Ambient Air Pollution and Cancer Mortality in the Cancer Prevention Study II

Background: The International Agency for Research on Cancer classified both outdoor air pollution and airborne particulate matter as carcinogenic to humans (Group 1) for lung cancer. There may be associations with cancer at other sites; however, the epidemiological evidence is limited. Objective: The aim of this study was to clarify whether ambient air pollution is associated with specific types of cancer other than lung cancer by examining associations of ambient air pollution with nonlung cancer death in the Cancer Prevention Study II (CPS-II). Methods: Analysis included 623,048 CPS-II participants who were followed for 22 y (1982–2004). Modeled estimates of particulate matter with aerodynamic diameter <2.5µm (PM2.5) (1999–2004), nitrogen dioxide (NO2) (2006), and ozone (O3) (2002–2004) concentrations were linked to the participant residence at enrollment. Cox proportional hazards models were used to estimate associations per each fifth percentile–mean increment with cancer mortality at 29 anatomic sites, adjusted for individual and ecological covariates. Results: We observed 43,320 nonlung cancer deaths. PM2.5 was significantly positively associated with death from cancers of the kidney {adjusted hazard ratio (HR) per 4.4 μg/m3=1.14 [95% confidence interval (CI): 1.03, 1.27]} and bladder [HR=1.13 (95% CI: 1.03, 1.23)]. NO2 was positively associated with colorectal cancer mortality [HR per 6.5 ppb=1.06 (95% CI: 1.02, 1.10). The results were similar in two-pollutant models including PM2.5 and NO2 and in three-pollutant models with O3. We observed no statistically significant positive associations with death from other types of cancer based on results from adjusted models. Conclusions: The results from this large prospective study suggest that ambient air pollution was not associated with death from most nonlung cancers, but associations with kidney, bladder, and colorectal cancer death warrant further investigation. https://doi.org/10.1289/EHP1249

Ambient air pollution represents a complex mixture of a broad range of carcinogenic and mutagenic substances that may play a role in chronic systemic inflammation, oxidative stress, and DNA damage in tissues other than the lung (Brook et al. 2010;Crouse et al. 2010;IARC 2012IARC , 2013Newby et al. 2015). As such, there may also be associations between ambient air pollution and other types of cancer; however, the epidemiological evidence for these associations is limited (IARC 2013).
Some previous studies reported positive associations of ambient air pollution with fatal bladder cancer (Liu et al. 2009), pancreatic cancer (Ancona et al. 2015), upper digestive tract, digestive accessory organs, and breast cancer (Wong et al. 2016), and with incident brain and cervical cancer (Raaschou-Nielsen et al. 2011), breast cancer (Crouse et al. 2010;Reding et al. 2015), hepatocellular carcinoma (Pan et al. 2015), and prostate cancer (Parent et al. 2013). Results from other studies of fatal brain cancer (McKean-Cowdin et al. 2009), incident brain tumors , and incident leukemia (Winters et al. 2015) were mixed or null. Previous studies are difficult to compare given variation in the specific cancers studied; whether outcomes were incident versus fatal cancers; and differences in study design, sample size, exposure assessment, and the availability of data on potential confounders.
We previously reported positive associations between longterm ambient air pollution exposure and lung cancer mortality in the American Cancer Society (ACS) Cancer Prevention Study II (CPS-II) Krewski et al. 2000Krewski et al. , 2009Turner et al. 2011Turner et al. , 2014Turner et al. , 2016. For example, we estimated that a 10-lg=m 3 increase in PM 2:5 concentrations was associated with a 9% (95% CI: 3, 16%) increase in lung cancer mortality (Turner et al. 2016, Table E4). The CPS-II is a large-scale prospective cohort in which the human health effects of ambient air pollution have been extensively investigated. The objective of this work was to clarify whether ambient air pollution is associated with specific types of cancer other than lung cancer by examining associations of ambient air pollution with nonlung cancer death among 623,048 CPS-II participants who were followed for 22 y.

