were computed only for highly aggregated poststrata. The variance calculations had to assume that the aggregate correction factors did not vary for individual poststrata within the larger aggregates. If for no other reason, it is likely that the variance estimates for the Revision II A.C.E. estimates are biased downward.
Two factors merit discussion for the role they played in producing an estimated net overcount in the 2000 census. They are computer-based, whole-person census imputations and duplicate census enumerations.
The much larger number of whole-person imputations in 2000 (5.8 million) compared with 1990 (1.9 million) helps explain one of the initial puzzles regarding the original A.C.E. estimates of net undercount.17 The puzzle was that the original A.C.E. correct enumeration and match rates were very similar to the PES rates (see Tables 6.5 and 6.6). Other things equal, these similarities should have produced similar estimates of net undercount. Yet the original A.C.E. estimates showed marked reductions in net undercount rates from 1990 levels for such groups as minorities, renters, and children and a consequent narrowing of differences in net undercount rates between historically less-well-counted and better-counted groups (see Table 6.7).
The explanation lies in those census cases that had to be excluded from the A.C.E. because they were wholly imputed and hence could not be matched or because they were available too late for matching—the II term in the DSE formula (see Section 5-A). There were so many more of these cases in 2000 than in 1990 that when they were added back to the census counts for comparison with the DSE population estimates, the result was to lower the net undercount estimates for 2000 compared with 1990.
The IIs in 2000 included 2.4 million reinstated cases from the special summer 2000 MAF unduplication operation, whereas the