, within house type and prefecture (24). Alternatives, such as mean radon level within prefecture, made little difference. To adjust variances for imputation, we selectively computed estimates by using multiple imputation (25
, 26
), but variance correction proved unnecessary because of high coverage of the exposure window.
Statistical analysis
We computed odds ratios adjusted for age, sex, prefecture, tobacco use, and, where appropriate, other factors by using unconditional logistic regression (27). We calculated 95 percent Wald confidence intervals for odds ratios and used a score statistic for tests of trend. We also fitted a linear model, odds ratio(x) = 1 + ß x, in which x was the mean radon level and ß was the excess odds ratio per becquerel/cubic meter. We computed likelihood-based confidence intervals for estimates of ß. Homogeneity of ß across categories of other factors was evaluated by using a likelihood ratio test.
Three important sources of error in assessing radon exposure were 1) detector measurement error, 2) use of contemporary measurements to estimate radon levels throughout the house and in prior years, and 3) missing radon values. Detector error was relatively small and was ignored. Estimation of radon induced classical error, while missing data induced Berkson error (3, 19
). To adjust for error, we restricted data to subjects for whom coverage of the exposure window was 70 percent or higher, thus minimizing Berkson error, so that classical error predominated.
Specifically, suppose that Xi was the true, but unobserved radon concentration in the ith house within the exposure period, Pi was the proportion of years spent in the ith house, and Zi was the estimated concentration. Further suppose that true radon exposure for a person was 25 times iPiXi, while observed radon exposure was 25 times
iPiZi. We assumed that each Xi was lognormally and independently distributed with parameters µ and
2 and that Ui was a multiplicative random error, independent of Xi and lognormally distributed with parameters 0 and
2. Then, Zi = Xi x Ui and is lognormally distributed with parameters µ and
2 1
2. Measurement data obtained from houses in the full study provided estimates of µ and
2 +
2, while the radon substudy provided an estimate of
2.
For houses included in the substudy, arithmetic means were 366.5, 338.4, 378.1, 361.0, and 343.2 Bq/m3 for underground cave dwellings, open-cut cave dwellings, ground cave dwellings, aboveground cave dwellings, and standard aboveground dwellings, respectively. The corresponding geometric means (and geometric standard deviations (GSDs)) were 338.7 Bq/m3 (1.52), 314.2 Bq/m3 (1.50), 347.6 Bq/m3 (1.52), 336.0 Bq/m3 (1.48), and 311.2 Bq/m3 (1.58). Apartments were not included in the substudy. Houses had not been modified extensively, so we estimated 2 by assuming that uncertainties resulted from random variations in radon concentration within houses and over time and that residential mobility was unrelated to radon, conditional on housing type. Use of a component of variance analysis estimated
2 as 0.16, or a GSD of about 1.5 for the error distribution. We evaluated a range of GSDs1.25, 1.5, and 1.6 (or coefficients of variation of 0.23, 0.42, and 0.50)to investigate the sensitivity of the error on the excess odds ratio.
Under our assumptions, the true value given the observed value, denoted Xi|Zi, was lognormally distributed with parameters (µ 2 + log(Zi)
2)/(
2 +
2) and
2
2/(
2 +
2) (18
). We used a Monte Carlo approach to evaluate error. For each subject's house, we randomly sampled from the Xi|Zi distribution, computed a "true" time-weighted average radon concentration, and estimated the excess odds ratio per becquerel/cubic meter. This process was repeated 1,000 times to obtain the empirical distribution of the estimated ß and its 95 percent confidence interval. This approach was less formal than the one used by Reeves et al. (19
) but was similar to the Lagarde et al. approach (20
).
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RESULTS |
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Most men smoked (92.3 percent), but most women did not (10.4 percent). The odds ratio for ever smokers compared with never smokers was 1.69 (95 percent confidence interval (CI): 1.2, 2.4) and was similar for males and females. Empirical analyses indicated that the increase in the logarithm of the odds ratio per manufactured cigarette smoked per day was one third the increase per liang (50 g) of tobacco smoked per month in hand-rolled cigarettes and about the same per liang of tobacco smoked per month in a pipe. We used these results to create cigarette-equivalents per day by summing number of cigarettes smoked per day, 3.0 times liang per month smoked in hand-rolled cigarettes, and liang per month smoked in pipes. Among smokers, cases and controls smoked 17.9 and 12.9 cigarette-equivalents per day for 30.3 and 29.7 years, respectively. We also created a smoking risk variable to account for duration and number of cigarette-equivalents smoked per day (table 2). Odds ratios increased with increasing tobacco exposure and were homogeneous by sex.