Study Population
Approximately 1.2 million CPS-II participants were enrolled by 77,000 volunteers in 1982 throughout the entire United States, Washington, DC, and Puerto Rico. Participants, largely friends and family members of volunteers, were ≥30 y old with a family member ≥45 y old. A detailed self-administered questionnaire was completed at enrollment to obtain demographic, medical, and behavioral information (CPS-II questionnaires are available at https://www.cancer.org/research/we-conduct-cancer-research/ epidemiology/cancer-prevention-questionnaires.html). Informed consent was implied by completion and return of the enrollment questionnaire. The Emory University School of Medicine Human Investigations Committee has approved all aspects of the CPS-II.
Volunteers ascertained the vital status of participants they had enrolled in 1984, 1986, and 1988, and death certificates, coded by trained nosologists, were used to determine the underlying cause of death, that is, "the disease or injury which initiated the train of morbid events leading directly to death" (WHO 1992). Since 1989, the National Death Index (NDI) has been used to determine vital status and underlying cause of death (Calle and Terrell 1993). Underlying cause of death was categorized according to the International Classification of Diseases (ICD) revisions 9 and 10 (WHO 1977(WHO , 1992. More than 99% of known deaths were assigned a cause.

Ambient Air Pollution Concentrations
PM 2:5 concentrations were assigned to the geocoded participant residence at enrollment based on a national-level hybrid land use regression (LUR) and Bayesian maximum entropy (BME) interpolation model ) used in previous work (Jerrett et al. 2016;Pope et al. 2015;Turner et al. 2014;. Details on the geocoding of participant residences are presented elsewhere (Pope et al. 2015). A total of 104,172 observations were collected monthly from 1,464 monitoring sites between 1999 and 2008. Approximately 10% of observations were reserved for cross-validation. PM 2:5 concentrations were predicted in a base LUR model that included traffic within 1 km (based on modeled traffic counts) and the cube of green space within 100 m. A BME interpolation model was then used to account for residual spatiotemporal variation in PM 2:5 concentrations, and the two estimates were combined. The cross-validation R 2 was 0.79, suggesting good spatiotemporal model prediction at locations other than those used to calibrate the model. Mean PM 2:5 concentrations at the enrollment residence for the period 1999-2004 were used in the present analysis to coincide with the cohort follow-up period.
NO 2 concentrations were assigned at the census block group level to each participant enrollment residence. A national LUR model was used that incorporated both hourly monitoring data from 423 monitors and ∼ 4 million satellite-based measurements in an approximate 10 km × 10 km grid scale, including additional data on population density, land use, and distance to roadways for the year 2006 (model R 2 = 0:78) (Novotny et al. 2011;Turner et al. 2016 (Turner et al. 2016).

Statistical Methods
Both single-and multipollutant Cox proportional hazards regression models were used to estimate associations of ambient PM 2:5 , NO 2 , and O 3 concentrations with death from 29 specific types of cancer according to an increment of the mean minus the fifth percentile for each pollutant (4:4 lg=m 3 for PM 2:5 , 6:5 ppb for NO 2 , and 6:9 ppb for O 3 ) ( Table 1). The fifth percentile-mean increment was used for comparability of results across pollutants and with previous work (Turner et al. 2016).
