than the Interphone analyses (30–59 years), so a considerable proportion of the national data [e.g., 58% for Sweden (Lönn et al. 2005)] was not included in the overall pooled analyses. The national publications need to be consid- ered, therefore, as additional semi-independent sources of evidence, not simply as subsets of the overall Interphone analysis. The Interphone publication (Interphone Study Group 2010) compared 2,708 glioma cases diagnosed at 30–59 years of age dur- ing 2000–2004 with 2,972 controls, and 2,409 meningioma cases with 2,662 controls. Participation rates were 64% for glioma cases, 78% for meningioma cases, and 53% for con- trols, with considerable variation among study centers; proxies were used for 13% of glioma cases, 2% of meningioma cases, and 1% of con- trols. Sensitivity analyses did not suggest, how- ever, that the results depended on participation rates across centers or on inclusion of proxies. Key findings were a significantly dimin- ished risk of both glioma and meningioma in regular users compared with people who were not users or were occasional users (“nonusers”); no trend in risk of either tumor type with cumulative hours of use but an apparent raised risk of glioma, and to a lesser extent menin- gioma, in those in the top decile of cumula- tive hours of use; and no relation of risk of either tumor type to cumulative number of calls, years of use, or years since first use. These results raise several important issues. Reduced Risk of Brain Tumors in Mobile Phone Users The Interphone Study, as well as some previous case–control studies (Inskip et al. 2001; Muscat et al. 2000) and the only large cohort study (Schüz et al. 2006), identified a reduced risk of brain tumors among mobile phone users com- pared with nonusers. In the Interphone study as a whole, ever-regular use was associated with an odds ratio (OR) of 0.79 [95% confidence inter- val (CI): 0.68, 0.91] for meningioma and 0.81 (95% CI: 0.70, 0.94) for glioma. The pattern was consistent across the Interphone study sites and statistically precise, calling for explanation. There is empirical evidence that the reduced risks were attributable partly to nonresponse bias (Vrijheid et al. 2009). Cases and controls who initially declined to participate but agreed to complete a short nonresponse questionnaire had lower frequencies of regular mobile phone use than did those who participated fully. The quantitative results from this nonresponse questionnaire imply that selection bias would produce an OR of 0.87–0.92 if the hypoth- esis were true. It seems unlikely that differen- tial response based on mobile phone use could explain the diminished risk entirely, because the reduction in risk was similar for study centers that did and did not reveal to potential partici- pants the study’s focus on mobile phone use. Even if the same pattern of diminished response by nonusers occurred for cases and controls, which it did not, the overall greater nonparticipation among controls because of refusal would result in a downward bias in the OR. Whereas only 11% of glioma and meningioma cases refused to participate, 30% of controls did so. Furthermore, the phone use of those who did not complete even the nonresponse questionnaire (e.g., because of refusal or death) is unknown, adding further uncertainty to the extent of the overall bias. Other likely contributors to the dimin- ished ORs in users are prodromal symptoms such as headaches and impaired cognition, which may have prevented recent initiation of mobile phone use among subjects with as yet undiagnosed brain tumors. Thus, some cases who would otherwise have become short-term users may have remained nonusers, leading to artifactually reduced ORs for brain tumor in phone users, especially short-term users (and low cumulative users, because short-term use will tend to result in low cumulative use). It seems likely that this accounts for at least part of the decreased risk in users, because the strongest reduction in glioma risk was found in the shortest-term users. Other potential con- tributors to diminished ORs can be hypothe- sized, but there is no evidence for them [see Supplemental Material, p. 1 (http://dx.doi. org/10.1289/ehp.1103693)]. The appropriate analytic approach and interpretation in light of this presumably non- causal reduction in risk are not obvious. One suggested response has been to alter the refer- ent group, by using low regular use rather than nonuse plus occasional use as the referent. This results in an upward shift in the ORs across the board, more for glioma than for meningioma, but no change in the magnitude of those ORs relative to one another across the range of exposure (Interphone Study Group 2010). However, whether this decreases or increases the bias depends on two factors: whether the diminished risk is attributable to nonresponse, and whether the biases apply to low-level users as well as nonusers. Neither of these factors is known, but to the extent that the diminished risk is attributable to prodromal symptoms, changing the referent group would produce upward bias. If short-term users (or low cumu- lative users) are used as the referent exposure group, the more pronounced risk reduction in this group caused by prodromal symptoms would bias relative risks for long-term users (or high cumulative users) upward. Risks after Prolonged and Heavy Mobile Phone Use If exposure to RF fields through mobile phone use were tumorigenic, people using mobile phones longest and those who were the heaviest users would be expected to show Environmental Health Perspectives • volume 119 | number 11 | November 2011 Mobile phones and brain tumor risk the highest risks of brain tumors. Reliability of recall of amount of use a decade ago is unknown, and the average amount of use is likely to have shifted over time as phone use has escalated universally. Validation studies of recall of phone use in the previous 6 months, and up to approximately 5 years in the past, have found that even in the short term, sub- jects on average underestimate the number of calls per month but overestimate dura- tion of calls, with moderate systematic error (underestimation by light users, overestima- tion by heavy users) and a large amount of random error (Vrijheid et al. 2006). Recall of number of calls was found to be better than recall of their duration. Furthermore, in the Interphone study cases more often than con- trols gave implausibly high estimates of daily time spent on calls (e.g., 10 cases and no con- trols reported average use of > 12 hr/day). A validation study that included both cases and controls found that cases overestimated phone use in distant time periods, which could cause positive bias in risk estimates (Vrijheid et al. 2009). It thus appears that recall of amount of use was appreciably erroneous and quite likely different for cases than for controls. It is possible that recall of year of first use, and hence duration of use, may have been more reliable than recall of amount of use. Notwithstanding the inherent unreli- ability of recalled amount of use, the only cumulative mobile phone exposure measures available in the Interphone study were dura- tion and amount. Neither yielded material evidence of a positive association with brain tumors. Specifically, for the longest-term users (≥ 10 years since first use), no association was found for glioma (OR = 0.98; 95% CI: 0.76, 1.26) or meningioma (OR = 0.83; 95% CI: 0.61, 1.14). Most ORs were < 1.0, and no dose–response pattern was found. This is con- sistent with results from a cohort study based on subscriber lists (Schüz et al. 2006) but in contrast with the raised risks for long-term use reported by Hardell et al. (2006a, 2006b). For heavy use measured by estimated total num- ber of calls, again, there was no positive asso- ciation with brain tumors: ORs were < 1.0 in all categories of number of calls, includ- ing those in the top decile, for both glioma and meningioma. For heavy use assessed by cumulative duration of calls, again, there was no dose–response effect for either type of tumor. For glioma, although the risk estimate for subjects in the highest decile of total call time (≥ 1,640 hr) was modestly raised at OR = 1.40 (95% CI: 1.03, 1.89), it was disjointed from the risk in the next heaviest users, the second highest decile, which had one of the lowest risk estimates: OR = 0.71 (95% CI: 0.53, 0.96). Similarly, for meningioma the OR in the highest decile of total call-time OR was 1.15 (95% CI: 0.81, 1.62), whereas in the 1535