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Radon exposure
Mean radon concentrations for cases and controls were 230.4 and 222.2 Bq/m3, respectively, and those for 81.6 percent of cases and 76.3 percent of controls were at or above 150 Bq/m3. Odds ratios increased significantly with increasing concentration (p < 0.001) (table 3, figure 2). The test for departure from linearity was not significant (p = 0.10). The estimated excess odds ratio at 100 Bq/m3 was 0.19 (95 percent CI: 0.05, 0.47).
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There were 297 histologically confirmed cases of lung cancer. The excess odds ratios for an exposure level of 100 Bq/m3 were 0.14 (95 percent CI: 0.03, 0.54) when confirmed cases were used and 0.20 (95 percent CI: 0.03, 0.55) when clinically diagnosed cases were used. Neither excess odds ratio differed significantly from the overall value of 0.19.
Table 4 shows the odds ratios for radon within categories of various other factors. There was no significant variation in radon effects by any factor except type of house. In addition, we found a suggestion of declining effects with reference age. Type of house, smokiness, and coal referred to the house in which a subject lived the longest, but results were similar for the current house. Heterogeneity was found in the excess odds ratio by type of house, with no trend for subjects living in standard aboveground dwellings or apartments. This difference in trend by house type was reduced when smoking and housing type were included as stratification variables and data were restricted. The excess odds ratios at 100 Bq/m3 (and p values for homogeneity) for underground and standard housing types were 0.33 and 0.03 (p = 0.15), 0.32 and 0.10 (p = 0.35), and 0.35 and 0.17 (p = 0.54) when coverage of the exposure window was 70 percent or higher, 90 percent or higher, and 100 percent, respectively.
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Adjustment for uncertainty about radon exposure
Among subjects for whom coverage of the exposure window was 70 percent or higher, the excess odds ratios at 100 Bq/m3, adjusted for error GSDs of 1.25, 1.5, and 1.6, were 0.27 (95 percent CI: 0.03, 0.69), 0.32 (95 percent CI: 0.08, 1.37), and 0.59 (95 percent CI: 0.14, 2.73), respectively, in contrast to the unadjusted estimate of 0.22 (95 percent CI: 0.06, 0.54). Excess odds ratio estimates, as well as the widths of the confidence intervals, increased with greater exposure error.
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DISCUSSION |
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There have been several case-control studies of residential radon and lung cancer in which long-term detectors were used (314
). Meta-analyses of those studies estimated an excess odds ratio of 0.10.2 at 100 Bq/m3 (3
, 15
, 16
). Extrapolations based on results from miners exposed to low radon concentrations result in excess odds ratios of about 0.12 (2
), similar to the combined estimate and slightly lower than the unadjusted estimate in our study.
Risk estimates from meta-analyses of residential studies and from pooled miner analyses do not account for errors in exposure assessment. Stability of our population may indicate improved exposure assessment, resulting in the higher excess odds ratio estimates. At 100 Bq/m3, estimates of excess odds ratios were 0.22, 0.26, and 0.31 when data were limited to subjects for whom coverage of the exposure window was 70 percent or higher, 90 percent or higher, and 100 percent, respectively.
Three recent studies with enhanced exposure assessments also suggest that the risk of lung cancer may be higher than previously estimated. A study of Missouri women based exposure on CR-39 surface measurement devices and reported an excess odds ratio of 0.63 (95 percent CI: 0.1, 1.9) at 100 Bq/m3 (11). These devices measure emissions from polonium-210 embedded in glass artifacts, such as picture frames and mirrors, which may better reflect historical exposure than contemporary air measurements, since the artifact serves as a continuous recording device. An Iowa study enrolled only long-term (20 years or more) residents, thereby minimizing uncertainties resulting from residential mobility (12
). The exposure assessment included measurements throughout each house, residential occupancy, and time spent in other buildings and outdoors (29
). The estimated excess odds ratios at 100 Bq/m3 ranged from 0.16 (95 percent CI: 0.0, 0.6) for all subjects to 0.33 (95 percent CI: 0.02, 1.23) for living subjects. A study in Finland enrolled residents of 20 years or more and estimated the excess odds ratio at 100 Bq/m3 as 0.11 (95 percent CI: 0.9, 1.3) (4
).
Precise characterization of exposure error and adjustment of risk estimates are problematic. Previous adjustments increased excess odds ratio estimates by about 50100 percent. In a southwest England study, the estimate of the excess odds ratio at 100 Bq/m3 increased after adjustment from 0.08 (95 percent CI: 0.03, 0.20) to 0.12 (95 percent CI: 0.05, 0.33) for all subjects and from 0.14 (95 percent CI: 0.01, 0.29) to 0.24 (95 percent CI: 0.01, 0.56) for subjects among whom coverage of the exposure window was complete (3). Depending on error assumptions, excess odds ratio estimates in a Swedish study increased from 0.10 to 0.150.20 (20
). Our best estimate of uncertainty increased the excess odds ratio by about 50 percent.