Models were stratified by 1-y age categories, gender, and race/ethnicity (white, black, other) to allow for separate baseline hazards according to these characteristics. In addition, in the multivariate model, we also adjusted for baseline values of education (<high school, high school, >high school); marital status (single, married, other); body mass index (BMI); BMI squared; smoking status (never cigarettes, pipes, or cigars; current cigarette smoker only; former cigarette smoker only; ever pipe/cigar); cigarettes per day and cigarettes per day squared (current and former cigarette smokers); years smoked and years smoked squared (current and former cigarette smokers); started smoking <18 y old (yes/ no) (current and former cigarette smokers); passive smoking (hours per day exposed to cigarette smoke of others at home, work, or other areas); usual dietary intake of vegetables, fruit, and fiber (combined, in quintiles) and fat (quintiles), plus a The land use regression Bayesian maximum entropy (LURBME) PM 2:5 model was created in three main steps: a) A base LUR model predicted PM 2:5 concentrations based on traffic within 1 km (based on modeled traffic counts) and the cube of green space within 100 m; b) A BME interpolation model was then used to interpolate residual spatiotemporal variation in PM 2:5 concentrations; c) The two estimates were then combined. Regional PM 2:5 concentrations therefore represent a residual with some observations less than zero.
category for missing/bad dietary intake data; usual consumption of beer (yes, no, missing), wine (yes, no, missing), and liquor (yes, no, missing); occupational exposures (ever regular exposure to asbestos, chemicals/acids/solvents, coal or stone dusts, coal tar/pitch/asphalt, formaldehyde, or diesel engine exhaust) (yes/ no); and an occupational dirtiness index to characterize workplace PM 2:5 exposure based on main lifetime occupation (six categories of exposure vs. a referent category, or missing) (Siemiatycki et al. 2003). In addition, we used 1990 census data to derive the following socioecomonic covariates defined at the ZIP-code level for each participant's residence at enrollment: median household income, percentage of African American residents, percentage of Hispanic residents, percentage of adults with post-secondary education, percentage of unemployed residents ≥16 y old, and percentage of residents with household incomes <125% of the poverty level. For each characteristic, we modeled the value for the ZIP code of residence and for a second variable indicating the difference between that value and the county-level mean (Jerrett et al. 2016;Pope et al. 2015;Turner et al. 2016; U.S. Department of Commerce Bureau of the Census 1993). The time axis was follow-up time in days. Participants lost to follow-up or who were alive at the end of follow-up were censored. For analysis of death from cancers of the uterus, cervix, or ovary, participants were excluded if they reported a previous hysterectomy or an artificial (vs. a natural) menopause (n = 108,956). An additional 7,143 participants were also excluded from analysis of ovarian cancer mortality if they reported having undergone an ovarian surgery because no information was available to distinguish partial from total oophorectomy.
A sensitivity analysis was performed to examine the influence of including additional baseline variables in the model including usual physical exercise (none, slight, moderate, heavy, missing), aspirin use in the past month (yes/no), and usual dietary intake of red meat (quintiles) for analysis of all cancer sites, as well as self-reported physician-diagnosed diabetes (yes/no). Detailed reproductive and hormonal variables, including age at menarche (<12 y old, 12-13 y old, ≥14 y old, missing), parity (0, 1, 2, 3, ≥4, missing), age at first birth (none, <20 y old, 20-29 y old, ≥30 y old, missing), ever oral contraceptive use (yes, no, missing), ever postmenopausal hormone use (yes, no, missing), and postmenopausal status (yes, no, missing) were also included in models for sensitivity analysis of female reproductive cancers. Additional analyses were also performed in postmenopausal women only.
Further, we decomposed PM 2:5 concentrations into nearsource and regional fractions to determine whether associations might vary for PM 2:5 from different air pollution sources (Turner et al. 2016). Near-source PM 2:5 concentrations were PM 2:5 concentrations predicted at each monitoring station based on landuse and traffic count data using the LUR model created in the first stage of the LURBME modeling process. Regional PM 2:5 was derived by subtracting the near-source PM 2:5 concentration from the total PM 2:5 concentrations from the LURBME model, thereby capturing spatial variation in PM 2:5 concentrations between monitoring stations.
Finally, where there were statistically significant positive associations observed, potential effect modification by gender, education (<high school, high school, >high school), and smoking status (never cigarettes, pipes, or cigars; current cigarette smoker only; former cigarette smoker only) was assessed on a multiplicative scale. Two-sided p-values were calculated according to the likelihood ratio statistic to compare models with and without multiplicative interaction terms between air pollutants and the potential modifiers. The proportional hazards assumption was tested by including interaction terms between ambient air pollution and follow-up time in the multivariate proportional hazards models. A p-value <0:05 was used to define statistical significance throughout this work. SAS (version 9.2; SAS Institute Inc.) was used to conduct all analyses. Ethics approval for analysis was obtained from the Ottawa Hospital Research Ethics Board.

Results
From 1982 through 2004 there were a total of 743,543 (62.8%) participants who were alive and 438,123 (37.0%) who had died. Participants were excluded from analysis because of missing or invalid residence information (n = 385,422) or selected baseline covariate data (n = 130,119) (Jerrett et al. 2016;Pope et al. 2015;Turner et al. 2016) or having a prevalent cancer (except nonmelanoma skin cancer) at enrollment (n = 45,998). Follow-up was censored in September 1988 for 2,921 (0.2%) of participants who had insufficient information to link to the NDI. The present analysis is based on 623,048 CPS-II participants among whom 43,320 nonlung cancer deaths were observed during 11,936,799 person-years of follow-up. Follow-up ranged from 0.01 y to 22.5 y with a mean (SD) of 19.2 (5.6) y.
Participant characteristics are presented in Table 2 and Table  S2. The majority of participants were 40 y old to 69 y old at enrollment, 94.5% of participants were white, 55.3% were female, and 57.4% had a greater than high school education. A total of 44.7% of study participants were never smokers.
There was little variation in ambient air pollution concentrations by participant characteristics, although somewhat higher PM 2:5 and NO 2 concentrations were observed in the younger and older age groups and among participants who were black or "other" race/ethnicity (Table 2). O 3 concentrations were slightly higher in older participants, in those with a low BMI (<18:5 kg=m 2 ), and among never smokers.

Discussion
Ambient air pollution was not associated with death from most nonlung cancers in our large prospective study population. However, there were statistically significant positive associations between PM 2:5 and death from bladder and kidney cancer, ranging from 13-14% increases in risk per each fifth percentile-mean increment (4:4 lg=m 3 ). For NO 2 , there was a statistically significant positive association with colorectal cancer mortality of 6% per 6:5 ppb. The results were similar in two-pollutant models including both PM 2:5 and NO 2 as well as in three-pollutant models with O 3 . The magnitudes of the associations here were somewhat stronger than, although compatible with, those of PM 2:5 and lung cancer mortality observed in previous work [9% (95% CI: 3, 16%) per 10 lg=m 3 or equivalently, 4% (95% CI: 1, 7%) per 4:4 lg=m 3 ] (Turner et al. 2016).