Our evaluation of exposure misclassification did not account for time spent in the house. Subjects reported spending about half their time indoors during adulthood. For males and females, mean occupancy during their adult years was 11.8 (49 percent) and 11.9 (50 percent) hours per day during summer months and 12.8 (53 percent) and 13.7 (57 percent) hours per day during winter months, respectively. Because most subjects were farmers, time not spent in their houses was likely spent outdoors. For our subjects, the length of occupancy was less than the assumed 6090 percent found in North American and European studies (30). In our data, occupancy was related to reference age, increasing 11.5 hours between ages 40 and 70 years. However, we had no data on the variation in occupancy throughout life, which changes substantially during adulthood (12
).
To our knowledge, there has been only one other large study of radon and lung cancer in China (14). That study, carried out in the northern industrial city of Shenyang, included 308 lung cancer cases and 356 controls. Odds ratios for all radon categories were less than one and were nonsignificant. The absence of significant findings may reflect the urban location, the higher levels of outdoor air pollution, or lower radon levels, which were measured in only a single house per subject. Subjects reported a median of 24 years of residency, which represents 19 years in the 530-year exposure-relevant period, and a median radon level of 85 Bq/m3, about 60 percent lower than in our study. For subjects in our study who had occupied their current house for at least 5 years, mean coverage was 28 of the previous 30 years, which represents 23 years during the exposure-relevant period.
Analyses of miners have suggested that the relative risk trend for radon is higher for never smokers and younger persons (1). Indoor radon studies, including the current one, show inconsistent patterns of risk by smoking status and age (table 5). Excess odds ratios seemed to decline with increasing attained age in the Finland-II (4
), Missouri-I (8
), Missouri-II (11
), southwest England (3
), and Stockholm, Sweden (6
) studies and suggestively in our study, but they did not vary in the Finland-I (9
) and Iowa (12
) studies. Data from the Stockholm study and living respondents in the Missouri-I study provide only suggestive evidence for a greater odds ratio trend with radon exposure for never smokers.
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Because miner studies included males only, these studies are uninformative about radon risks for females. Risk extrapolations to females for residential exposures have relied on the assumption of equivalent susceptibility (1). There is evidence from the Finland-II (4
) and southwest England (3
) studies that the excess odds ratio for radon is higher for males than for females; however, our study suggests no such difference (table 5). The question of differential effects by sex remains unresolved.
A potential confounder in our study was indoor air pollution, since most subjects used coal, wood, or sticks in a stove or kang for cooking and heating. In April 1995, we measured particle-bound polycyclic aromatic hydrocarbons, particulate matter smaller than 10 microns (PM10), carbon monoxide, nitrogen dioxide, sulfur dioxide, and air exchange rate in 25 dwellings (22, 32
). Ventilation rates were high, averaging 1.5 air exchanges per hour, and resulted in pollutant levels that were episodic and elevated only during stove use. Except for carbon monoxide and PM10, mean values were below US Ambient Air Quality Standards (for more information, refer to the following Internet Web site: http://www.epa.gov/airs/criteria.html). We did not have air pollutant measurements for each study house; however, odds ratios for radon did not vary significantly with level of indoor smokiness as reported by the respondent. Odds ratio trends for radon were similar by house type after we included smoking risk and house type as stratification variables.
Information on more than half of our cases came from their next of kin, who may have been less knowledgeable about life events, raising the possibility that results might have been affected by differential misclassification. However, odds ratios were similar after adjustment for source of information or when data were restricted to subject respondents only.
In conclusion, radon concentrations were high in our study, exceeding those found in most previous indoor studies, and the population was stable and rural. Results provide evidence that high levels of residential radon increase the risk of lung cancer and support the findings from meta-analyses of indoor studies and from miners. In addition, our estimates suggest that effects of residential radon may equal or exceed miner-based estimates, which are currently used to evaluate risk.
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ACKNOWLEDGMENTS |
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The authors acknowledge the contributions of Drs. Jiguo Shi, Yuelan Kang, Datong Zhao, Fuhui Du, and Yucheng Wang for identifying lung cancer cases, and they thank the many local investigators from Pingliang and Qingyang Prefectures for their assistance with data collection. The authors also thank all those persons who contributed to part of the project as well as Margaret Pacious and Lew Gaul of WESTAT, Inc. for program management and data processing support for this study.
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NOTES |
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REFERENCES |
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