A small but growing body of literature has examined associations between ambient air pollution and nonlung cancer risk; however, the evidence for such associations is limited. For bladder cancer, a Spanish hospital-based case-control study including 1,219 incident cases and 1,271 controls reported that living Table 3. Adjusted HRs (95% CIs) a for nonlung cancer mortality per each fifth percentile-mean increment in air pollutant concentrations, single-pollutant models, CPS-II cohort, United States   Age, race/ethnicity, gender stratified and adjusted for baseline values of education; marital status; body mass index; body mass index squared; smoking status; cigarettes per day; cigarettes per day squared; duration of smoking; duration of smoking squared; age started smoking; passive smoking, vegetable/fruit/fiber consumption; fat consumption; beer, wine, liquor consumption; industrial exposures; occupation dirtiness index; and 1990 ecological covariates. b As above but not gender stratified. Analysis limited to 344,593 women for analysis of breast cancer mortality, 235,637 women for analysis of uterine or cervical cancer mortality, and 228,494 women for analysis of ovarian cancer mortality. Analysis limited to 278,455 men for analysis of prostate cancer mortality. c Women reporting a previous hysterectomy or an artificial menopause excluded here (n = 108,956). d Women reporting having undergone an ovarian surgery also excluded here (n = 7,143). >40 y in a city of >100,000 inhabitants was associated with a significantly higher risk of the disease [odds ratio ðORÞ = 1:30 (95% CI: 1.04, 1.63)] (Castaño-Vinyals et al. 2008). In a death certificate-based case-control study in Taiwan, including 680 bladder cancer deaths and 680 matched noncancer or nongenitourinary death controls, there were significant positive trends with increasing tertiles of particulate matter with aerodynamic diameter <10 lm (PM 10 ), NO 2 , and sulfur dioxide (SO 2 ), but not with carbon monoxide (CO) or O 3 concentrations (Liu et al. 2009). There was a positive, nonsignificant association between nitrogen oxides (NO x ) concentrations and bladder cancer incidence (n = 221) [incidence rate ratio (IRR) per 100 lg= m 3 = 1:32 (95% CI: 0.80, 2.19)] in an analysis of 54,304 participants in the Danish Diet Cancer and Health cohort (Raaschou-Nielsen et al. 2011). In a retrospective cohort study of 85,559 individuals in Malagrotta (Rome, Italy), there was a positive association between hydrogen sulfide (H 2 S) concentrations from a municipal waste landfill and bladder cancer mortality in women [HR per 0:043 lg=m 3 = 1:35 (95% CI: 1.00, 1.82), n = 12], but not in men [HR = 0:88 (95% CI: 0.51, 1.52), n = 61] (Ancona et al. 2015). Taken together, the results from our study and those from most previous studies are generally consistent with the recent IARC evaluation noting a positive association in studies of outdoor air pollution and bladder cancer risk (IARC 2013).
There was no association [HR per 10 lg=m 3 = 0:98 (95% CI: 0.58, 1.64)] between PM 2:5 and urinary cancer mortality (including 155 bladder and kidney cancer deaths combined) in a cohort of 66,820 elderly Hong Kong residents (Wong et al. 2016). The Danish Diet Cancer and Health cohort reported a positive, although imprecise, association of NO x with kidney cancer incidence [IRR = 1:73 (95% CI: 0.89, 3.73) per 100 lg=m 3 (Raaschou-Nielsen et al. 2011). A total of 95 kidney cancer cases were observed in that study.
Few studies of colorectal cancer have been reported to date. Although our study showed a significant positive association between NO 2 and death from colorectal cancer and a borderline association with PM 2:5 , a Hong Kong cohort reported no association [HR per 10 lg=m 3 = 1:01 (95% CI: 0.79, 1.30)] between PM 2:5 concentrations and mortality from cancers of the lower digestive tract (n = 719) (Wong et al. 2016). There were nonsignificant inverse associations in men and women for PM 10 concentrations from a waste incinerator and colorectal cancer mortality (n = 149 deaths in men and 95 deaths in women) in Malagrotta (Ancona et al. 2015). There were also nonsignificant inverse associations for NO x concentrations and both colon and rectal cancer incidence in Denmark (n = 414 and 246 cases, respectively) (Raaschou-Nielsen et al. 2011). The number of included colorectal cancer deaths (>6,000) in CPS-II here is substantially larger than the numbers of cases/deaths in previous studies.
When PM 2:5 concentrations were decomposed into nearsource and regional components, there were stronger associations for near-source PM 2:5 with colorectal (and stomach) cancer mortality, supporting similar findings for NO 2 , and for regional PM 2:5 with both bladder and kidney cancer mortality. Little is known regarding the role of different air pollution sources or components in cancer, including at sites other than the lung. We have not specifically examined what PM 2:5 sources were further related to this decomposition.
Studies of occupational exposure to diesel engine exhaust and polycyclic aromatic hydrocarbons (PAHs), which are formed during incomplete combustion processes, have suggested positive associations with both bladder and kidney cancer (Brown et al. 2012;IARC 2005;Siemiatycki et al. 2004). Recent populationbased case-control studies of men in Canada reported positive associations between any exposure to high concentrations of occupational diesel exhaust and both bladder [OR = 1:64 (95% CI: 0.87, 3.08), n = 658 cases and 1,360 controls] and rectal [OR = 1:98 (95% CI: 1.09, 3.60), n = 840 cases and 1,360 controls] cancer incidence, which increased among those with >10 years of exposure (Kachuri et al. 2016;Latifovic et al. 2015). Studies of workers in dusty occupations (e.g., mineral dust, wood dust), as well as in steel and iron processing, have also noted positive associations with colorectal and stomach cancer (Kreuzer et al. 2012;Oddone et al. 2014;Raj et al. 2003;Santibañez et al. 2012).
The largest HR was for PM 2:5 in relation to death from cervical cancer, although the findings were based on only 115 deaths and were not statistically significant. The positive association with NO 2 was weaker. In the Danish Diet Cancer and Health study, the strongest association was between NO x and cervical cancer incidence [i.e., IRR = 2:45 (95% CI: 1.01, 5.93) per 100 lg=m 3 ], but it was based on only 35 cases (Raaschou-Nielsen et al. 2011). In both studies, there were no data on human papillomavirus (HPV) infection (or on other potentially relevant infections for cancer mortality at other sites, e.g., Helicobacter pylori for stomach cancer or hepatitis B or C for liver cancer), and confounding cannot be ruled out. There was also no information on access to or compliance with cervical cancer screening programs, which may differ in areas with differing levels of ambient air pollution. However, we adjusted our models for a range of area-level socioecomonic covariates and screening rates in a subset of CPS-II participants for both breast (>90%) and colorectal (>65%) cancer were high (Patel et al. 2003;Stevens et al. 2011). A recent cross-sectional study of women from a clinical trial of cervical disease diagnostic techniques in Texas reported a positive association of cervical dysplasia with residential census-tract level estimates of ambient benzene, diesel particulate matter, and PAH concentrations (Scheurer et al. 2014).
Although we found no significant associations with death from breast cancer based on fully adjusted models, other studies reported positive associations between ambient air pollution and breast cancer incidence (Crouse et al. 2010;Mordukhovich et al. 2016;Reding et al. 2015). Crouse et al. (2010) reported a positive association between NO 2 concentrations and postmenopausal breast cancer incidence [OR per 5 ppb = 1:31 (95% CI: 1.00, 1.71)] in a hospital-based case-control study including 383 case and 416 control participants. The prospective U.S. Sister Study reported no association between PM 10 , PM 2:5 , or NO 2 concentrations and incident breast cancer overall in an analysis of 47,591 participants including 1,749 breast cancer cases, but a positive association was reported between NO 2 and estrogen receptor ðERÞ + = progesterone receptor ðPRÞ + disease [HR per 5:8 ppb = 1:10 (95% CI: 1.02, 1.19)] (Reding et al. 2015). In our mortality-based study, no information on ER or PR status was available.
There was no positive association of any ambient air pollutant and total leukemia mortality in the present study. There was a positive association of NO 2 concentrations at the residence with incident adult acute myeloid leukemia [OR per 10 mg=m 3 = 1:31 (95% CI: 1.02, 1.65), n = 531] but not with other leukemia subtypes in a Danish case-control study ). There was no positive association of PM 2:5 with total incident leukemia or with chronic lymphocytic leukemia in a Canadian casecontrol study of 1,064 total case and 5,039 control participants (Winters et al. 2015). There was no information on leukemia subtype in CPS-II.
There was no association of any ambient air pollutant with total brain cancer mortality in this analysis, which included 1,591 brain cancer deaths. The Danish Diet Cancer and Health cohort reported a positive association of NO x with incident brain tumor risk [IRR per 100 lg=m 3 NO x = 2:28 (  ). There was no information on tumor histology or cranial location in CPS-II.
In the present study, there were no clear positive, and some cases there were significant inverse, associations observed with O 3 in both single-and multipollutant models possibly owing to broader spatial patterns in ambient air pollution concentrations, to negative correlations with other air pollutants, and to the larger spatial scale of O 3 concentrations, which were unable to capture fine-scale variation and scavenging effects in urban areas (Williams et al. 2014). O 3 is thought to increase DNA damage (U.S. EPA 2013), which plays a role in several types of cancer. However, in previous studies of the CPS-II cohort, O 3 was not associated with lung cancer death Krewski et al. 2009;Pope et al. 2002;Turner et al. 2016). A positive association between ambient O 3 and incident male lung cancer was reported in the Adventist Health Study on Smog (AHSMOG), although few lung cancers were observed (n = 16) (Beeson et al. 1998).
Strengths of this study include a large-scale, well-established cohort design with large numbers of nonlung cancer deaths observed at many cancer sites. Air pollution exposures were estimated at each participant's residence using national-level exposure surfaces that have previously been used to examine mortality associations in CPS-II (Jerrett et al. 2016;Pope et al. 2015;Turner et al. 2014;. Detailed data were collected at enrollment on a variety of cancer risk factors including cigarette smoking, occupation, diet, and various hormonal and reproductive factors. The main limitation of this study is the use of cancer mortality rather than cancer incidence end points, with inferences of associations of ambient air pollution here reflecting both disease incidence and survival following diagnosis. Because lung cancer is rapidly fatal, with 5-y survival rates ranging from ∼ 13-18% for the periods 1987-1989 and 2005-2011 respectively, the use of mortality data reasonably approximates disease incidence (American Cancer Society 2016). Other rapidly fatal cancer sites include the pancreas (5-year survival = 4-8%), the liver and intrahepatic bile duct (5-18%), the esophagus (10-20%), the stomach (20-30%), and the brain and other nervous system sites (29-35%) (American Cancer Society 2016). In contrast, survival is greater for cancer at other sites, including the urinary bladder, the kidney and renal pelvis, and the colorectum, of interest here (5-y survival ranging from 57-79%), with survival from disease playing an increasing role in associations with ambient air pollution observed here (American Cancer Society 2016).
There is little research on whether ambient air pollution may be related to cancer progression or survival. One recent study of >350,000 California lung cancer patients reported that higher residential ambient air pollution concentrations (NO 2 , PM 2:5 , PM 10 ) were associated with poorer survival, particularly among patients diagnosed in earlier disease states (i.e., with localized disease) (Eckel et al. 2016). Mean ambient air pollution concentrations were also somewhat higher among those diagnosed with more advanced disease and among those with an unknown stage at diagnosis (i.e., either patients who were dying before stage information was obtained or who had limited workup performed), possibly reflecting differences in access to medical care, which may vary by levels of ambient air pollution. Little is known regarding possible impacts at other cancer sites, although reduced breast cancer survival was associated with higher PM 2:5 and PM 10 concentrations in another study (Hu et al. 2013). Further studies of nonlung cancer incidence are needed to disentangle the observed associations with ambient air pollution.
Covariate data were only available at enrollment and were not updated over the follow-up period in CPS-II. It is unlikely that CPS-II participants would begin smoking cigarettes over the follow-up time, given the mean age of >55 years at enrollment, although participants may increasingly become former smokers. There were also limited data on occupational exposure history. Participant residence data were also only available at enrollment. Changes in participant residence after enrollment as well as changes in coding of the underlying cause of death over the follow-up period would likely be nondifferential and would result in attenuation of the magnitude of the associations observed. A Canadian study observed little impact of accounting for residential mobility on PM 2:5 or O 3 mortality associations; however, associations with more spatially resolved NO 2 strengthened somewhat (Crouse et al. 2015).
There is also a lack of historical ambient air pollution data, although correlations between PM 2:5 and O 3 concentrations assigned to CPS-II participants over recent decades were moderately strong, ranging from approximately 0.6 to >0:8, indicating that the use of more recent ambient air pollution estimates may be reflective of longer-term exposure patterns (Krewski et al. 2009;Pope et al. 2002;Turner et al. 2016). The rank ordering of U.S. cities was also similar over time in the context of generally declining ambient air pollution concentrations. The use of recent ambient air pollution estimates may result in somewhat inflated HRs because increments of recent concentrations represent greater contrasts of historically higher concentrations (Krewski et al. 2009;Pope et al. 2002;Turner et al. 2016). Little is known regarding potential latency periods for cancer development in relation with ambient air pollution concentrations, which may also differ from those of lung cancer.
Correlations among pollutants were generally weak. Ambient air pollution concentrations were estimated using different approaches at different time periods and different geographic units of scale, possibly complicating interpretation of the correlation structure among pollutants. The LURBME model outperformed a range of other geostatistical and remote sensing PM 2:5 models in the CPS-II (Jerrett et al. 2016). Similar positive cardiorespiratory mortality-O 3 associations were observed in recent work with O 3 concentrations estimated at either 12 km × 12 km or 36 km × 36 km scales in the Eastern United States (Turner et al. 2016).
Owing to multiple testing and to the large number of cancer sites evaluated (n = 29), it is possible that some of the significant associations observed may be due to chance. As such, the results of this study should be replicated in other studies, particularly in studies of cancer incidence. Finally, our findings may not be entirely generalizable because CPS-II participants are of generally higher socioeconomic status and more limited racial/ethnic composition than the broader U.S. population.
Potential mechanisms through which ambient air pollution may be associated with other nonlung cancers remain to be fully elucidated; however, ambient air pollution represents a complex mixture of exposure to a broad range of carcinogenic and mutagenic substances, including PAHs and other aromatic hydrocarbons, benzene, metals, and xenoestrogens, which may be transported and metabolized in the body (Crouse et al. 2010;IARC 2012IARC , 2013). Populations exposed to outdoor air pollutants and to diesel engine exhaust have elevated urinary 1-hydroxypyrene and hemoglobin adducts of nitro-PAHs and low-molecular-weight alkenes (Ciarrocca et al. 2014;Duan et al. 2016;IARC 2012IARC , 2013. A recent study of 23,820 participants in Taiwan, including 464 incident hepatocellular carcinoma (HCC) cases, reported a positive association of PM 2:5 with HCC incidence, which may have been mediated by alanine transaminase levels, suggesting that PM 2:5 exposure may lead to HCC via chronic inflammation (Pan et al. 2015).
Active cigarette smoking has also been associated with a range of other nonlung cancers, including cancers in various urinary and digestive sites such as the bladder, the kidney, and the colorectum, among others (Carter et al. 2015;IARC 2009). However, levels of exposure to PM 2:5 from ambient air pollution are substantially lower and of differing chemical composition and toxicity compared with those from active cigarette smoking or occupational exposure .
Patients with a history of diabetes, which has also been linked with ambient air pollution (Eze et al. 2015;Pope et al. 2015), have also been observed to be at increased risk for bladder, kidney, and colorectal cancers, although it is unclear whether this is due to shared underlying risk factors (such as obesity) or to other metabolic features of the disease (such as hyperinsulinemia, hyperglycemia, or chronic inflammation) (Campbell et al. 2012;Giovannucci et al. 2010). The results herein were virtually unchanged with further adjustment for prevalent diabetes at enrollment (not shown).

Conclusion
The results from this large prospective study suggest that ambient air pollution was not associated with most nonlung cancer causes of death. Nonetheless, observed associations with mortality from kidney, bladder, and colorectal cancer merit further research, particularly in studies of cancer incidence.